G * Power 3.1 manual
June 1, 2023
This manual is not yet complete. We will be adding help on more tests in the future. If you cannot find help for your testin this version of the manual, then please check the G*Power website to see if a more up-to-date version of the manual
has been made available.

Contents

21 t test: Means - difference between two independent
means (two groups)
52

1

Introduction

2

2

The G * Power calculator

7

3

Exact: Correlation - Difference from constant (onesample case)922 Wilcoxon signed-rank test: Means - difference fromconstant (one sample case)53
23 Wilcoxon signed-rank test: (matched pairs)

4

Exact: Proportion - difference from constant (onesample case)
11

5

Exact: Proportion - inequality, two dependentgroups (McNemar)
14

6

Exact: Proportions - inequality of two independentgroups (Fisher’s exact-test)
17

55

24 Wilcoxon-Mann-Whitney test of a difference between two independent means
59
25 t test: Generic case63

26 χ2 test: Variance - difference from constant (onesample case)64
27 z test: Correlation - inequality of two independent
Pearson r’s
65

7

Exact test: Multiple Regression - random model

18

8

Exact: Proportion - sign test

22

28 z test: Correlation - inequality of two dependent
Pearson r’s
66

9

Exact: Generic binomial test

23

29 Z test: Multiple Logistic Regression

70

24

30 Z test: Poisson Regression

75

31 Z test: Tetrachoric Correlation

80

References

84

10 F test: Fixed effects ANOVA - one way

11 F test: Fixed effects ANOVA - special, main effects
and interactions
26
12 t test: Linear Regression (size of slope, one group) 31
13 F test: Multiple Regression - omnibus (deviation of
R2 from zero), fixed model
33
14 F test: Multiple Regression - special (increase of
R2 ), fixed model
36
15 F test: Inequality of two Variances

39

16 t test: Correlation - point biserial model40

17 t test: Linear Regression (two groups)

42

18 t test: Linear Regression (two groups)

45

19 t test: Means - difference between two dependent
means (matched pairs)
48
20 t test: Means - difference from constant (one sample case)
50

1

1

Introduction

Distribution-based approach to test selection First selectthe family of the test statistic (i.e., exact, F −, t−, χ2 , or ztest) using the Test family menu in the main window. The
Statistical test menu adapts accordingly, showing a list of all
tests available for the test family.

G * Power (Fig. 1 shows the main window of the program)covers statistical power analyses for many different statistical tests of the• F test,

Example: For the two groups t-test, first select the test family
based on the t distribution.

• t test,• χ2 -test and• z test families and some
• exact tests.
G * Power provides effect size calculators and graphics
options. G * Power supports both a distribution-based anda design-based input mode. It contains also a calculator that
supports many central and noncentral probability distributions.
G * Power is free software and available for Mac OS X
and Windows XP/Vista/7/8.

1.1

Then select Means: Difference between two independent means(two groups) option in the Statictical test menu.

Types of analysis

G * Power offers five different types of statistical power
analysis:
1. A priori (sample size N is computed as a function of
power level 1 − β, significance level α, and the to-bedetected population effect size)
2. Compromise (both α and 1 − β are computed as functions of effect size, N, and an error probability ratioq = β/α)3. Criterion (α and the associated decision criterion arecomputed as a function of 1 − β, the effect size, and N)

Design-based approach to the test selection Alternatively, one might use the design-based approach. With theTests pull-down menu in the top row it is possible to select

4. Post-hoc (1 − β is computed as a function of α, the population effect size, and N)• the parameter class the statistical test refers to (i.e.,correlations and regression coefficients, means, proportions, or variances), and

5. Sensitivity (population effect size is computed as afunction of α, 1 − β, and N)

1.2• the design of the study (e.g., number of groups, independent vs. dependent samples, etc.).

Program handling

Perform a Power Analysis Using G * Power typically involves the following three steps:

The design-based approach has the advantage that test options referring to the same parameter class (e.g., means) arelocated in close proximity, whereas they may be scatteredacross different distribution families in the distributionbased approach.

1. Select the statistical test appropriate for your problem.
2. Choose one of the five types of power analysis available
3. Provide the input parameters required for the analysis
and click "Calculate".

Example: In the Tests menu, select Means, then select Two independent groups" to specify the two-groups t test.

Plot parameters In order to help you explore the parameter space relevant to your power analysis, one parameter(α, power (1 − β), effect size, or sample size) can be plotted
as a function of another parameter.
1.2.1

Select the statistical test appropriate for your problem

In Step 1, the statistical test is chosen using the distributionbased or the design-based approach.
2

Figure 1: The main window of G * Power

1.2.2• the required power level (1 − β),Choose one of the five types of power analysis
available• the pre-specified significance level α, andIn Step 2, the Type of power analysis menu in the center ofthe main window is used to choose the appropriate analysis
type and the input and output parameters in the windowchange accordingly.

• the population effect size to be detected with probability (1 − β).
In a criterion power analysis, α (and the associated decision criterion) is computed as a function of

Example: If you choose the first item from the Type of power
analysis menu the main window will display input and output
parameters appropriate for an a priori power analysis (for t tests
for independent groups if you followed the example provided
in Step 1).

• 1-β,• the effect size, and• a given sample size.
In a compromise power analysis both α and 1 − β arecomputed as functions of• the effect size,• N, and• an error probability ratio q = β/α.
In a post-hoc power analysis the power (1 − β) is computed as a function of

In an a priori power analysis, sample size N is computed
as a function of
3

• α,

Because Cohen’s book on power analysis Cohen (1988)
appears to be well known in the social and behavioral sciences, we made use of his effect size measures wheneverpossible. In addition, wherever available G * Power provides his definitions of "‘small"’, "‘medium"’, and "‘large"’
effects as "‘Tool tips"’. The tool tips may be optained bymoving the cursor over the "‘effect size"’ input parameterfield (see below). However, note that these conventions may
have different meanings for different tests.

• the population effect size parameter, and• the sample size(s) used in a study.
In a sensitivity power analysis the critical population effect size is computed as a function of• α,• 1 − β, andExample: The tooltip showing Cohen’s measures for the effectsize d used in the two groups t test

• N.
1.2.3

Provide the input parameters required for the analysis

In Step 3, you specify the power analysis input parametersin the lower left of the main window.
Example: An a priori power analysis for a two groups t test
would require a decision between a one-tailed and a two-tailed
test, a specification of Cohen’s (1988) effect size measure d under H1 , the significance level α, the required power (1 − β) of
the test, and the preferred group size allocation ratio n2 /n1 .
Let us specify input parameters forIf you are not familiar with Cohen’s measures, if you
think they are inadequate for your test problem, or if you
have more detailed information about the size of the to-beexpected effect (e.g., the results of similar prior studies),
then you may want to compute Cohen’s measures from
more basic parameters. In this case, click on the Determine
button to the left the effect size input field. A drawer will
open next to the main window and provide access to aneffect size calculator tailored to the selected test.

• a one-tailed t test,• a medium effect size of d = .5,• α = .05,
• (1 − β) = .95, and• an allocation ratio of n2 /n1 = 1

Example: For the two-group t-test users can, for instance, specify the means µ1 , µ2 and the common standard deviation (σ =σ1 = σ2 ) in the populations underlying the groups to calculate Cohen’s d = |µ1 − µ2 |/σ. Clicking the Calculate andtransfer to main window button copies the computed effectsize to the appropriate field in the main window

This would result in a total sample size of N = 176 (i.e., 88
observation units in each group). The noncentrality parameterδ defining the t distribution under H1 , the decision criterion
to be used (i.e., the critical value of the t statistic), the degrees
of freedom of the t test and the actual power value are also
displayed.

In addition to the numerical output, G * Power displays
the central (H0 ) and the noncentral (H1 ) test statistic distributions along with the decision criterion and the associatederror probabilities in the upper part of the main window.
This supports understanding the effects of the input parameters and is likely to be a useful visualization tool in
the teaching of, or the learning about, inferential statistics.

Note that the actual power will often be slightly larger
than the pre-specified power in a priori power analyses. The
reason is that non-integer sample sizes are always rounded
up by G * Power to obtain integer values consistent with apower level not less than the pre-specified one.
4

The distributions plot may be copied, saved, or printed byclicking the right mouse button inside the plot area.

The button X-Y plot for a range of values at to bottom of
the main window opens the plot window.

Example: The menu appearing in the distribution plot for thet-test after right clicking into the plot.

By selecting the appropriate parameters for the y and thex axis, one parameter (α, power (1 − β), effect size, or sample size) can be plotted as a function of another parameter. Of the remaining two parameters, one can be chosen to
draw a family of graphs, while the fourth parameter is kept
constant. For instance, power (1 − β) can be drawn as afunction of the sample size for several different population
effects sizes, keeping α at a particular value.
The plot may be printed, saved, or copied by clicking theright mouse button inside the plot area.
Selecting the Table tab reveals the data underlying theplot (see Fig. 3); they may be copied to other applications
by selecting, cut and paste.

The input and output of each power calculation in a
G*Power session are automatically written to a protocolthat can be displayed by selecting the "Protocol of power
analyses" tab in the main window. You can clear the protocol, or to save, print, and copy the protocol in the same way
as the distributions plot.

Note: The Power Plot window inherits all input parameters of the analysis that is active when the X-Y plot
for a range of values button is pressed. Only someof these parameters can be directly manipulated in thePower Plot window. For instance, switching from a plotof a two-tailed test to that of a one-tailed test requires
choosing the Tail(s): one option in the main window, followed by pressing the X-Y plot for range of values button.

(Part of) the protocol window.

1.2.4

Plotting of parameters

G * Power provides to possibility to generate plots of oneof the parameters α, effectsize, power and sample size, depending on a range of values of the remaining parameters.
The Power Plot window (see Fig. 2) is opened by clicking the X-Y plot for a range of values button locatedin the lower right corner of the main window. To ensurethat all relevant parameters have valid values, this button is
only enabled if an analysis has successfully been computed(by clicking on calculate).
The main output parameter of the type of analysis selected in the main window is by default selected as the dependent variable y. In an a prior analysis, for instance, thisis the sample size.
5

Figure 2: The plot window of G * Power

Figure 3: The table view of the data for the graphs shown in Fig. 2

6

2

The G * Power calculator• sign(x) - Sign of x: x < 0 → −1, x = 0 → 0, x > 0 →1.

G * Power contains a simple but powerful calculator thatcan be opened by selecting the menu label "Calculator" inthe main window. Figure 4 shows an example session. This
small example script calculates the power for the one-tailedt test for matched pairs and demonstrates most of the available features:• lngamma(x) Natural logarithm of the gamma functionln(Γ( x ))• frac(x) - Fractional part of floating point x: frac(1.56)
is 0.56.
• int(x) - Integer part of float point x: int(1.56) is 1.

• There can be any number of expressions• min(x,y) - Minimum of x and y• The result is set to the value of the last expression in
the script• max(x,y) - Maximum of x and y• Several expression on a line are separated by a semicolon• uround(x,m) - round x up to a multiple of muround(2.3, 1) is 3, uround(2.3, 2) = 4.

• Expressions can be assigned to variables that can beused in following expressions

Supported distribution functions (CDF = cumulative
distribution function, PDF = probability density function, Quantile = inverse of the CDF). For information about the properties of these distributions check
http://mathworld.wolfram.com/.

• The character # starts a comment. The rest of the line
following # is ignored• Many standard mathematical functions like squareroot, sin, cos etc are supported (for a list, see below)• zcdf(x) - CDF
zpdf(x) - PDF
zinv(p) - Quantileof the standard normal distribution.

• Many important statistical distributions are supported(see list below)• normcdf(x,m,s) - CDF
normpdf(x,m,s) - PDF
norminv(p,m,s) - Quantileof the normal distribution with mean m and standard
deviation s.

• The script can be easily saved and loaded. In this way
a number of useful helper scripts can be created.
The calculator supports the following arithmetic operations (shown in descending precedence):• Power: ^• Multiply: ∗• Divide: /• chi2cdf(x,df) - CDF
chi2pdf(x,df) - PDF
chi2inv(p,df) - Quantileof the chi square distribution with d f degrees of freedom: χ2d f ( x ).

(2^3 = 8)(2 ∗ 2 = 4)
(6/2 = 3)• Plus: +(2 + 3 = 5)

• Minus: -(3 − 2 = 1)

• fcdf(x,df1,df2) - CDF
fpdf(x,df1,df2) - PDF
finv(p,df1,df2) - Quantileof the F distribution with d f 1 numerator and d f 2 denominator degrees of freedom Fd f1 ,d f2 ( x ).

Supported general functions• abs(x) - Absolute value | x |• tcdf(x,df) - CDF
tpdf(x,df) - PDF
tinv(p,df) - Quantileof the Student t distribution with d f degrees of freedom td f ( x ).

• sin(x) - Sine of x
• asin(x) - Arc sine of x
• cos(x) - Cosine of x
• acos(x) - Arc cosine of x

• ncx2cdf(x,df,nc) - CDF
ncx2pdf(x,df,nc) - PDF
ncx2inv(p,df,nc) - Quantile
of noncentral chi square distribution with d f degrees
of freedom and noncentrality parameter nc.

• tan(x) - Tangent of x
• atan(x) - Arc tangent of x
• atan2(x,y) - Arc tangent of y/x• ncfcdf(x,df1,df2,nc) - CDF
ncfpdf(x,df1,df2,nc) - PDF
ncfinv(p,df1,df2,nc) - Quantile
of noncentral F distribution with d f 1 numerator andd f 2 denominator degrees of freedom and noncentrality
parameter nc.

• exp(x) - Exponential e x• log(x) - Natural logarithm ln( x )√• sqrt(x) - Square root x• sqr(x) - Square x2

7

Figure 4: The G * Power calculator• mr2cdf(R2 , ρ2 ,k,N) - CDF
mr2pdf(R2 , ρ2 ,k,N) - PDF
mr2inv(p,ρ2 ,k,N) - Quantileof the distribution of the sample squared multiple correlation coefficient R2 for population squared multiple
correlation coefficient ρ2 , k − 1 predictors, and samplesof size N.

• nctcdf(x,df,nc) - CDF
nctpdf(x,df,nc) - PDF
nctinv(p,df,nc) - Quantile
of noncentral Student t distribution with d f degrees of
freedom and noncentrality parameter nc.
• betacdf(x,a,b) - CDF
betapdf(x,a,b) - PDF
betainv(p,a,b) - Quantileof the beta distribution with shape parameters a and b.

• logncdf(x,m,s) - CDF
lognpdf(x,m,s) - PDF
logninv(p,m,s) - Quantileof the log-normal distribution, where m, s denote meanand standard deviation of the associated normal distribution.

• poisscdf(x,λ) - CDF
poisspdf(x,λ) - PDF
poissinv(p,λ) - Quantile
poissmean(x,λ) - Meanof the poisson distribution with mean λ.

• laplcdf(x,m,s) - CDF
laplpdf(x,m,s) - PDF
laplinv(p,m,s) - Quantileof the Laplace distribution, where m, s denote locationand scale parameter.

• binocdf(x,N,π) - CDF
binopdf(x,N,π) - PDF
binoinv(p,N,π) - Quantileof the binomial distribution for sample size N and success probability π.

• expcdf(x,λ - CDF
exppdf(x,λ) - PDF
expinv(p,λ - Quantileof the exponential distribution with parameter λ.

• hygecdf(x,N,ns,nt) - CDF
hygepdf(x,N,ns,nt) - PDF
hygeinv(p,N,ns,nt) - Quantileof the hypergeometric distribution for samples of size
N from a population of total size nt with ns successes.

• unicdf(x,a,b) - CDF
unipdf(x,a,b) - PDF
uniinv(p,a,b) - Quantileof the uniform distribution in the intervall [ a, b].

• corrcdf(r,ρ,N) - CDF
corrpdf(r,ρ,N) - PDF
corrinv(p,ρ,N) - Quantileof the distribution of the sample correlation coefficientr for population correlation ρ and samples of size N.

8

3

Exact: Correlation - Difference fromconstant (one sample case)

asymptotically identical, that is, they produce essentially the same results if N is large. Therefore, a threshold value x for N can be specified that determines thetransition between both procedures. The exact procedure is used if N < x, the approximation otherwise.

The null hypothesis is that in the population the true correlation ρ between two bivariate normally distributed random variables has the fixed value ρ0 . The (two-sided) alternative hypothesis is that the correlation coefficient has adifferent value: ρ ̸= ρ0 :
H0 :
H1 :

2. Use large sample approximation (Fisher Z). With this option you select always to use the approximation.
There are two properties of the output that can be used
to discern which of the procedures was actually used: The
option field of the output in the protocol, and the namingof the critical values in the main window, in the distribution
plot, and in the protocol (r is used for the exact distribution
and z for the approximation).

ρ − ρ0 = 0ρ − ρ0 ̸= 0.

A common special case is ρ0 = 0 (see e.g. Cohen, 1969,
Chap. 3). The two-sided test (“two tails”) should be used
if there is no restriction on the direction of the deviationof the sample r from ρ0 . Otherwise use the one-sided test(“one tail”).

3.1

3.3

In the null hypothesis we assume ρ0 = 0.60 to be the correlation coefficient in the population. We further assume that
our treatment increases the correlation to ρ = 0.65. If werequire α = β = 0.05, how many subjects do we need in a
two-sided test?

Effect size index

To specify the effect size, the conjectured alternative correlation coefficient ρ should be given. ρ must conform to thefollowing restrictions: −1 + ε < ρ < 1 − ε, with ε = 10−6 .
The proper effect size is the difference between ρ and ρ0 :
ρ − ρ0 . Zero effect sizes are not allowed in a priori analyses.
G * Power therefore imposes the additional restriction that|ρ − ρ0 | > ε in this case.
For the special case ρ0 = 0, Cohen (1969, p.76) defines thefollowing effect size conventions:• Select
Type of power analysis: A priori• OptionsUse exact distribution if N <: 10000• Input
Tail(s): Two
Correlation ρ H1: 0.65
α err prob: 0.05
Power (1-β err prob): 0.95
Correlation ρ H0: 0.60

• small ρ = 0.1• medium ρ = 0.3• large ρ = 0.5
Pressing the Determine button on the left side of the effect size label opens the effect size drawer (see Fig. 5). You
can use it to calculate |ρ| from the coefficient of determination r2 .

• Output
Lower critical r: 0.570748
Upper critical r: 0.627920
Total sample size: 1928
Actual power: 0.950028
In this case we would reject the null hypothesis if we observed a sample correlation coefficient outside the interval [0.571, 0.627]. The total sample size required to ensure
a power (1 − β) > 0.95 is 1928; the actual power for this N
is 0.950028.
In the example just discussed, using the large sample approximation leads to almost the same sample size N =1929. Actually, the approximation is very good in mostcases. We now consider a small sample case, where thedeviation is more pronounced: In a post hoc analysis ofa two-sided test with ρ0 = 0.8, ρ = 0.3, sample size 8, and
α = 0.05 the exact power is 0.482927. The approximation
gives the slightly lower value 0.422599.

Figure 5: Effect size dialog to determine the coefficient of determination from the correlation coefficient ρ.

3.2

ExamplesOptions

The procedure uses either the exact distribution of the correlation coefficient or a large sample approximation basedon the z distribution. The options dialog offers the following choices:

3.4

Related tests

Similar tests in G * Power 3.0:

1. Use exact distribution if N < x. The computation time ofthe exact distribution increases with N, whereas thatof the approximation does not. Both procedures are• Correlation: Point biserial model• Correlations: Two independent Pearson r’s (two samples)
93.5

Implementation notes

Exact distribution. The H0 -distribution is the sample correlation coefficient distribution sr (ρ0 , N ), the H1 distribution is sr (ρ, N ), where N denotes the total sample size, ρ0 denotes the value of the baseline correlation
assumed in the null hypothesis, and ρ denotes the ‘alternative correlation’. The (implicit) effect size is ρ − ρ0 . The
algorithm described in Barabesi and Greco (2002) is used to
calculate the CDF of the sample coefficient distribution.
Large sample approximation. The H0 -distribution is thestandard normal distribution N (0, 1), the H1-distribution is
N ( Fz (ρ) − Fz (ρ0 ))/σ, 1), with Fz (r )p= ln((1 + r )/(1 − r ))/2
(Fisher z transformation) and σ = 1/( N − 3).

3.6

Validation

The results in the special case of ρ0 = 0 were comparedwith the tabulated values published in Cohen (1969). The
results in the general case were checked against the values
produced by PASS (Hintze, 2006).

10

4

Exact: Proportion - difference fromconstant (one sample case)

The relational value given in the input field on the left sideand the two proportions given in the two input fields on theright side are automatically synchronized if you leave oneof the input fields. You may also press the Sync values
button to synchronize manually.
Press the Calculate button to preview the effect size gresulting from your input values. Press the Transfer tomain window button to (1) to calculate the effect size g =π − π0 = P2 − P1 and (2) to change, in the main window,
the Constant proportion field to P1 and the Effect sizeg field to g as calculated.

The problem considered in this case is whether the probability π of an event in a given population has the constantvalue π0 (null hypothesis). The null and the alternative hypothesis can be stated as:
H0 :
H1 :

π − π0 = 0π − π0 ̸= 0.

A two-tailed binomial tests should be performed to test
this undirected hypothesis. If it is possible to predict a priori the direction of the deviation of sample proportions p
from π0 , e.g. p − π0 < 0, then a one-tailed binomial test
should be chosen.

4.1

4.2

Options

The binomial distribution is discrete. It is thus not normallypossible to arrive exactly at the nominal α-level. For twosided tests this leads to the problem how to “distribute” α
to the two sides. G * Power offers the three options listed
here, the first option being selected by default:

Effect size index

The effect size g is defined as the deviation from the constant probability π0 , that is, g = π − π0 .
The definition of g implies the following restriction: ε ≤(π0 + g) ≤ 1 − ε. In an a priori analysis we need to respect the additional restriction | g| > ε (this is in accordancewith the general rule that zero effect hypotheses are undefined in a priori analyses). With respect to these constraints,
G * Power sets ε = 10−6 .
Pressing the Determine button on the left side of the effect size label opens the effect size drawer:

1. Assign α/2 to both sides: Both sides are handled independently in exactly the same way as in a one-sided
test. The only difference is that α/2 is used instead of
α. Of the three options offered by G * Power , this one
leads to the greatest deviation from the actual α (in post
hoc analyses).
2. Assign to minor tail α/2, then rest to major tail (α2 =
α/2, α1 = α − α2 ): First α/2 is applied to the side of
the central distribution that is farther away from thenoncentral distribution (minor tail). The criterion used
for the other side is then α − α1 , where α1 is the actual
α found on the minor side. Since α1 ≤ α/2 one canconclude that (in post hoc analyses) the sum of the actual values α1 + α2 is in general closer to the nominal
α-level than it would be if α/2 were assigned to bothside (see Option 1).
3. Assign α/2 to both sides, then increase to minimize the difference of α1 + α2 to α: The first step is exactly the sameas in Option 1. Then, in the second step, the criticalvalues on both sides of the distribution are increased(using the lower of the two potential incremental αvalues) until the sum of both actual α values is as close
as possible to the nominal α.

You can use this dialog to calculate the effect size g fromπ0 (called P1 in the dialog above) and π (called P2 in the
dialog above) or from several relations between them. If you
open the effect dialog, the value of P1 is set to the value inthe constant proportion input field in the main window.
There are four different ways to specify P2:

Press the Options button in the main window to selectone of these options.

4.3

ExamplesWe assume a constant proportion π0 = 0.65 in the population and an effect size g = 0.15, i.e. π = 0.65 + 0.15 = 0.8.
We want to know the power of a one-sided test given
α = .05 and a total sample size of N = 20.

1. Direct input: Specify P2 in the corresponding input field
below P1
2. Difference: Choose difference P2-P1 and insert thedifference into the text field on the left side (the difference is identical to g).

• Select
Type of power analysis: Post hoc

3. Ratio: Choose ratio P2/P1 and insert the ratio value
into the text field on the left side

• Options
Alpha balancing in two-sided tests: Assign α/2 on bothsides

4. Odds ratio: Choose odds ratio and insert the odds ratio ( P2/(1 − P2))/( P1/(1 − P1)) between P1 and P2
into the text field on the left side.
11

Figure 6: Distribution plot for the example (see text)

would choose N = 16 as the result of a search for the sample size that leads to a power of at least 0.3. All types ofpower analyses except post hoc are confronted with similar problems. To ensure that the intended result has beenfound, we recommend to check the results from these typesof power analysis by a power vs. sample size plot.

• Input
Tail(s): One
Effect size g: 0.15
α err prob: 0.05
Total sample size: 20
Constant proportion: 0.65
• Output
Lower critical N: 17
Upper critical N: 17
Power (1-β err prob): 0.411449
Actual α: 0.044376

4.4

Related tests

Similar tests in G * Power 3.0:
• Proportions: Sign test.

The results show that we should reject the null hypothesis of π = 0.65 if in 17 out of the 20 possible cases therelevant event is observed. Using this criterion, the actual α
is 0.044, that is, it is slightly lower than the requested α of
5%. The power is 0.41.
Figure 6 shows the distribution plots for the example. The
red and blue curves show the binomial distribution underH0 and H1 , respectively. The vertical line is positioned at
the critical value N = 17. The horizontal portions of thegraph should be interpreted as the top of bars ranging fromN − 0.5 to N + 0.5 around an integer N, where the heightof the bars correspond to p( N ).
We now use the graphics window to plot power values for a range of sample sizes. Press the X-Y plot fora range of values button at the bottom of the main window to open the Power Plot window. We select to plot thepower as a function of total sample size. We choose a rangeof samples sizes from 10 in steps of 1 through to 50. Next,
we select to plot just one graph with α = 0.05 and effectsize g = 0.15. Pressing the Draw Plot button produces theplot shown in Fig. 7. It can be seen that the power does notincrease monotonically but in a zig-zag fashion. This behavior is due to the discrete nature of the binomial distributionthat prevents that arbitrary α value can be realized. Thus,
the curve should not be interpreted to show that the power
for a fixed α sometimes decreases with increasing sample
size. The real reason for the non-monotonic behaviour isthat the actual α level that can be realized deviates more orless from the nominal α level for different sample sizes.
This non-monotonic behavior of the power curve poses a
problem if we want to determine, in an a priori analysis, theminimal sample size needed to achieve a certain power. In
these cases G * Power always tries to find the lowest sample size for which the power is not less than the specified
value. In the case depicted in Fig. 7, for instance, G * Power

4.5

Implementation notes

The H0 -distribution is the Binomial distribution B( N, π0 ),the H1 -distribution the Binomial distribution B( N, g + π0 ).
N denotes the total sample size, π0 the constant proportion
assumed in the null hypothesis, and g the effect size index
as defined above.

4.6

Validation

The results of G * Power for the special case of the sign test,
that is π0 = 0.5, were checked against the tabulated values
given in Cohen (1969, chapter 5). Cohen always chose fromthe realizable α values the one that is closest to the nominal
value even if it is larger then the nominal value. G * Power, in contrast, always requires the actual α to be lower then
the nominal value. In cases where the α value chosen by
Cohen happens to be lower then the nominal α, the results
computed with G * Power were very similar to the tabulated values. In the other cases, the power values computedby G * Power were lower then the tabulated ones.
In the general case (π0 ̸= 0.5) the results of post hoc
analyses for a number of parameters were checked against
the results produced by PASS (Hintze, 2006). No differenceswere found in one-sided tests. The results for two-sided
tests were also identical if the alpha balancing method “Assign α/2 to both sides” was chosen in G * Power .

12

Figure 7: Plot of power vs. sample size in the binomial test (see text)13

5

Exact: Proportion - inequality, twodependent groups (McNemar)

the central distribution that is farther away from thenoncentral distribution (minor tail). The criterion usedon the other side is then α − α1 , where α1 is the actual
α found on the minor side. Since α1 ≤ α/2 one canconclude that (in post hoc analyses) the sum of the actual values α1 + α2 is in general closer to the nominal
α-level than it would be if α/2 were assigned to bothsides (see Option 1).

This procedure relates to tests of paired binary responses.
Such data can be represented in a 2 × 2 table:
Treatment
YesNo

Standard
Yes
No
π11
π12
π21
π22
πs 1 − πs3. Assign α/2 to both sides, then increase to minimize the difference of α1 + α2 to α: The first step is exactly the sameas in Option 1. Then, in the second step, the criticalvalues on both sides of the distribution are increased(using the lower of the two potential incremental αvalues) until the sum of both actual α values is as close
as possible to the nominal α.

πt1 − πt1

where πij denotes the probability of the respective response. The probability π D of discordant pairs, that is, theprobability of yes/no-response pairs, is given by π D =π12 + π21 . The hypothesis of interest is that πs = πt , which
is formally identical to the statement π12 = π21 .
Using this fact, the null hypothesis states (in a ratio notation) that π12 is identical to π21 , and the alternative hypothesis states that π12 and π21 are different:
H0 :
H1 :

5.2.2

You may choose between an exact procedure and a fasterapproximation (see implementation notes for details):
1. Exact (unconditional) power if N < x. The computation
time of the exact procedure increases much faster withsample size N than that of the approximation. Given
that both procedures usually produce very similar results for large sample sizes, a threshold value x for N
can be specified which determines the transition between both procedures. The exact procedure is used if
N < x; the approximation is used otherwise.

π12 /π21 = 1
π12 /π21 ̸= 1.

In the context of the McNemar test the term odds ratio (OR)denotes the ratio π12 /π21 that is used in the formulation of
H0 and H1 .

5.1

Effect size index

Note: G * Power does not show distribution plots forexact computations.

The Odds ratio π12 /π21 is used to specify the effect size.
The odds ratio must lie inside the interval [10−6 , 106 ]. An
odds ratio of 1 corresponds to a null effect. Therefore thisvalue must not be used in a priori analyses.
In addition to the odds ratio, the proportion of discordantpairs, i.e. π D , must be given in the input parameter field
called Prop discordant pairs. The values for this proportion must lie inside the interval [ε, 1 − ε], with ε = 10−6 .
If π D and d = π12 − π21 are given, then the odds ratio
may be calculated as: OR = (d + π D )/(d − π D ).

5.2

2. Faster approximation (assumes number of discordant pairs
to be constant). Choosing this option instructs G * Powerto always use the approximation.

5.3

ExamplesAs an example we replicate the computations in O’Brien(2002, p. 161-163). The assumed table is:Treatment
Yes
No

Options

Press the Options button in the main window to select oneof the following options.
5.2.1

Computation

Standard
Yes No.54 .08.32 .06.86 .14.62.38
1

In this table the proportion of discordant pairs is π D =.32 + .08 = 0.4 and the Odds Ratio OR = π12 /π21 =0.08/.32 = 0.25. We want to compute the exact power fora one-sided test. The sample size N, that is, the number ofpairs, is 50 and α = 0.05.

Alpha balancing in two-sided tests

The binomial distribution is discrete. It is therefore notnormally possible to arrive at the exact nominal α-level.
For two-sided tests this leads to the problem how to “distribute” α to the two sides. G * Power offers the three options listed here, the first option being selected by default:• Select
Type of power analysis: Post hoc

1. Assign α/2 to both sides: Both sides are handled independently in exactly the same way as in a one-sided
test. The only difference is that α/2 is used instead of
α. Of the three options offered by G * Power , this one
leads to the greatest deviation from the actual α (in post
hoc analyses).

• Options
Computation: Exact• Input
Tail(s): One
Odds ratio: 0.25
α err prob: 0.05
Total sample size: 50
Prop discordant pairs: 0.4

2. Assign to minor tail α/2, then rest to major tail (α2 =
α/2, α1 = α − α2 ): First α/2 is applied to the side of14• Output
Power (1-β err prob): 0.839343
Actual α: 0.032578
Proportion p12: 0.08
Proportion p21: 0.32

also in the two-tailed case when the alpha balancing Option 2 (“Assign to minor tail α/2, then rest to major tail”,
see above) was chosen in G * Power .
We also compared the exact results of G * Power generated for a large range of parameters to the results produced
by PASS (Hintze, 2006) for the same scenarios. We foundcomplete correspondence in one-sided test. In two-sided
tests PASS uses an alpha balancing strategy corresponding to Option 1 in G * Power (“Assign α/2 on both sides”,
see above). With two-sided tests we found small deviations
between G * Power and PASS (about ±1 in the third decimal place), especially for small sample sizes. These deviations were always much smaller than those resulting froma change of the balancing strategy. All comparisons withPASS were restricted to N < 2000, since for larger N the exact routine in PASS sometimes produced nonsensical values(this restriction is noted in the PASS manual).

The power calculated by G * Power (0.839343) corresponds
within the given precision to the result computed by
O’Brien (0.839). Now we use the Power Plot window to calculate the power for several other sample sizes and to generate a graph that gives us an overview of a section of theparameter space. The Power Plot window can be opened
by pressing the X-Y plot for a range of values buttonin the lower part of the main window.
In the Power Plot window we choose to plot the poweron the Y-Axis (with markers and displaying the values
in the plot) as a function of total sample size. The sample
sizes shall range from 50 in steps of 25 through to 150. We
choose to draw a single plot. We specify α = 0.05 and Odds
ratio = 0.25.
The results shown in figure 8 replicate exactly the values
in the table in O’Brien (2002, p. 163)
To replicate the values for the two-sided case, we must
decide how the α error should be distributed to the twosides. The method chosen by O’Brien corresponds to Option 2 in G * Power (“Assign to minor tail α/2, then rest
to major tail”, see above). In the main window, we selectTail(s) "Two" and set the other input parameters exactly
as shown in the example above. For sample sizes 50, 75, 100,
125, 150 we get power values 0.798241, 0.930639, 0.980441,
0.994839, and 0.998658, respectively, which are again equal
to the values given in O’Brien’s table.

5.4

Related tests

5.5

Implementation notes

Exact (unconditional) test . In this case G * Power calculates the unconditional power for the exact conditional test:
The number of discordant pairs ND is a random variablewith binomial distribution B( N, π D ), where N denotes thetotal number of pairs, and π D = π12 + π21 denotes theprobability of discordant pairs. Thus P( ND ) = ( NND )(π11 +π22 ) ND (π12 + π21 ) N − ND . Conditional on ND , the frequency
f 12 has a binomial distribution B( ND , π0 = π12 /π D ) and
we test the H0 : π0 = 0.5. Given the conditional ‘binomial power’ Pow( ND , π0 | ND = i ) the exact unconditional
power is ∑iN P( ND = i )Pow( ND , π0 | Nd = i ). The summation starts at the most probable value for ND and then steps
outward until the values are small enough to be ignored.
Fast approximation . In this case an ordinary one
sample binomial power calculation with H0 -distribution
B( Nπ D , 0.5), and H1 -Distribution B( Nπ D , OR/(OR + 1)) is
performed.

5.6

Validation

The results of the exact procedure were checked against
the values given on pages 161-163 in O’Brien (2002). Complete correspondence was found in the one-tailed case and15

Figure 8: Result of the sample McNemar test (see text for details).

16

6

6.1

Exact: Proportions - inequality of
two independent groups (Fisher’s
exact-test)

where
Pr ( T ≥ tα |m, H1 ) =This procedure calculates power and sample size for tests
comparing two independent binomial populations withprobabilities π1 and π2 , respectively. The results of sampling from these two populations can be given in a 2 × 2
contingency table X:
Success
FailureTotal

Group 2x2
n2 − x2
n2

Total
m
N−m
N

T = − ln,

" n n #
( x11 )( x22 )N(m)

2. Persons’s exact test:( x j − mn j /N )2 [(n j − x j ) − ( N − m)n j /N ]2+
mn j /N( N − m)n j /N

2

T=∑

j =1

The choice of the test statistics only influences the way inwhich α is distributed on both sides of the null distribution.
For one-sided tests the test statistic is T = x2 .
6.6.2

Large sample approximation

The large sample approximation is based on a continuity
corrected χ2 test with pooled variances. To permit a twosided test, a z test version is used: The H0 distribution is
the standard normal distribution N (0, 1), and the H1 distribution given by the normal distribution N (m(k), σ ), withq1
σ=p1 (1 − p1 )/n1 + p2 (1 − p2 )/n2
σ0

Options

This test has no options.

6.4

Examples6.5

Related tests

6.6

Implementation notes

6.6.1

Exact unconditional power!

3. Likelihood ratio exact test:
"
#
"
#!
2
xj
nj − xj
T = 2 ∑ x j ln
+ (n j − x j ) ln
mn j /N( N − m)n j /N
j =1

Effect size index

m(k) =s

The procedure computes the exact unconditional power ofthe (conditional) test.
The exact probability of table X (see introduction) underH0 , conditional on x1 + x2 = m, is given by:
Pr ( X |m, H0 ) =B12

1. Fisher’s exact test:

The effect size is determined by directly specifying the two
proportions π1 and π2 .

6.3

X ∈ Mt α ∑ X ∈ M

For two-sided tests G * Power provides three common
test statistics that are asymptotically equivalent:

Here, n1 , n2 are the sample sizes, and x1 , x2 the observed
number of successes in the two populations. N = n1 + n2 is
the total sample size, and m = x1 + x2 the total number ofsuccesses.
The null hypothesis states that π1 = π2 , whereas the alternative hypothesis assumes different probabilities in both
populations:
H0 : π 1 − π 2 = 0
H1 : π1 − π2 ̸= 0.

6.2

B12

P(m) = Pr ( x1 + x2 = m|H1 ) = B12 , and
 
 
n1
x1
n1 − x1 n 2
π1 (1 − π1 )B12 =π2x2 (1 − π2 )n2 − x2x1x2

Introduction

Group 1x1
n1 − x1
n1

∑

σ0 =

(nx11 )(nx22 )
N
(m
)

6.7

Let T be a test statistic, t a possible value of T, and M theset of all tables X with a total number of successes equal
to m. We define Mt = { X ∈ M : T ≥ t}, i.e. Mt is thesubset of M containing all tables for which the value of thetest statistic is equal to or exceeds the value t. The exact
null distribution of T is obtained by calculating Pr ( T ≥t|m, H0 ) = ∑ X ∈ Mt Pr ( X |m, H0 ) for all possible t. The criticalvalue tα is the smallest value such that Pr ( T ≥ tα |m, H0 ) ≤α. The power is then defined as:1[ p2 − p1 − k(1/n1 + 1/n2 )/2], withσ0
n1 (1 − p1 ) + n2 (1 − p2 ) n1 p1 + n2 p2·
n1 n2
n1 + n2
p1 < p2 : −1k=p1 ≥ p2 : +1

Validation

The results were checked against the values produced by
GPower 2.0.

N

1 − β = ∑ P(m) Pr ( T ≥ tα |m, H1 ),
m =0

17

7

Exact test: Multiple Regression - random model

In multiple regression analyses, the relation of a dependent
variable Y to m independent factors X = ( X1 , ..., Xm ) is
studied. The present procedure refers to the so-called unconditional or random factors model of multiple regression(Gatsonis & Sampson, 1989; Sampson, 1974), that is, it is assumed that Y and X1 , . . . , Xm are random variables, where(Y, X1 , . . . , Xm ) have a joint multivariate normal distribution with a positive definite covariance matrix:
 2
′σY ΣYXΣYX Σ Xand mean (µY , µ X ). Without loss of generality we may assume centered variables with µY = 0, µ Xi = 0.
The squared population multiple correlation coefficient
between Y and X is given by:
′
1
2
2
= ΣYX
Σ−
ρYX
X ΣYX /σY .

and the regression coefficient vector γ, the analog of β in1
the fixed model, by γ = Σ−X ΣYX . The maximum likelihood
estimates of the regression coefficient vector and the residuals are the same under both the fixed and the random
model (see theorem 1 in Sampson (1974)); the models differ, however, with respect to power.
The present procedure allows power analyses for the test
2 has
that the population squared correlations coefficient ρYX
the value ρ20 . The null and alternate hypotheses are:
H0 :
H1 :

Figure 9: Effect size drawer to calculate ρ2 from a confidence interval or via predictor correlations (see text).

2 = ρ2ρYX
0
2 ̸ = ρ2 .
ρYX
0value can range from 0 to 1, where 0, 0.5 and 1 corresponds to the left, central and right position inside the interval, respectively. The output fields C.I. lower ρ2 and C.I.
upper ρ2 contain the left and right border of the two-sided
100(1 − α) percent confidence interval for ρ2 . The output
fields Statistical lower bound and Statistical upperbound show the one-sided (0, R) and ( L, 1) intervals, respectively.

An important special case is ρ0 = 0 (corresponding to theassumption ΣYX = 0). A commonly used test statistic for
2 / (1 − R2 ), which has
this case is F = [( N − m − 1)/p] RYX
YXa central F distribution with d f 1 = m, and d f 2 = N − m − 1.
This is the same test statistic that is used in the fixed model.
The power differs, however, in both cases.

7.1

Effect size from predictor correlations By choosing theoption "From predictor correlation matrix" (see Fig. (9)) one
may compute ρ2 from the matrix of correlation among thepredictor variables and the correlations between predictors
and the dependent variable Y. Pressing the "Insert/edit
matrix"-button opens a window, in which one can specify (1) the row vector u containing the correlations between
each of the m predictors Xi and the dependent variable Y
and (2) the m × m matrix B of correlations among the predictors. The squared multiple correlation coefficient is thengiven by ρ2 = uB−1 u′ . Each input correlation must lie in
the interval [−1, 1], the matrix B must be positive-definite,
and the resulting ρ2 must lie in the interval [0, 1]. Pressing
the Button "Calc ρ2 " tries to calculate ρ2 from the input andchecks the positive-definiteness of matrix B.

Effect size index

The effect size is the population squared correlation coefficient H1 ρ2 under the alternative hypothesis. To fully specify the effect size, you also need to give the populationsquared correlation coefficient H0 ρ2 under the null hypothesis.
Pressing the button Determine on the left of the effect sizelabel in the main window opens the effect size drawer (seeFig. 9) that may be used to calculateρ2 either from the con2
fidence interval for the population ρYXgiven an observed
2squared multiple correlation RYX or from predictor correlations.
Effect size from C.I. Figure (9) shows an example of how
the H1 ρ2 can be determined from the confidence interval
computed for an observed R2 . You have to input the sample size, the number of predictors, the observed R2 and theconfidence level of the confidence interval. In the remaining input field a relative position inside the confidence interval can be given that determines the H1 ρ2 value. The

Relation of ρ2 to effect size f 2 The relation between ρ2and effect size f 2 used in the fixed factors model is:
f2 =

18

ρ2
1 − ρ2

Figure 10: Input of correlations between predictors and Y (top) and the matrix of correlations among the predictors (bottom).

and conversely:f2
ρ2 =1 + f2

Examples7.3.1

Power and sample size

Example 1 We replicate an example given for the procedure for the fixed model, but now under the assumptionthat the predictors are not fixed but random samples: We
assume that a dependent variable Y is predicted by as set
2 is 0.10, that is that the 5 preB of 5 predictors and that ρYX
dictors account for 10% of the variance of Y. The sample
size is N = 95 subjects. What is the power of the F test that2 = 0 at α = 0.05?
ρYX
We choose the following settings in G * Power to calculate the power:

Cohen (1988, p. 412) defines the following conventional
values for the effect size f 2 :• small f 2 = 0.02
• medium f 2 = 0.15
• large f 2 = 0.35
which translate into the following values for ρ2 :• small ρ2 = 0.02
• medium ρ2 = 0.13

• Select
Type of power analysis: Post hoc• large ρ2 = 0.26

7.2

7.3

• Input
Tail(s): One
H1 ρ2 : 0.1
α err prob: 0.05
Total sample size: 95
Number of predictors: 5
H0 ρ2 : 0.0

Options

You can switch between an exact procedure for the calculation of the distribution of the squared multiple correlation
coefficient ρ2 and a three-moment F approximation suggested by Lee (1971, p.123). The latter is slightly faster andmay be used to check the results of the exact routine.
19

The results show that N should not be less than 153. Thisconfirms the results in Shieh and Kung (2007).

• Output
Lower critical R2 : 0.115170
Upper critical R2 : 0.115170
Power (1- β): 0.662627

7.3.2

The output shows that the power of this test is about 0.663
which is slightly lower than the power 0.674 found in the
fixed model. This observation holds in general: The powerin the random model is never larger than that found for thesame scenario in the fixed model.

Using confidence intervals to determine the effectsize

Suppose that in a regression analysis with 5 predictors andN = 50 we observed a squared multiple correlation coefficient R2 = 0.3 and we want to use the lower boundaryof the 95% confidence interval for ρ2 as H1 ρ2 . Pressing theDetermine-button next to the effect size field in the main
window opens the effect size drawer. After selecting input
mode "From confidence interval" we insert the above values (50, 5, 0.3, 0.95) in the corresponding input field and
set Rel C.I. pos to use (0=left, 1=right) to 0 to select the left interval border. Pressing calculate computes
the lower, upper, and 95% two-sided confidence intervals:[0, 4245], [0.0589, 1] and [0.0337, 0.4606]. The left boundaryof the two-sided interval (0.0337) is transfered to the field
H1 ρ2 .

Example 2 We now replicate the test of the hypotheses
H0 : ρ2 ≤ 0.3 versus H1 : ρ2 > 0.3 given in Shieh and Kung
(2007, p.733), for N = 100, α = 0.05, and m = 5 predictors.
We assume that H1ρ2 = 0.4 . The settings and output inthis case are:• Select
Type of power analysis: Post hoc• Input
Tail(s): One
H1 ρ2 : 0.4
α err prob: 0.05
Total sample size: 100
Number of predictors: 5
H0 ρ2 : 0.3

7.3.3

Using predictor correlations to determine effectsize

We may use assumptions about the (m × m) correlation matrix between a set of m predictors, and the m correlations
between predictor variables and the dependent variable Y
to determine ρ2 . Pressing the Determine-button next to theeffect size field in the main window opens the effect size
drawer. After selecting input mode "From predictor correlations" we insert the number of predictors in the corresponding field and press "Insert/edit matrix". This opens
a input dialog (see Fig. (10)). Suppose that we have 4 predictors and that the 4 correlations between Xi and Y areu = (0.3, 0.1, −0.2, 0.2). We insert this values in the tab"Corr between predictors and outcome". Assume further
that the correlations between X1 and X3 and between X2
and X4 are 0.5 and 0.2, respectively, whereas all other predictor pairs are uncorrelated. We insert the correlation matrix


1
0 0.5
0
 0
1
0 0.2 

B=
 0.5
0
1
0 
0 0.2
0
1

• Output
Lower critical R2 : 0.456625
Upper critical R2 : 0.456625
Power (1- β): 0.346482
The results show, that H0 should be rejected if the observed
R2 is larger than 0.457. The power of the test is about 0.346.
Assume we observed R2 = 0.5. To calculate the associatedp-value we may use the G * Power -calculator. The syntaxof the CDF of the squared sample multiple correlation coefficient is mr2cdf(R2 ,ρ2 ,m+1,N). Thus for the present case
we insert 1-mr2cdf(0.5,0.3,6,100) in the calculator andpressing Calculate gives 0.01278. These values replicate
those given in Shieh and Kung (2007).
Example 3 We now ask for the minimum sample size required for testing the hypothesis H0 : ρ2 ≥ 0.2 vs. the specific alternative hypothesis H1 : ρ2 = 0.05 with 5 predictors
to achieve power=0.9 and α = 0.05 (Example 2 in Shieh andKung (2007)). The inputs and outputs are:under the "Corr between predictors" tab. Pressing the "Calc
ρ2 "-button computes ρ2 = uB−1 u′ = 0.297083, which also
confirms that B is positive-definite and thus a correct correlation matrix.

• Select
Type of power analysis: A priori

7.4• Input
Tail(s): One
H1 ρ2 : 0.05
α err prob: 0.05
Power (1- β): 0.9
Number of predictors: 5
H0 ρ2 : 0.2

Related tests

Similar tests in G * Power 3.0:
• Linear Multiple Regression: Deviation of R2 from zero.
• Linear Multiple Regression: Increase of R2 .

7.5

• Output
Lower critical R2 : 0.132309
Upper critical R2 : 0.132309
Total sample size: 153
Actual power: 0.901051

Implementation notes

The procedure uses the exact sampling distribution of thesquared multiple correlation coefficient (MRC-distribution)
Lee (1971, 1972). The parameters of this distribution are thepopulation squared multiple correlation coefficient ρ2 , the20

number of predictors m, and the sample size N. The only
difference between the H0 and H1 distribution is that the
population multiple correlation coefficient is set to "H0 ρ2 "in the former and to "H1 ρ2 " in the latter case.
Several algorithms for the computation of the exact
or approximate CDF of the distribution have been proposed (Benton & Krishnamoorthy, 2003; Ding, 1996; Ding &Bargmann, 1991; Lee, 1971, 1972). Benton and Krishnamoorthy (2003) have shown, that the implementation proposed
by Ding and Bargmann (1991) (that is used in Dunlap,
Xin, and Myers (2004)) may produce grossly false resultsin some cases. The implementation of Ding (1996) has the
disadvantage that it overflows for large sample sizes, because factorials occuring in ratios are explicitly evaluated.
This can easily be avoided by using the log of the gammafunction in the computation instead.
In G * Power we use the procedure of Benton and Krishnamoorthy (2003) to compute the exact CDF and a modified version of the procedure given in Ding (1996) to compute the exact PDF of the distribution. Optionally, one canchoose to use the 3-moment noncentral F approximation
proposed by Lee (1971) to compute the CDF. The latter procedure has also been used by Steiger and Fouladi (1992) intheir R2 program, which provides similar functionality.

7.6

Validation

The power and sample size results were checked against
the values produced by R2 (Steiger & Fouladi, 1992), the tables in Gatsonis and Sampson (1989), and results reported
in Dunlap et al. (2004) and Shieh and Kung (2007). Slightdeviations from the values computed with R2 were found,which are due to the approximation used in R2, whereas
complete correspondence was found in all other tests made.
The confidence intervals were checked against values computed in R2, the results reported in Shieh and Kung (2007),and the tables given in Mendoza and Stafford (2001).

7.7

References

See Chapter 9 in Cohen (1988) for a description of the fixedmodel. The random model is described in Gatsonis andSampson (1989) and Sampson (1974).

21

8

Exact: Proportion - sign test

The sign test is equivalent to a test that the probability π of an event in the populations has the value
π0 = 0.5. It is identical to the special case π0 =0.5 of the test Exact: Proportion - difference fromconstant (one sample case). For a more thorough description see the comments for that test.

8.1

Effect size index

The effect size index is g = π − 0.5.
(Cohen, 1969, p. 142) defines the following effect size
conventions:• small g = 0.05
• medium g = 0.15
• large g = 0.25

8.2

Options

See comments for Exact: Proportion - difference
from constant (one sample case) in chapter 4 (page 11).

8.3

Examples8.4

Related tests

Similar tests in G * Power 3.0:
• Exact: Proportion - difference from constant (one sample case).

8.5

Implementation notes

See comments for Exact: Proportion - difference
from constant (one sample case) in chapter 4 (page 11).

8.6

Validation

The results were checked against the tabulated values in Cohen (1969, chap. 5). For more information
see comments for Exact: Proportion - difference fromconstant (one sample case) in chapter 4 (page 11).

22

9Exact: Generic binomial test

9.1

Effect size index

9.2

Options

Since the binomial distribution is discrete, it is normallynot possible to achieve exactly the nominal α-level. For twosided tests this leads to the problem how to “distribute” αon the two sides. G * Power offers three options (case 1 isthe default):
1. Assign α/2 on both sides: Both sides are handled independently in exactly the same way as in a one-sided
test. The only difference is that here α/2 is used instead of α. From the three options this one leads to thegreatest deviation from the actual α (in post hoc analyses).
2. Assign to minor tail α/2, then rest to major tail (α2 =
α/2, α1 = α − α2 ): First α/2 is applied on the side of
the central distribution that is farther away from thenoncentral distribution (minor tail). The criterion usedon the other side is then α − α1 , where α1 is the actual
α found on the minor side. Since α1 ≤ α/2 one canconclude that (in post hoc analyses) the sum of the actual values α1 + α2 is in general closer to the nominal
α-level than in case 1.
3. Assign α/2 on both sides, then increase to minimize the difference of α1 + α2 to α: The first step is exactly the sameas in case 1. Then, in the second step, the critical values on both sides of the distribution are increased (using the lower of the two potential incremental α-values)
until the sum of both actual α’s is as close as possible
to the nominal α.

9.3

Examples9.4

Related tests

9.5

Implementation notes

9.6

Validation

The results were checked against the values produced by
GPower 2.0.

9.7

References

Cohen...

23

10

F test: Fixed effects ANOVA - oneway

The fixed effects one-way ANOVA tests whether there areany differences between the means µi of k ≥ 2 normally
distributed random variables with equal variance σ. The
random variables represent measurements of a variable X
in k fixed populations. The one-way ANOVA can be viewed
as an extension of the two group t test for a difference ofmeans to more than two groups.
The null hypothesis is that all k means are identical H0 :µ1 = µ2 = . . . = µk . The alternative hypothesis states thatat least two of the k means differ. H1 : µi ̸= µ j , for at least
one pair i, j with 1 ≤ i, j ≤ k.

10.1

Effect size index

The effect size f is defined as: f = σm /σ. In this equation σm is the standard deviation of the group means µi
and σ the common standard deviation within each of the
2 + σ2 . A difk groups. The total variance is then σt2 = σmferent but equivalent way to specify the effect size is in
2 /σ2 . That is, η 2
terms of η 2 , which is defined as η 2 = σmt
2 andis the ratio between the between-groups variance σm2
the total variance σt and can be interpreted as “proportionof variance explained by group membership”. The relationship p
between η 2 and f is: η 2 = f 2 /(1 + f 2 ) or solved for f :f = η 2 / (1 − η 2 ).
Cohen (1969, p.348) defines the following effect size conventions:• small f = 0.10
• medium f = 0.25
• large f = 0.40
If the mean µi and size ni of all k groups are known thenthe standard deviation σm can be calculated in the following
way:
µ̄

=

σm

=

k

wi µ i ,∑q i =1(grand mean),k∑i=1 wi (µi − µ̄)2 .

where wi = ni /(n1 + n2 + · · · + nk ) stands for the relative
size of group i.
Pressing the Determine button to the left of the effect sizelabel opens the effect size drawer. You can use this drawer
to calculate the effect size f from variances, from η 2 or fromthe group means and group sizes. The drawer essentiallycontains two different dialogs and you can use the Selectprocedure selection field to choose one of them.
10.1.1Figure 11: Effect size dialogs to calculate f

this button fills the size column of the table with the chosen
value.
Clicking on the Calculate button provides a preview ofthe effect size that results from your inputs. If you clickon the Calculate and transfer to main window button
then G * Power calculates the effect size and transfers theresult into the effect size field in the main window. If thenumber of groups or the total sample size given in the effect size drawer differ from the corresponding values in themain window, you will be asked whether you want to adjust the values in the main window to the ones in the effectsize drawer.

Effect size from means

In this dialog (see left side of Fig. 11) you normally start by
setting the number of groups. G * Power then provides you
with a mean and group size table of appropriate size. Insert
the standard deviation σ common to all groups in the SD
σ within each group field. Then you need to specify themean µi and size ni for each group. If all group sizes areequal then you may insert the common group size in the
input field to the right of the Equal n button. Clicking on
24

10.1.2

10.4

Effect size from varianceThis dialog offers two ways to specify f . If you choose
From Variances then you need to insert the variance of
2 , into the Variance explained
the group means, that is σmby special effect field, and the square of the common
standard deviation within each group, that is σ2 , into
the Variance within groups field. Alternatively, you may
choose the option Direct and then specify the effect size f
via η 2 .

10.2

• ANOVA: Fixed effects, special, main effects and interactions• ANOVA: Repeated measures, between factors

10.5

Implementation notes

The distribution under H0 is the central F (k − 1, N − k ) distribution with numerator d f 1 = k − 1 and denominatord f 2 = N − k. The distribution under H1 is the noncentral
F (k − 1, N − k, λ) distribution with the same df’s and noncentrality parameter λ = f 2 N. (k is the number of groups,
N is the total sample size.)

Options

This test has no options.

10.3

Related tests

Examples10.6

We compare 10 groups, and we have reason to expect a"medium" effect size ( f = .25). How many subjects do weneed in a test with α = 0.05 to achieve a power of 0.95?

Validation

The results were checked against the values produced by
GPower 2.0.

• Select
Type of power analysis: A priori• Input
Effect size f : 0.25
α err prob: 0.05
Power (1-β err prob): 0.95
Number of groups: 10• Output
Noncentrality parameter λ: 24.375000
Critical F: 1.904538
Numerator df: 9
Denominator df: 380
Total sample size: 390
Actual Power: 0.952363
Thus, we need 39 subjects in each of the 10 groups. What
if we had only 200 subjects available? Assuming that both αand β error are equally costly (i.e., the ratio q := beta/alpha= 1) which probably is the default in basic research, we cancompute the following compromise power analysis:• Select
Type of power analysis: Compromise• Input
Effect size f : 0.25
β/α ratio: 1
Total sample size: 200
Number of groups: 10• Output
Noncentrality parameter λ: 12.500000
Critical F: 1.476210
Numerator df: 9
Denominator df: 190
α err prob: 0.159194
β err prob: 0.159194
Power (1-β err prob): 0.840806

25

11

F test: Fixed effects ANOVA - special, main effects and interactions

In testing the three-factor interactions, the residuals δijk
of the group means after subtraction of all main effects andall two-factor interactions are considered. In a three factor
design there is only one possible three-factor interaction.
The 3 × 3 × 4 = 36 residuals in the example are calculated
as δijk = µijk − µi⋆⋆ − µ⋆ j⋆ − µ⋆⋆k − δij⋆ − δi⋆k − δ⋆ jk . The
null hypothesis of no interaction states that all residuals
are equal. Thus,

This procedure may be used to calculate the power of main
effects and interactions in fixed effects ANOVAs with factorial designs. It can also be used to compute power forplanned comparisons. We will discuss both applications inturn.
11.0.1H0 : δijk = δlmn for all combinations of index triples
i, j, k and l, m, n.

Main effects and interactions

H1 : δijk ̸= δlmn for at least one combination of index
triples i, j, k and l, m, n.

To illustrate the concepts underlying tests of main effects
and interactions we will consider the specific example of an
A × B × C factorial design, with i = 3 levels of A, j = 3
levels of B, and k = 4 levels of C. This design has a total
number of 3 × 3 × 4 = 36 groups. A general assumption isthat all groups have the same size and that in each groupthe dependent variable is normally distributed with identical variance.
In a three factor design we may test three main effectsof the factors A, B, C, three two-factor interactions A × B,
A × C, B × C, and one three-factor interaction A × B × C.
We write µijk for the mean of group A = i, B = j, C = k.
To indicate the mean of means across a dimension we writea star (⋆) in the corresponding index. Thus, in the exampleµij⋆ is the mean of the groups A = i, B = j, C = 1, 2, 3, 4.
To simplify the discussion we assume that the grand meanµ⋆⋆⋆ over all groups is zero. This can always be achieved bysubtracting a given non-zero grand mean from each groupmean.
In testing the main effects, the null hypothesis is that
all means of the corresponding factor are identical. For themain effect of factor A the hypotheses are, for instance:

The assumption that the grand mean is zero implies that∑i,j,k δijk = 0. The above hypotheses are therefore equivalenttoH0 : δijk = 0 for all i, j, k
H1 : δijk ̸= 0 for at least one i, j, k.
It should be obvious how the reasoning outlined above
can be generalized to designs with 4 and more factors.
11.0.2

Planned comparisons

Planned comparison are specific tests between levels of afactor planned before the experiment was conducted.
One application is the comparison between two sets oflevels. The general idea is to subtract the means across
two sets of levels that should be compared from each otherand to test whether the difference is zero. Formally this isdone by calculating the sum of the component-wise product of the mean vector ⃗µ and a nonzero contrast vector⃗c (i.e. the scalar product of ⃗µ and c): C = ∑ik=1 ci µi . The
contrast vector c contains negative weights for levels on
one side of the comparison, positive weights for the levels on the other side of the comparison and zero for levels
that are not part of the comparison. The sum of weights
is always zero. Assume, for instance, that we have a factor with 4 levels and mean vector ⃗µ = (2, 3, 1, 2) and that
we want to test whether the means in the first two levels are identical to the means in the last two levels. In
this case we define ⃗c = (−1/2, −1/2, 1/2, 1/2) and get
C = ∑i ⃗µi⃗ci = −1 − 3/2 + 1/2 + 1 = −1.
A second application is testing polygonal contrasts in atrend analysis. In this case it is normally assumed that thefactor represents a quantitative variable and that the levels of the factor that correspond to specific values of thisquantitative variable are equally spaced (for more details,
see e.g. Hays (1988, p. 706ff)). In a factor with k levels k − 1
orthogonal polynomial trends can be tested.
In planned comparisons the null hypothesis is: H0 : C =0, and the alternative hypothesis H1 : C ̸= 0.

H0 : µ1⋆⋆ = µ2⋆⋆ = µ3⋆⋆
H1 : µi⋆⋆ ̸= µ j⋆⋆ for at least one index pair i, j.
The assumption that the grand mean is zero implies that∑i µi⋆⋆ = ∑ j µ⋆ j⋆ = ∑k µ⋆⋆k = 0. The above hypotheses are
therefore equivalent to
H0 : µi⋆⋆ = 0 for all iH1 : µi⋆⋆ ̸= 0 for at least one i.
In testing two-factor interactions, the residuals δij⋆ , δi⋆k ,and δ⋆ik of the groups means after subtraction of the main
effects are considered. For the A × B interaction of the example, the 3 × 3 = 9 relevant residuals are δij⋆ = µij⋆ −µi⋆⋆ − µ⋆ j⋆ . The null hypothesis of no interaction effectstates that all residuals are identical. The hypotheses for
the A × B interaction are, for example:
H0 : δij⋆ = δkl ⋆ for all index pairs i, j and k, l.
H1 : δij⋆ ̸= δkl ⋆ for at least one combination of i, j and
k, l.

11.1

The assumption that the grand mean is zero implies that∑i,j δij⋆ = ∑i,k δi⋆k = ∑ j,k δ⋆ jk = 0. The above hypotheses are
therefore equivalent toEffect size index

The effect size f is defined as: f = σm /σ. In this equation σmis the standard deviation of the effects that we want to testand σ the common standard deviation within each of the
2 +groups in the design. The total variance is then σt2 = σm
2
σ . A different but equivalent way to specify the effect size

H0 : δij⋆ = 0 for all i, j
H1 : δij⋆ ̸= 0 for at least one i, j.
26

2 /σ2 . That is,
is in terms of η 2 , which is defined as η 2 = σm
t
2
2 and
η is the ratio between the between-groups variance σm2
the total variance σt and can be interpreted as “proportionof variance explained by the effect under consideration”.
The relationship between
η 2 and f is: η 2 = f 2 /(1 + f 2 ) orpsolved for f : f = η 2 /(1 − η 2 ).
Cohen (1969, p.348) defines the following effect size conventions:is not essential but makes the calculation and discussion
easier). Next, we estimate the common variance σ withineach group by calculating the mean variance of all cells,
that is, σ2 = 1/36 ∑i s2i = 1.71296.
Main effects To calculate the power for the A, B, and C
main effects we need to know the effect size f = σm /σ.
We already know σ2 to be 1.71296 but need to calculate the2 for each factor. The procedure isvariance of the means σmanalogous for all three main effects. We therefore demonstrate only the calculations necessary for the main effect offactor A.
We first calculate the three means for factor A: µi⋆⋆ ={−0.722231, 1.30556, −0.583331}. Due to the fact that wehave first subtracted the grand mean from each cell wehave ∑i µi⋆⋆ = 0, and we can easily compute the vari2 = 1/3
ance of these means as mean square: σmµ2 =√∑i i⋆⋆
σm /σ =
0.85546.
With these values we calculate f =√0.85546/1.71296 = 0.7066856. The effect size drawer
in G * Power can be used to do the last calculation: We
choose From Variances and insert 0.85546 in the Varianceexplained by special effect and 1.71296 in the Error
Variance field. Pressing the Calculate button gives theabove value for f and a partial η 2 of 0.3330686. Notethat the partial η 2 given by G * Power is calculated fromf according to the formula η 2 = f 2 /(1 + f 2 ) and is not
identical to the SPSS partial η 2 , which is based on sample estimates. The relation between the two is “SPSS η02 ”= η 2 N/( N + k(η 2 − 1)), where N denotes the total sample
size, k the total number of groups in the design and η 2 theG * Power value. Thus η02 = 0.33306806 · 108/(108 − 36 +0.33306806 · 36) = 0.42828, which is the value given in the
SPSS output.
We now use G * Power to calculate the power for α = 0.05
and a total sample size 3 × 3 × 4 × 3 = 108. We set• small f = 0.10
• medium f = 0.25
• large f = 0.40

Figure 12: Effect size dialog to calculate f

Clicking on the Determine button to the left of the effectsize label opens the effect size drawer (see Fig. 12). You can
use this drawer to calculate the effect size f from variances
or from η 2 . If you choose From Variances then you need toinsert the variance explained by the effect under considera2 , into the Variance explained by specialtion, that is σmeffect field, and the square of the common standard deviation within each group, that is σ2 , into the Variancewithin groups field. Alternatively, you may choose the option Direct and then specify the effect size f via η 2 .
See examples section below for information on how to
calculate the effect size f in tests of main effects and interactions and tests of planned comparisons.

11.2• Select
Type of power analysis: Post hoc• Input
Effect size f : 0.7066856
α err prob: 0.05
Total sample size: 108
Numerator df: 2(number of factor levels - 1, A has
3 levels)
Number of groups: 36(total number of groups inthe design)

Options

This test has no options.

11.3

Examples11.3.1

Effect sizes from means and standard deviations• Output
Noncentrality parameter λ: 53.935690
Critical F: 3.123907
Denominator df: 72
Power (1-β err prob): 0.99999

To illustrate the test of main effects and interaction we assume the specific values for our A × B × C example shownin Table 13. Table 14 shows the results of a SPSS analysis(GLM univariate) done for these data. In the following wewill show how to reproduce the values in the Observed
Power column in the SPSS output with G * Power .
As a first step we calculate the grand mean of the data.
Since all groups have the same size (n = 3) this is justthe arithmetic mean of all 36 groups means: m g = 3.1382.
We then subtract this grand mean from all cells (this step

The value of the noncentrality parameter and the power
computed by G * Power are identical to the values in the
SPSS output.
Two-factor interactions To calculate the power for twofactor interactions A × B, A × C, and A × B we need to
calculate the effect size f corresponding to the values given
in table 13. The procedure is analogous for each of the three27

Figure 13: Hypothetical means (m) and standard deviations (s) of a 3 × 3 × 4 design.

Figure 14: Results computed with SPSS for the values given in table 13

Number of groups: 36
the design)

two-factor interactions and we thus restrict ourselves to the
A × B interaction.
2 are the 3 × 3 = 9
The values needed to calculate σmresiduals δij⋆ . They are given by δij⋆ = µij⋆ − µi⋆⋆ −µ⋆ j⋆ = {0.555564, -0.361111, -0.194453, -0.388903, 0.444447,-0.0555444, -0.166661, -0.0833361, 0.249997}. The mean ofthese values is zero (as a consequence of subtracting
the grand mean). Thus, the variance σm is given by
1/9 ∑i,j δij2 ⋆ = 0.102881. This results in an effect size√
f = 0.102881/1.71296 = 0.2450722 and a partial η 2 =0.0557195. Using the formula given in the previous section
on main effects it can be checked that this corresponds to a
“SPSS η02 ” of 0.0813, which is identical to that given in the
SPSS output.
We use G * Power to calculate the power for α = 0.05 anda total sample size 3 × 3 × 4 × 3 = 108. We set:

(total number of groups in• Output
Noncentrality parameter λ: 6.486521
Critical F: 2.498919
Denominator df: 72
Power (1-β err prob): 0.475635(The notation #A in the comment above means number oflevels in factor A). A check reveals that the value of the noncentrality parameter and the power computed by G * Power
are identical to the values for (A * B) in the SPSS output.
Three-factor interations To calculate the effect size of thethree-factor interaction corresponding to the values given
in table 13 we need the variance of the 36 residuals δijk =µijk − µi⋆⋆ − µ⋆ j⋆ − µ⋆⋆k − δij⋆ − δi⋆ j − δ⋆ jk = {0.333336,
0.777792, -0.555564, -0.555564, -0.416656, -0.305567,
0.361111, 0.361111, 0.0833194, -0.472225, 0.194453, 0.194453,
0.166669, -0.944475, 0.388903, 0.388903, 0.666653, 0.222242,-0.444447, -0.444447, -0.833322, 0.722233, 0.0555444,
0.0555444, -0.500006, 0.166683, 0.166661, 0.166661, -0.249997,
0.083325, 0.0833361, 0.0833361, 0.750003, -0.250008, -• Select
Type of power analysis: Post hoc• Input
Effect size f : 0.2450722
α err prob: 0.05
Total sample size: 108
Numerator df: 4(#A-1)(#B-1) = (3-1)(3-1)
28

0.249997, -0.249997}. The mean of these values is zero(as a consequence of subtracting the grand mean). Thus,
2 = 0.185189. This
the variance σm is given by 1/36 ∑i,j,k δijk√
results in an effect size f = 0.185189/1.71296 = 0.3288016
and a partial η 2 = 0.09756294. Using the formula given in
the previous section on main effects it can be checked that
this corresponds to a “SPSS η 2 ” of 0.140, which is identicalto that given in the SPSS output.
We use G * Power to calculate the power for α = 0.05 anda total sample size 3 × 3 × 4 × 3 = 108. We therefore choose• Output
Noncentrality parameter λ: 22.830000
Critical F: 1.942507
Denominator df: 2253
Total sample size: 2283
Actual power: 0.950078
G * Power calculates a total sample size of 2283. Please
note that this sample size is not a multiple of the group size
30 (2283/30 = 76.1)! If you want to ensure that your haveequal group sizes, round this value up to a multiple of 30
by choosing a total sample size of 30*77=2310. A post hoc
analysis with this sample size reveals that this increases thepower to 0.952674.

• Select
Type of power analysis: Post hoc• Input
Effect size f : 0.3288016
α err prob: 0.05
Total sample size: 108
Numerator df: 12(#A-1)(#B-1)(#C-1) = (3-1)(3-1)(41)
Number of groups: 36(total number of groups inthe design)

11.3.3

To calculate the effect size f = σm /σ for a given comparison C = ∑ik=1 µi ci we need to know—besides the standard
deviation σ within each group—the standard deviation σmof the effect. It is given by:• Output
Noncentrality parameter λ: 11.675933
Critical F: 1.889242
Denominator df: 72
Power (1-β err prob): 0.513442|C |σm = sk

N ∑ c2i /ni
i =1

where N, ni denote total sample size and sample size ingroup i, respectively.
Given the mean vector µ = (1.5, 2, 3, 4), sample size ni =5 in each group, and standard deviation σ = 2 within eachgroup, we want to calculate the power for the following
contrasts:(The notation #A in the comment above means number oflevels in factor A). Again a check reveals that the value ofthe noncentrality parameter and the power computed by
G * Power are identical to the values for (A * B * C) in the
SPSS output.
11.3.2

Power for planned comparisonscontrast
1,2 vs. 3,4
lin. trend
quad. trend

Using conventional effect sizes

In the example given in the previous section, we assumed
that we know the true values of the mean and variancesin all groups. We are, however, seldom in that position. Instead, we usually only have rough estimates of the expected
effect sizes. In these cases we may resort to the conventional
effect sizes proposed by Cohen.
Assume that we want to calculate the total sample size
needed to achieve a power of 0.95 in testing the A × C twofactor interaction at α level 0.05. Assume further that the
total design in this scenario is A × B × C with 3 × 2 × 5
factor levels, that is, 30 groups. Theoretical considerations
suggest that there should be a small interaction. We thus
use the conventional value f = 0.1 defined by Cohen (1969)
as small effect. The inputs into and outputs of G * Powerfor this scenario are:weights c
1
- 21
2
-3 -1
1
1
-1 -1

- 12

1
2

3
1

σµ
0.875
0.950
0.125

f
0.438
0.475
0.063

η2
0.161
0.184
0.004

Each contrast has a numerator d f = 1. The denominator
dfs are N − k, where k is the number of levels (4 in theexample).
To calculate the power of the linear trend at α = 0.05 we
specify:• Select
Type of power analysis: A priori• Input
Effect size f : 0.475164
α err prob: 0.05
Total sample size: 20
Numerator df: 1
Number of groups: 4• Select
Type of power analysis: A priori• Output
Noncentrality parameter λ: 4.515617
Critical F: 4.493998
Denominator df: 16
Power (1-β err prob): 0.514736• Input
Effect size f : 0.1
α err prob: 0.05
Power (1-β err prob): 0.95
Numerator df: 8(#A-1)(#C-1) = (3-1)(5-1)
Number of groups: 30(total number of groups inthe design)

Inserting the f ’s for the other two contrasts yields a
power of 0.451898 for the comparison of “1,2 vs. 3,4”, and apower of 0.057970 for the test of a quadratic trend.
29

11.4

Related tests• ANOVA: One-way

11.5

Implementation notes

The distribution under H0 is the central F (d f 1 , N − k ) distribution. The numerator d f 1 is specified in the input and thedenominator df is d f 2 = N − k, where N is the total sample
size and k the total number of groups in the design. The
distribution under H1 is the noncentral F (d f 1 , N − k, λ) distribution with the same df’s and noncentrality parameter
λ = f 2 N.

11.6

Validation

The results were checked against the values produced by
GPower 2.0.

30

12

t test: Linear Regression (size of 12.2 Options
slope, one group)This test has no options.

A linear regression is used to estimate the parameters a, bof a linear relationship Y = a + bX between the dependent
variable Y and the independent variable X. X is assumed to
be a set of fixed values, whereas Yi is modeled as a randomvariable: Yi = a + bXi + ε i , where ε i denotes normally distributed random errors with mean 0 and standard deviation
σi . A common assumption also adopted here is that all σi ’s
are identical, that is σi = σ. The standard deviation of theerror is also called the standard deviation of the residuals.
A common task in linear regression analysis is to test
whether the slope b is identical to a fixed value b0 or not.
The null and the two-sided alternative hypotheses are:
H0 :
H1 :

12.1

12.3

We replicate an example given on page 593 in Dupont andPlummer (1998). The question investigated in this example
is, whether the actual average time spent per day exercising
is related to the body mass index (BMI) after 6 month ona training program. The estimated standard deviation of
exercise time of participants is σx = 7.5 minutes. From aprevious study the standard deviation of the BMI in thegroup of participants is estimated to be σy = 4. The sample
size is N = 100 and α = 0.05. We want to determine thepower with which a slope b0 = −0.0667 of the regression
line (corresponding to a drop of BMI by 2 per 30 min/day
exercise) can be detected.

b − b0 = 0b − b0 ̸= 0.

• Select
Type of power analysis: Post hoc

Effect size index• Input
Tail(s): Two
Effect size Slope H1: -0.0667
α err prob: 0.05
Total sample size: 100
Slope H0: 0
Std dev σ_x: 7.5
Std dev σ_y: 4

Slope H1, the slope b of the linear relationship assumed
under H1 is used as effect size measure. To fully specify theeffect size, the following additional inputs must be given:• Slope H0This is the slope b0 assumed under H0.
• Std dev σ_xTheq standard deviation σx of the values in X: σx =n1
2
N ∑i =1 ( Xi − X̄ ) . The standard deviation must be >0.

• Output
Noncentrality parameter δ: 1.260522
Critical t:-1.984467
Df: 98
Power (1- β): 0.238969• Std dev σ_y
The standard deviation σy > 0 of the Y-values. Important relationships of σy to other relevant measures are:σyσy

= (bσx )/ρ
q= σ/ 1 − ρ2

Examples(1)

The output shows that the power of this test is about 0.24.
This confirms the value estimated by Dupont and Plummer(1998, p. 596) for this example.

(2)

12.3.1

Relation to Multiple Regression: Omnibus

The present procedure is a special case of Multiple regression, or better: a different interface to the same procedureusing more convenient variables. To show this, we demonstrate how the MRC procedure can be used to compute theexample above. First, we determine R2 = ρ2 from the relation b = ρ · σy /σx , which implies ρ2 = (b · σx /σy )2 . Entering (-0.0667*7.5/4)^2 into the G * Power calculator givesρ2 = 0.01564062. We enter this value in the effect size dialog of the MRC procedure and compute an effect size
f 2 = 0.0158891. Selecting a post hoc analysis and setting
α err prob to 0.05, Total sample size to 100 and Numberof predictors to 1, we get exactly the same power as given
above.

where σ denotes the standard deviation of the residuals Yi − ( aX + b) and ρ the correlation coefficient between X and Y.
The effect size dialog may be used to determine Std dev
σ_y and/or Slope H1 from other values based on Eqns (1)and (2) given above.
Pressing the button Determine on the left side of the effect size label in the main window opens the effect size
drawer (see Fig. 15).
The right panel in Fig 15 shows the combinations of input and output values in different input modes. The input
variables stand on the left side of the arrow ’=>’, the output
variables on the right side. The input values must conformto the usual restrictions, that is, σ > 0, σx > 0, σy > 0, −1 <
ρ < 1. In addition, Eqn. (1) together with the restriction onρ implies the additional restriction −1 < b · σx /σy < 1.
Clicking on the button Calculate and transfer tomain window copies the values given in Slope H1, Std dev
σ_y and Std dev σ_x to the corresponding input fields inthe main window.

12.4

Related tests

Similar tests in G * Power 3.0:
• Multiple Regression: Omnibus (R2 deviation fromzero).
• Correlation: Point biserial model
31

Figure 15: Effect size drawer to calculate σy and/or slope b from various inputs constellations (see right panel).

12.5

Implementation notes

The H0-distribution is the central t distribution with d f 2 =N − 2 degrees of freedom, where N is the sample size.
The H1-distribution is the noncentral t distribution withthe same
degrees of freedom and noncentrality parameter√δ = N [σx (b − b0 )]/σy .

12.6

Validation

The results were checked against the values produced by by
PASS (Hintze, 2006) and perfect correspondance was found.

32

13

F test: Multiple Regression - omnibus (deviation of R2 from zero),
fixed model

In multiple regression analyses the relation of a dependent
variable Y to p independent factors X1 , ..., Xm is studied.
The present procedure refers to the so-called conditionalor fixed factors model of multiple regression (Gatsonis &Sampson, 1989; Sampson, 1974), that is, it is assumed thatY = Xβ + εwhere X = (1X1 X2 · · · Xm ) is a N × (m + 1) matrix of aconstant term and fixed and known predictor variables Xi .
The elements of the column vector β of length m + 1 are
the regression weights, and the column vector ε of length N
contains error terms, with ε i ∼ N (0, σ ).
This procedure allows power analyses for the test that the
proportion of variance of a dependent variable Y explained
2 , is zero. The null andby a set of predictors B, that is RY·B
alternate hypotheses are:
H0 :
H1 :

Figure 16: Effect size drawer to calculate f 2 from either R2 or frompredictor correlations.

Effect size from predictor correlations By choosing theoption "From predictor correlations" (see Fig. (17)) one may
compute ρ2 from the matrix of correlations among the predictor variables and the correlations between predictors andthe dependent variable Y. Pressing the "Insert/edit matrix"button opens a window, in which one can specify (a) therow vector u containing the correlations between each of
the m predictors Xi and the dependent variable Y, and (b)the m × m matrix B of correlations among the predictors.
The squared multiple correlation coefficient is then givenby ρ2 = uB−1 u′ . Each input correlation must lie in the interval [−1, 1], the matrix B must be positive-definite, and theresulting ρ2 must lie in the interval [0, 1]. Pressing the Button "Calc ρ2 " tries to calculate ρ2 from the input and checks
the positive-definiteness of matrix B and the restriction on
ρ2 .

2
RY
·B = 02
RY· B > 0.

As will be shown in the examples section, the MRC procedure is quite flexible and can be used as a substitute forsome other tests.

13.1

Effect size index

The general definition of the effect size index f 2 used inthis procedure is: f 2 = VS /VE , where VS is the proportion of variance explained by a set of predictors, and VE theresidual or error variance (VE + VS = 1). In the special caseconsidered here (case 0 in Cohen (1988, p. 407ff.)) the pro2
portion of variance explained is given by VS = RY
· B and the2residual variance by VE = 1 − RY · B . Thus:

13.2

2
RY
·B
f =
2
1 − RY·B
2

Options

This test has no options.

and conversely:
f2
2
RY
·B =
1 + f2

13.3

Examples13.3.1

Basic example

We assume that a dependent variable Y is predicted by as2
set B of 5 predictors and that the population RY· B is 0.10,
that is that the 5 predictors account for 10% of the varianceof Y. The sample size is N = 95 subjects. What is the powerof the F test at α = 0.05?
First, by inserting R2 = 0.10 in the effect size dialog wecalculate the corresponding effect size f 2 = 0.1111111. We
then use the following settings in G * Power to calculate thepower:

Cohen (1988, p. 412) defines the following conventional
values for the effect size f 2 :• small f 2 = 0.02
• medium f 2 = 0.15
• large f 2 = 0.35
Pressing the button Determine on the left side of the effect size label in the main window opens the effect size
drawer (see Fig. 16).

• Select
Type of power analysis: Post hoc• Input
Effect size f 2 : 0.1111111
α err prob: 0.05
Total sample size: 95

Effect size from squared multiple correlation coefficient
Choosing input mode "From correlation coefficient" allows
to calculate effect size f 2 from the squared multiple corre2 .
lation coefficient RY·B33

Figure 17: Input of correlations between predictors and Y (top) and the matrix of the correlations among predictors (see text).

Number of predictors: 5

An equivalent analysis could be done using the MRC
procedure. To this end we set the effect size of the MRC
procedure to f 2 = 0.593090442 = 0.351756 and the numberof predictors to (number of groups -1) — in the example tok − 1 = 3. Choosing the remaining parameters α and total
sample size exactly as in the one-way ANOVA case leads toidentical result.
From the fact that the two-sided t-tests for the difference
in means of two independent groups is a special case of
the one-way ANOVA, it can be concluded that this test canalso be regarded as a special case of the MRC procedure.
The relation between the effect size d of the t-test and f 2 isas follows: f 2 = (d/2)2 .

• Output
Noncentrality parameter λ: 10.555555
Critical F: 2.316858
Numerator df: 5
Denominator df: 89
Power (1- β): 0.673586
The output shows that the power of this test is about 0.67.
This confirms the value estimated by Cohen (1988, p. 424)
in his example 9.1, which uses identical values.
13.3.2

Example showing relations to a one-way ANOVA
and the two-sample t-test

13.3.3

We assume the means 2, 3, 2, 5 for the k = 4 experimental groups in a one-factor design. The sample sizes in thegroup are 5, 6, 6, 5, respectively, and the common standard
deviation is assumed to be σ = 2. Using the effect size dialog of the one-way ANOVA procedure we calculate fromthese values the effect size f = 0.5930904. With α = 0.05, 4
groups, and a total sample size of 22, a power of 0.536011
is computed.

Example showing the relation to two-sided testsof point-biserial correlations

For testing whether a point biserial correlation r is different from zero, using the special procedure provided in
G * Power is recommended. But the power analysis of thetwo-sided test can also be done as well with the current
MRC procedure. We just need to set R2 = r2 and Numberof predictor = 1.
Given the correlation r = 0.5 (r2 = 0.25) we get f 2 =34

0.25/(1 − 0.25) = 0.333. For α = 0.05 and total sample size
N = 12 a power of 0.439627 is computed from both procedures.

13.4

Related tests

Similar tests in G * Power 3.0:
• Multiple Regression: Special (R2 increase).
• ANOVA: Fixed effects, omnibus, one-way.
• Means: Difference between two independent means(two groups)• Correlation: Point biserial model

13.5

Implementation notes

The H0-distribution is the central F distribution with numerator degrees of freedom d f 1 = m, and denominator degrees of freedom d f 2 = N − m − 1, where N is the sample size and p the number of predictors in the set B ex2 . The H1plaining the proportion of variance given by RY·B
distribution is the noncentral F distribution with the same
degrees of freedom and noncentrality parameter λ = f 2 N.

13.6

Validation

The results were checked against the values produced by
GPower 2.0 and those produced by PASS (Hintze, 2006).
Slight deviations were found to the values tabulated in Cohen (1988). This is due to an approximation used by Cohen(1988) that underestimates the noncentrality parameter λand therefore also the power. This issue is discussed morethoroughly in Erdfelder, Faul, and Buchner (1996).

13.7

References

See Chapter 9 in Cohen (1988) .

35

14

F test: Multiple Regression - special(increase of R2 ), fixed model

Using this definition, f can alternatively be written interms of the partial R2 :f2 =In multiple regression analyses the relation of a dependent
variable Y to m independent factors X1 , ..., Xm is studied.
The present procedure refers to the so-called conditionalor fixed factors model of multiple regression (Gatsonis &Sampson, 1989; Sampson, 1974), that is, it is assumed that

2. In a second special case (case 2 in Cohen (1988, p.
407ff.)), the same effect variance VS is considered, but it
is assumed that there is a third set of predictors C that
also accounts for parts of the variance of Y and thus
2
reduces the error variance: VE = 1 − RY
· A,B,C . In thiscase, the effect size is

Y = Xβ + εwhere X = (1X1 X2 · · · Xm ) is a N × (m + 1) matrix of aconstant term and fixed and known predictor variables Xi .
The elements of the column vector β of length m + 1 are
the regression weights, and the column vector ε of length N
contains error terms, with ε i ∼ N (0, σ ).
This procedure allows power analyses for the test,
whether the proportion of variance of variable Y explainedby a set of predictors A is increased if an additional
nonempty predictor set B is considered. The variance explained by predictor sets A, B, and A ∪ B is denoted by
2 , R2 , and R2
RY
·A
Y·B
Y · A,B , respectively.
Using this notation, the null and alternate hypotheses
are:
H0 :
H1 :

f2 =

2
2
RY
· A,B − RY · A
2
1 − RY· A,B,CWe may again define a partial R2x as:
R2x :=

2
RYB
·A
2
2
1 − ( RY
· A,B,C − RYB· A )and with this quantity we get
f2 =

2
2
RY
· A,B − RY · A = 0
2
2
RY · A,B − RY
· A > 0.

R2x
1 − R2x

Note: Case 1 is the special case of case 2, where C isthe empty set.

2
The directional form of H1 is due to the fact that RY· A,B ,
that is the proportion of variance explained by sets A and2
B combined, cannot be lower than the proportion RY· A explained by A alone.
As will be shown in the examples section, the MRC procedure is quite flexible and can be used as a substitute forsome other tests.

14.1

2
RYB
·A
2
1 − RYB·A

Pressing the button Determine on the left side of the effect size label in the main window opens the effect size
drawer (see Fig. 18) that may be used to calculate f 2 fromthe variances VS and VE , or alternatively from the partial
R2 .

Effect size index

The general definition of the effect size index f 2 used inthis procedure is: f 2 = VS /VE , where VS is the proportionof variance explained by a set of predictors, and VE is theresidual or error variance.
1. In the first special cases considered here (case 1 in
Cohen (1988, p. 407ff.)), the proportion of varianceexplained by the additional predictor set B is given
22by VS = RY
· A,B − RY · A and the residual variance by
2
VE = 1 − RY
· A,B . Thus:2

f =Figure 18: Effect size drawer to calculate f 2 from variances or fromthe partial R2 .

22RY
· A,B − RY · A
2
1 − RY· A,B(Cohen, 1988, p. 412) defines the following conventional
values for the effect size f 2 :2
2
The quantity RY· A,B − RY,A is also called the squared
semipartial multiple correlation and symbolized by
2
RY·( B· A) . A further interesting quantity is the squared
partial multiple correlation coefficient, which is defined as• small f 2 = 0.02
• medium f 2 = 0.15
• large f 2 = 0.35

2
2
2
RY
RY
·( B· A)
· A,B − RY · A
2
RYB
:
=
=
·A
2
2
1 − RY · A
1 − RY·A

14.2

Options

This test has no options.
36

14.3

Examples14.3.1

Basic example for case 1

2
to 0.16, thus RY· A,B = 0.16. Considering in addition the 4predictors in set C, increases the explained variance further2
to 0.2, thus RY· A,B,C = 0.2. We want to calculate the powerof a test for the increase in variance explained by the inclusion of B in addition to A, given α = 0.01 and a total
sample size of 200. This is a case 2 scenario, because the hypothesis only involves sets A and B, whereas set C should
be included in the calculation of the residual variance.
We use the option From variances in the effectsize drawer to calculate the effect size. In the input
field Variance explained by special effect we insert
22
RY· A,B − RY · A = 0.16 − 0.1 = 0.06, and as Residual
2
variance we insert 1 − RY
· A,B,C = 1 − 0.2 = 0.8. Clicking
on Calculate and transfer to main window shows that
this corresponds to a partial R2 = 0.06976744 and to an effect size f = 0.075. We then set the input field Numerator
df in the main window to 3, the number of predictors in set
B (which are responsible to a potential increase in varianceexplained), and Number of predictors to the total numberof predictors in A, B and C (which all influence the residual
variance), that is to 5 + 3 + 4 = 12.
This leads to the following analysis in G * Power :We make the following assumptions: A dependent variable
Y is predicted by two sets of predictors A and B. The 5
predictors in A alone account for 25% of the variation of
2
Y, thus RY· A = 0.25. Including the 4 predictors in set B
increases the proportion of variance explained to 0.3, thus
2
RY
· A,B = 0.3. We want to calculate the power of a test for
the increase due to the inclusion of B, given α = 0.01 and a
total sample size of 90.
First we use the option From variances in the effect size drawer to calculate the effect size. In the input field Variance explained by special effect we in2
2
sert RY· A,B − RY · A = 0.3 − 0.25 = 0.05, and as Residual
2
variance we insert 1 − RY
· A,B = 1 − 0.3 = 0.7. After clicking on Calculate and transfer to main window we seethat this corresponds to a partial R2 of about 0.0666 andto an effect size f = 0.07142857. We then set the input
field Numerator df in the main window to 4, the number ofpredictors in set B, and Number of predictors to the total
number of predictors in A and B, that is to 4 + 5 = 9.
This leads to the following analysis in G * Power :• Select
Type of power analysis: Post hoc• Select
Type of power analysis: Post hoc• Input
Effect size f 2 : 0.075
α err prob: 0.01
Total sample size: 200
Number of tested predictors: 3
Total number of predictors: 12• Input
Effect size f 2 : 0.0714286
α err prob: 0.01
Total sample size: 90
Number of tested predictors: 4
Total number of predictors: 9• Output
Noncentrality parameter λ:15.000000
Critical F: 3.888052
Numerator df: 3
Denominator df: 187
Power (1- β): 0.766990• Output
Noncentrality parameter λ: 6.428574
Critical F: 3.563110
Numerator df: 4
Denominator df: 80
Power (1- β): 0.241297

We find that the power of this test is about 0.767. In this casethe power is slightly larger than the power value 0.74 estimated by Cohen (1988, p. 439) in his example 9.13, whichuses identical values. This is due to the fact that his approximation for λ = 14.3 underestimates the true value λ = 15
given in the output above.

We find that the power of this test is very low, namely
about 0.24. This confirms the result estimated by Cohen(1988, p. 434) in his example 9.10, which uses identical values. It should be noted, however, that Cohen (1988) uses anapproximation to the correct formula for the noncentrality
parameter λ that in general underestimates the true λ andthus also the true power. In this particular case, Cohen estimates λ = 6.1, which is only slightly lower than the correct
value 6.429 given in the output above.
By using an a priori analysis, we can compute how large
the sample size must be to achieve a power of 0.80. We findthat the required sample size is N = 242.
14.3.214.3.3

Example showing the relation to factorial ANOVA
designs

We assume a 2 × 3 × 4 design with three factors U, V,
W. We want to test main effects, two-way interactions(U × V, U × W, V × W) and the three-way interaction (U ×V × W). We may use the procedure “ANOVA: Fixed effects, special, main effects and interactions” in G * Powerto do this analysis (see the corresponding entry in the manual for details). As an example, we consider the test of the
V × W interaction. Assuming that Variance explained by
the interaction = 0.422 and Error variance = 6.75, leads
to an effect size f = 0.25 (a mean effect size according toCohen (1988)). Numerator df = 6 corresponds to (levels of
V - 1)(levels of W - 1), and the Number of Groups = 24 is
the total number of cells (2 · 3 · 4 = 24) in the design. With

Basis example for case 2

Here we make the following assumptions: A dependent
variable Y is predicted by three sets of predictors A, B
and C, which stand in the following causal relationshipA ⇒ B ⇒ C. The 5 predictors in A alone account for 10% of
2the variation of Y, thus RY· A = 0.10. Including the 3 predictors in set B increases the proportion of variance explained

37

14.6

α = 0.05 and total sample size 120 we compute a power of
0.470.
We now demonstrate, how these analyses can be donewith the MRC procedure. A main factor with k levels corresponds to k − 1 predictors in the MRC analysis, thus
the number of predictors is 1, 2, and 3 for the three factors U, V, and W. The number of predictors in interactions
is the product of the number of predictors involved. The
V × W interaction, for instance, corresponds to a set of(3 − 1)(4 − 1) = (2)(3) = 6 predictors.
To test an effect with MRC we need to isolate the relative
contribution of this source to the total variance, that is weneed to determine VS . We illustrate this for the V × W inter2
action. In this case we must find RY·V ×W by excluding fromthe proportion of variance that is explained by V, W, V × W
2together, that is RY·V,W,V ×W , the contribution of main ef222
fects, that is: RY·V ×W = RY ·V,W,V ×W − RY ·V,W . The residual
variance VE is the variance of Y from which the variance ofall sources in the design have been removed.
This is a case 2 scenario, in which V × W corresponds to
set B with 2 · 3 = 6 predictors, V ∪ W corresponds to set A
with 2 + 3 = 5 predictors, and all other sources of variance,that is U, U × V, U × W, U × V × W, correspond to set C
with (1 + (1 · 2) + (1 · 3) + (1 · 2 · 3)) = 1 + 2 + 3 + 6 = 12
predictors. Thus, the total number of predictors is (6 + 5 +12) = 23. Note: The total number of predictors is always(number of cells in the design -1).
We now specify these contributions numerically:
2
RY · A,B,C = 0.325, RY · A,B − RY
= RY2 ·V ×W =Ȧ
0.0422. Inserting these values in the effect size dialog (Variance explained by special effect = 0.0422,
Residual variance = 1-0.325 = 0.675) yields an effect size
f 2 = 0.06251852 (compare these values with those chosen
in the ANOVA analysis and note that f 2 = 0.252 = 0.0625).
To calculate power, we set Number of tested predictors= 6 (= number of predictors in set B), and Total numberof predictors = 23. With α = 0.05 and N = 120, we get—as expected—the same power 0.470 as from the equivalent
ANOVA analysis.

14.4

The results were checked against the values produced by
GPower 2.0 and those produced by PASS (Hintze, 2006).
Slight deviations were found to the values tabulated in Cohen (1988). This is due to an approximation used by Cohen(1988) that underestimates the noncentrality parameter λand therefore also the power. This issue is discussed morethoroughly in Erdfelder et al. (1996).

14.7

References

See Chapter 9 in Cohen (1988) .

Related tests

Similar tests in G * Power 3.0:
• Multiple Regression: Omnibus (R2 deviation fromzero).
• ANOVA: Fixed effects, special, main effect and interactions.

14.5

Validation

Implementation notes

The H0-distribution is the central F distribution with numerator degrees of freedom d f 1 = q, and denominator degrees of freedom d f 2 = N − m − 1, where N is the sample
size, q the number of tested predictors in set B, which is responsible for a potential increase in explained variance, andm is the total number of predictors. In case 1 (see section“effect size index”), m = q + w, in case 2, m = q + w + v,
where w is the number of predictors in set A and v thenumber of predictors in set C. The H1-distribution is thenoncentral F distribution with the same degrees of freedom
and noncentrality parameter λ = f 2 N.
38

15

F test: Inequality of two Variances

Power (1- β): 0.80
Allocation ratio N2/N1: 1

This procedure allows power analyses for the test thatthe population variances σ02 and σ12 of two normally distributed random variables are identical. The null and (twosided)alternate hypothesis of this test are:
H0 :
H1 :

• Output
Lower critical F: 0.752964
Upper critical F: 1.328085
Numerator df: 192
Denominator df: 192
Sample size group 1: 193
Sample size group 2: 193
Actual power : 0.800105

σ1 − σ0 = 0
σ1 − σ0 ̸= 0.

The two-sided test (“two tails”) should be used if there is
no a priori restriction on the sign of the deviation assumed
in the alternate hypothesis. Otherwise use the one-sided
test (“one tail”).

15.1

The output shows that we need at least 386 subjects (193
in each group) in order to achieve the desired level of the
α and β error. To apply the test, we would estimate bothvariances s21 and s20 from samples of size N1 and N0 , respectively. The two-sided test would be significant at α = 0.05
if the statistic x = s21 /s20 were either smaller than the lower
critical value 0.753 or greater then the upper critical value
1.328.
By setting “Allocation ratio N2/N1 = 2”, we can easily
check that a much larger total sample size, namely N =443 (148 and 295 in group 1 and 2, respectively), would
be required if the sample sizes in both groups are clearly
different.

Effect size index

The ratio σ12 /σ02 of the two variances is used as effect size
measure. This ratio is 1 if H0 is true, that is, if both variances are identical. In an a priori analysis a ratio close
or even identical to 1 would imply an exceedingly large
sample size. Thus, G * Power prohibits inputs in the range[0.999, 1.001] in this case.
Pressing the button Determine on the left side of the effect size label in the main window opens the effect size
drawer (see Fig. 19) that may be used to calculate the ratio
from two variances. Insert the variances σ02 and σ12 in thecorresponding input fields.

15.4

Related tests

Similar tests in G * Power 3.0:
• Variance: Difference from constant (two sample case).

15.5

It is assumed that both populations are normally distributed and that the means are not known in advance but
estimated from samples of size N1 and N0 , respectively. Under these assumptions, the H0-distribution of s21 /s20 is thecentral F distribution with N1 − 1 numerator and N0 − 1
denominator degrees of freedom (FN1 −1,N0 −1 ), and the H1distribution is the same central F distribution scaled withthe variance ratio, that is, (σ12 /σ02 ) · FN1 −1,N0 −1 .

Figure 19: Effect size drawer to calculate variance ratios.

15.2

Implementation notes

Options

This test has no options.

15.6

15.3

The results were successfully checked against values produced by PASS (Hintze, 2006) and in a Monte Carlo simulation.

Examples

We want to test whether the variance σ12 in population B
is different from the variance σ02 in population A. We here
regard a ratio σ12 /σ02 > 1.5( or, the other way around, <1/1.5 = 0.6666) as a substantial difference.
We want to realize identical sample sizes in both groups.
How many subjects are needed to achieve the error levels
α = 0.05 and β = 0.2 in this test? This question can be
answered by using the following settings in G * Power :• Select
Type of power analysis: A priori• Input
Tail(s): Two
Ratio var1/var0: 1.5
α err prob: 0.05
39

Validation

16

t test: Correlation - point biserialmodel

The point biserial correlation is a measure of association
between a continuous variable X and a binary variable Y,
the latter of which takes on values 0 and 1. It is assumed
that the continuous variables X at Y = 0 and Y = 1 arenormally distributed with means µ0 , µ1 and equal varianceσ. If π is the proportion of values with Y = 1 then the point
biserial correlation coefficient is defined as:p( µ1 − µ0 ) π (1 − π )ρ=σxFigure 20: Effect size dialog to compute ρ from the coefficient ofdetermination ρ2 .

16.3

where σx = σ + (µ1 − µ0 )2 /4.
The point biserial correlation is identical to a Pearson correlation between two vectors x, y, where xi contains a value
from X at Y = j, and yi = j codes the group from which
the X was taken.
The statistical model is the same as that underlying atest for a differences in means µ0 and µ1 in two independent groups. The relation between the effect size d =(µ1 − µ0 )/σ used in that test and the point biserial correlation ρ considered here is given by:
ρ= q

We want to know how many subjects it takes to detect r =.25 in the population, given α = β = .05. Thus, H0 : ρ = 0,
H1 : ρ = 0.25.
• Select
Type of power analysis: A priori• Input
Tail(s): One
Effect size |ρ|: 0.25
α err prob: 0.05
Power (1-β err prob): 0.95d

• Outputnoncentrality parameter δ: 3.306559
Critical t: 1.654314
df: 162
Total sample size: 164
Actual power: 0.950308

2d2 + nN0 n1where n0 , n1 denote the sizes of the two groups and N =n0 + n1 .
The power procedure refers to a t-test used to evaluate
the null hypothesis that there is no (point-biserial) correlation in the population (ρ = 0). The alternative hypothesis isthat the correlation coefficient has a non-zero value r.
H0 :
H1 :

The results indicate that we need at least N = 164 subjects to ensure a power > 0.95. The actual power achieved
with this N (0.950308) is slightly higher than the requested
power.
To illustrate the connection to the two groups t test, we
calculate the corresponding effect size d for equal sample
sizes n0 = n1 = 82:

ρ=0ρ = r.

The two-sided (“two tailed”) test should be used if there
is no restriction on the sign of ρ under the alternative hypothesis. Otherwise use the one-sided (“one tailed”) test.

16.1

d= p

Effect size index

Nrn0 n1 (1 − r 2 )= p

164 · 0.25
82 · 82 · (1 − 0.252 )= 0.51639778

Performing a power analysis for the one-tailed two group t
test with this d, n0 = n1 = 82, and α = 0.05 leads to exactly
the same power 0.930308 as in the example above. If weassume unequal sample sizes in both groups, for example
n0 = 64, n1 = 100, then we would compute a different value
for d:
Nr
164 · 0.25
d= p
= p
= 0.52930772
n0 n1 (1 − r 2 )100 · 64 · (1 − 0.252 )The effect size index |ρ| is the absolute value of the correlation coefficient in the population as postulated in thealternative hypothesis. From this definition it follows that0 ≤ |ρ| < 1.
Cohen (1969, p.79) defines the following effect size conventions for |ρ|:• small ρ = 0.1

but we would again arrive at the same power. It thus posesno restriction of generality that we only input the total sample size and not the individual group sizes in the t test forcorrelation procedure.

• medium ρ = 0.3• large ρ = 0.5
Pressing the Determine button on the left side of the effect size label opens the effect size drawer (see Fig. 20). You
can use it to calculate |ρ| from the coefficient of determination ρ2 .

16.2

Examples16.4

Related tests

Similar tests in G * Power 3.0:
• Exact test for the difference of one (Pearson) correlationfrom a constant

Options

This test has no options.

• Test for the difference of two (Pearson) correlations
40

16.5

Implementation notes

The H0 -distribution is the central t-distribution with d f =N − 2 degrees of freedom. The H1 -distribution is the noncentral t-distribution with d f = N − 2 and noncentrality
parameter δ wheres| ρ |2 Nδ=
1.0 − |ρ|2
N represents the total sample size and |ρ| represents theeffect size index as defined above.

16.6

Validation

The results were checked against the values produced by
GPower 2.0 Faul and Erdfelder (1992).

41

17

t test: Linear
groups)Regression(two

output variables on the right side. The input values must
conform to the usual restrictions, that is, σxi > 0, σyi > 0,
−1 < ρi < 1. In addition, Eq (8) together with the restriction
on ρi implies the additional restriction −1 < b · σxi /σyi < 1.
Clicking on the button Calculate and transfer tomain window copies the values given in Std dev σ_x1,
Std dev σ_x2, Std dev residual σ, Allocation ration
N2/N1, and |∆ slope | to the corresponding input fields inthe main window.

A linear regression is used to estimate the parameters a, bof a linear relationship Y = a + bX between the dependent
variable Y and the independent variable X. X is assumed to
be a set of fixed values, whereas Yi is modeled as a randomvariable: Yi = a + bXi + ε i , where ε i denotes normally distributed random errors with mean 0 and standard deviation
σi . A common assumption also adopted here is that all σi ’s
are identical, that is σi = σ. The standard deviation of theerror is also called the standard deviation of the residuals.
If we have determined the linear relationships between X
and Y in two groups: Y1 = a1 + b1 X1 , Y2 = a2 + b2 X2 , we
may ask whether the slopes b1 , b2 are identical.
The null and the two-sided alternative hypotheses are
H0 :
H1 :

17.1

17.2

This test has no options.

17.3

Effect size index• Std dev residual σ
The standard deviation σ > 0 of the residuals in the
combined data set (i.e. the square root of the weighted
sum of the residual variances in the two data sets): If
σr21 and σr22 denote the variance of the residuals r1 =( a1 + b1 X1 ) − Y1 and r2 = ( a2 + b2 X2 ) − Y2 in the twogroups, and n1 , n2 the respective sample sizes, thensn1 σr21 + n2 σr22(3)
σ=n1 + n2• Std dev σ_x1
The standard deviation σx1 > 0 of the X-values ingroup 1.
• Std dev σ_x2

• Select
Type of power analysis: A priori

The standard deviation σx2 > 0 of the X-values ingroup 2.

• Input
Tail(s): Two|∆ slope|: 0.01592
α err prob: 0.05
Power (1- β): 0.80
Allocation ratio N2/N1: 1.571428
Std dev residual σ: 0.5578413
Std dev σ_x1: 9.02914
Std dev σ_x2: 11.86779

Important relationships between the standard deviations
σxi of Xi , σyi of Yi , the slopes bi of the regression lines, andthe correlation coefficient ρi between Xi and Yi are:σyi

= (bi σxi )/ρiq
= σri / 1 − ρ2i

ExamplesWe replicate an example given on page 594 in Dupont andPlummer (1998) that refers to an example in Armitage,
Berry, and Matthews (2002, p. 325). The data and relevant
statistics are shown in Fig. (24). Note: Contrary to Dupont
and Plummer (1998), we here consider the data as hypothesized true values and normalize the variance by N not( N − 1).
The relation of age and vital capacity for two groupsof men working in the cadmium industry is investigated.
Group 1 includes n1 = 28 worker with less than 10 years
of cadmium exposure, and Group 2 n2 = 44 workers never
exposed to cadmium. The standard deviation of the ages inboth groups are σx1 = 9.029 and σx2 = 11.87. Regressing vital capacity on age gives the following slopes of the regression lines β 1 = −0.04653 and β 2 = −0.03061. To calculate
the pooled standard deviation of the residuals we use theeffect size dialog: We use input mode “σ_x, σ_y, slope=> residual σ, ρ” and insert the values given above, thestandard deviations σy1 , σy2 of y (capacity) as given in Fig.
24, and the allocation ratio n2 /n1 = 44/28 = 1.571428. This
results in an pooled standard deviation of the residuals of
σ = 0.5578413 (compare the right panel in Fig. 23).
We want to recruit enough workers to detect a true difference in slope of |(−0.03061) − (−0.04653)| = 0.01592 with
80% power, α = 0.05 and the same allocation ratio to thetwo groups as in the sample data.

b1 − b2 = 0
b1 − b2 ̸= 0.

The absolute value of the difference between the slopes|∆slope| = |b1 − b2 | is used as effect size. To fully specify the effect size, the following additional inputs must be
given:

σyi

Options(4)(5)

where σi denotes the standard deviation of the residuals
Yi − (bi X + ai ).
The effect size dialog may be used to determine Std
dev residual σ and |∆ slope| from other values basedon Eqns (7), (8) and (9) given above. Pressing the button
Determine on the left side of the effect size label in the main
window opens the effect size drawer (see Fig. 23).
The left panel in Fig 23 shows the combinations of input and output values in different input modes. The input variables stand on the left side of the arrow ’=>’, the• Output
Noncentrality parameter δ:2.811598
Critical t: 1.965697
Df: 415
Sample size group 1: 163
Sample size group 2: 256
Total sample size: 419
Actual power: 0.800980
42

Figure 21: Effect size drawer to calculate the pooled standard deviation of the residuals σ and the effect size |∆ slope | from various
inputs constellations (see left panel). The right panel shows the inputs for the example discussed below.

Group 1: exposure
age capacity
age capacity
29
5.21
50
3.50
29
5.17
45
5.06
33
4.88
48
4.06
32
4.50
51
4.51
31
4.47
46
4.66
29
5.12
58
2.88
29
4.51
30
4.85
21
5.22
28
4.62
23
5.07
35
3.64
38
3.64
38
5.09
43
4.61
39
4.73
38
4.58
42
5.12
43
3.89
43
4.62
37
4.30
50
2.70
n1 =
28
sx1 =
9.029147
sy1 =0.669424b1 =-0.046532
a1 =6.230031
r1 =-0.627620

Group 2: no exposure
age capacity
age capacity
27
5.29
41
3.77
25
3.67
41
4.22
24
5.82
37
4.94
32
4.77
42
4.04
23
5.71
39
4.51
25
4.47
41
4.06
32
4.55
43
4.02
18
4.61
41
4.99
19
5.86
48
3.86
26
5.20
47
4.68
33
4.44
53
4.74
27
5.52
49
3.76
33
4.97
54
3.98
25
4.99
48
5.00
42
4.89
49
3.31
35
4.09
47
3.11
35
4.24
52
4.76
41
3.88
58
3.95
38
4.85
62
4.60
41
4.79
65
4.83
36
4.36
62
3.18
36
4.02
59
3.03
n2 =
44
sx2 =
11.867794
sy2 =0.684350
b2 =-0.030613
a2 =5.680291
r2 =-0.530876

Figure 22: Data for the example discussed Dupont and Plumer

43

17.5

The output shows that we need 419 workers in total, with
163 in group 1 and 256 in group 2. These values are close
to those reported in Dupont and Plummer (1998, p. 596) for
this example (166 + 261 = 427). The slight difference is dueto the fact that they normalize the variances by N − 1, anduse shifted central t-distributions instead of non-central t
distributions.
17.3.1

Implementation notes

The procedure implements a slight variant of the algorithm
proposed in Dupont and Plummer (1998). The only difference is, that we replaced the approximation by shifted central t-distributions used in their paper with noncentral tdistributions. In most cases this makes no great difference.
The H0-distribution is the central t distribution with d f =n1 + n2 − 4 degrees of freedom, where n1 and n2 denotethe sample size in the two groups, and the H1 distribution
is the non-central t distribution with the same degrees of√
freedom and the noncentrality parameter δ = ∆ n2 .

Relation to Multiple Regression: Special

The present procedure is essentially a special case of Multiple regression, but provides a more convenient interface. To
show this, we demonstrate how the MRC procedure can beused to compute the example above (see also Dupont andPlummer (1998, p.597)).
First, the data are combined into a data set of size n1 +n2 = 28 + 44 = 72. With respect to this combined data set,
we define the following variables (vectors of length 72):

Statistical test. The power is calculated for the t-test forequal slopes as described in Armitage et al. (2002) in chapter 11. The test statistic is (see Eqn 11.18, 11.19, 11.20):t=• y contains the measured vital capacity

12with d f = n1 − n2 − 4 degrees of freedom.
Let for group i ∈ {1, 2}, Sxi , Syi , denote the sum ofsquares in X and Y (i.e. the variance times ni ). Then apooled estimate of the residual variance be obtained by
s2r = (Sy1 + Sy2 )/(n1 + n2 − 4). The standard error of thedifference of slopes is
s 

1
1
2
+
sb̂ −b̂ = sr
2
1
Sx1
Sx2

• x1 contains the age data
• x2 codes group membership (0 = not exposed, 1 = exposed)• x3 contains the element-wise product of x1 and x2
The multiple regression model:
y = β 0 + β 1 x1 + β 2 x2 + β 3 x3 + ε ireduces to y = β 0 + β 1 x1 , and y = ( β 0 + β 2 x2 ) + ( β 1 +β 3 ) x1 , for unexposed and exposed workers, respectively. In
this model β 3 represents the difference in slope between
both groups, which is assumed to be zero under the null
hypothesis. Thus, the above model reduces toPower of the test: In the procedure
for equal slopes the√noncentrality parameter is δ = ∆ n, with ∆ = |∆slope|/σR
and
v
!
u
u
1
1
σR = tσ2 1 +
(6)
+ 2
2
mσx1
σx2

y = β 0 + β 1 x1 + β 2 x2 + ε iwhere m = n1 /n2 , σx1 and σx2 are the standard deviationsof X in group 1 and 2, respectively, and σ the common standard deviation of the residuals.

if the null hypothesis is true.
Performing a multiple regression analysis with the fullmodel leads to β 1 = −0.01592 and R21 = 0.3243. With the
reduced model assumed in the null hypothesis one finds
R20 = 0.3115. From these values we compute the following
effect size:

17.6

Validation

The results were checked for a range of input scenarios
against the values produced by the program PS publishedby Dupont and Plummer (1998). Only slight deviations
were found that are probably due to the use of the noncentral t-distribution in G * Power instead of the shifted
central t-distributions that are used in PS.

R2 − R20
0.3243 − 0.3115== 0.018870
f2 = 12
1 − 0.3243
1 − R1
Selecting an a priori analysis and setting α err prob to
0.05, Power (1-β err prob) to 0.80, Numerator df to 1
and Number of predictors to 3, we get N = 418, that is
almost the same result as in the example above.

17.4

b̂1 − b̂2sb̂ −b̂Related tests

Similar tests in G * Power 3.0:
• Multiple Regression: Omnibus (R2 deviation fromzero).
• Correlation: Point biserial model

44

18

t test: Linear
groups)Regression(two

output variables on the right side. The input values must
conform to the usual restrictions, that is, σxi > 0, σyi > 0,
−1 < ρi < 1. In addition, Eq (8) together with the restriction
on ρi implies the additional restriction −1 < b · σxi /σyi < 1.
Clicking on the button Calculate and transfer tomain window copies the values given in Std dev σ_x1,
Std dev σ_x2, Std dev residual σ, Allocation ration
N2/N1, and |∆ slope | to the corresponding input fields inthe main window.

A linear regression is used to estimate the parameters a, bof a linear relationship Y = a + bX between the dependent
variable Y and the independent variable X. X is assumed to
be a set of fixed values, whereas Yi is modeled as a randomvariable: Yi = a + bXi + ε i , where ε i denotes normally distributed random errors with mean 0 and standard deviation
σi . A common assumption also adopted here is that all σi ’s
are identical, that is σi = σ. The standard deviation of theerror is also called the standard deviation of the residuals.
If we have determined the linear relationships between X
and Y in two groups: Y1 = a1 + b1 X1 , Y2 = a2 + b2 X2 , we
may ask whether the slopes b1 , b2 are identical.
The null and the two-sided alternative hypotheses are
H0 :
H1 :

18.1

18.2

This test has no options.

18.3

Effect size index• Std dev residual σ
The standard deviation σ > 0 of the residuals in the
combined data set (i.e. the square root of the weighted
sum of the residual variances in the two data sets): If
σr21 and σr22 denote the variance of the residuals r1 =( a1 + b1 X1 ) − Y1 and r2 = ( a2 + b2 X2 ) − Y2 in the twogroups, and n1 , n2 the respective sample sizes, thensn1 σr21 + n2 σr22(7)
σ=n1 + n2• Std dev σ_x1
The standard deviation σx1 > 0 of the X-values ingroup 1.

• Select
Type of power analysis: A priori• Std dev σ_x2
The standard deviation σx2 > 0 of the X-values ingroup 2.

• Input
Tail(s): Two|∆ slope|: 0.01592
α err prob: 0.05
Power (1- β): 0.80
Allocation ratio N2/N1: 1.571428
Std dev residual σ: 0.5578413
Std dev σ_x1: 9.02914
Std dev σ_x2: 11.86779

Important relationships between the standard deviations
σxi of Xi , σyi of Yi , the slopes bi of the regression lines, andthe correlation coefficient ρi between Xi and Yi are:σyi

= (bi σxi )/ρiq
= σri / 1 − ρ2i

ExamplesWe replicate an example given on page 594 in Dupont andPlummer (1998) that refers to an example in Armitage et al.
(2002, p. 325). The data and relevant statistics are shown in
Fig. (24). Note: Contrary to Dupont and Plummer (1998),
we here consider the data as hypothesized true values andnormalize the variance by N not ( N − 1).
The relation of age and vital capacity for two groupsof men working in the cadmium industry is investigated.
Group 1 includes n1 = 28 worker with less than 10 years
of cadmium exposure, and Group 2 n2 = 44 workers never
exposed to cadmium. The standard deviation of the ages inboth groups are σx1 = 9.029 and σx2 = 11.87. Regressing vital capacity on age gives the following slopes of the regression lines β 1 = −0.04653 and β 2 = −0.03061. To calculate
the pooled standard deviation of the residuals we use theeffect size dialog: We use input mode “σ_x, σ_y, slope=> residual σ, ρ” and insert the values given above, thestandard deviations σy1 , σy2 of y (capacity) as given in Fig.
24, and the allocation ratio n2 /n1 = 44/28 = 1.571428. This
results in an pooled standard deviation of the residuals of
σ = 0.5578413 (compare the right panel in Fig. 23).
We want to recruit enough workers to detect a true difference in slope of |(−0.03061) − (−0.04653)| = 0.01592 with
80% power, α = 0.05 and the same allocation ratio to thetwo groups as in the sample data.

b1 − b2 = 0
b1 − b2 ̸= 0.

The absolute value of the difference between the slopes|∆slope| = |b1 − b2 | is used as effect size. To fully specify the effect size, the following additional inputs must be
given:

σyi

Options(8)(9)

where σi denotes the standard deviation of the residuals
Yi − (bi X + ai ).
The effect size dialog may be used to determine Std
dev residual σ and |∆ slope| from other values basedon Eqns (7), (8) and (9) given above. Pressing the button
Determine on the left side of the effect size label in the main
window opens the effect size drawer (see Fig. 23).
The left panel in Fig 23 shows the combinations of input and output values in different input modes. The input variables stand on the left side of the arrow ’=>’, the• Output
Noncentrality parameter δ:2.811598
Critical t: 1.965697
Df: 415
Sample size group 1: 163
Sample size group 2: 256
Total sample size: 419
Actual power: 0.800980

45

Figure 23: Effect size drawer to calculate the pooled standard deviation of the residuals σ and the effect size |∆ slope | from various
inputs constellations (see left panel). The right panel shows the inputs for the example discussed below.

Group 1: exposure
age capacity
age capacity
29
5.21
50
3.50
29
5.17
45
5.06
33
4.88
48
4.06
32
4.50
51
4.51
31
4.47
46
4.66
29
5.12
58
2.88
29
4.51
30
4.85
21
5.22
28
4.62
23
5.07
35
3.64
38
3.64
38
5.09
43
4.61
39
4.73
38
4.58
42
5.12
43
3.89
43
4.62
37
4.30
50
2.70
n1 =
28
sx1 =
9.029147
sy1 =0.669424b1 =-0.046532
a1 =6.230031
r1 =-0.627620

Group 2: no exposure
age capacity
age capacity
27
5.29
41
3.77
25
3.67
41
4.22
24
5.82
37
4.94
32
4.77
42
4.04
23
5.71
39
4.51
25
4.47
41
4.06
32
4.55
43
4.02
18
4.61
41
4.99
19
5.86
48
3.86
26
5.20
47
4.68
33
4.44
53
4.74
27
5.52
49
3.76
33
4.97
54
3.98
25
4.99
48
5.00
42
4.89
49
3.31
35
4.09
47
3.11
35
4.24
52
4.76
41
3.88
58
3.95
38
4.85
62
4.60
41
4.79
65
4.83
36
4.36
62
3.18
36
4.02
59
3.03
n2 =
44
sx2 =
11.867794
sy2 =0.684350
b2 =-0.030613
a2 =5.680291
r2 =-0.530876

Figure 24: Data for the example discussed Dupont and Plumer

46

18.5

The output shows that we need 419 workers in total, with
163 in group 1 and 256 in group 2. These values are close
to those reported in Dupont and Plummer (1998, p. 596) for
this example (166 + 261 = 427). The slight difference is dueto the fact that they normalize the variances by N − 1, anduse shifted central t-distributions instead of non-central t
distributions.
18.3.1

Implementation notes

The procedure implements a slight variant of the algorithm
proposed in Dupont and Plummer (1998). The only difference is, that we replaced the approximation by shifted central t-distributions used in their paper with noncentral tdistributions. In most cases this makes no great difference.
The H0-distribution is the central t distribution with d f =n1 + n2 − 4 degrees of freedom, where n1 and n2 denotethe sample size in the two groups, and the H1 distribution
is the non-central t distribution with the same degrees of√
freedom and the noncentrality parameter δ = ∆ n2 .

Relation to Multiple Regression: Special

The present procedure is essentially a special case of Multiple regression, but provides a more convenient interface. To
show this, we demonstrate how the MRC procedure can beused to compute the example above (see also Dupont andPlummer (1998, p.597)).
First, the data are combined into a data set of size n1 +n2 = 28 + 44 = 72. With respect to this combined data set,
we define the following variables (vectors of length 72):

Statistical test. The power is calculated for the t-test forequal slopes as described in Armitage et al. (2002) in chapter 11. The test statistic is (see Eqn 11.18, 11.19, 11.20):t=• y contains the measured vital capacity

12with d f = n1 − n2 − 4 degrees of freedom.
Let for group i ∈ {1, 2}, Sxi , Syi , denote the sum ofsquares in X and Y (i.e. the variance times ni ). Then apooled estimate of the residual variance be obtained by
s2r = (Sy1 + Sy2 )/(n1 + n2 − 4). The standard error of thedifference of slopes is
s 

1
1
2
+
sb̂ −b̂ = sr
2
1
Sx1
Sx2

• x1 contains the age data
• x2 codes group membership (0 = not exposed, 1 = exposed)• x3 contains the element-wise product of x1 and x2
The multiple regression model:
y = β 0 + β 1 x1 + β 2 x2 + β 3 x3 + ε ireduces to y = β 0 + β 1 x1 , and y = ( β 0 + β 2 x2 ) + ( β 1 +β 3 ) x1 , for unexposed and exposed workers, respectively. In
this model β 3 represents the difference in slope between
both groups, which is assumed to be zero under the null
hypothesis. Thus, the above model reduces toPower of the test: In the procedure
for equal slopes the√noncentrality parameter is δ = ∆ n, with ∆ = |∆slope|/σR
and
v
!
u
u
1
1
σR = tσ2 1 +
(10)
+ 2
2
mσx1
σx2

y = β 0 + β 1 x1 + β 2 x2 + ε iwhere m = n1 /n2 , σx1 and σx2 are the standard deviationsof X in group 1 and 2, respectively, and σ the common standard deviation of the residuals.

if the null hypothesis is true.
Performing a multiple regression analysis with the fullmodel leads to β 1 = −0.01592 and R21 = 0.3243. With the
reduced model assumed in the null hypothesis one finds
R20 = 0.3115. From these values we compute the following
effect size:

18.6

Validation

The results were checked for a range of input scenarios
against the values produced by the program PS publishedby Dupont and Plummer (1998). Only slight deviations
were found that are probably due to the use of the noncentral t-distribution in G * Power instead of the shifted
central t-distributions that are used in PS.

R2 − R20
0.3243 − 0.3115== 0.018870
f2 = 12
1 − 0.3243
1 − R1
Selecting an a priori analysis and setting α err prob to
0.05, Power (1-β err prob) to 0.80, Numerator df to 1
and Number of predictors to 3, we get N = 418, that is
almost the same result as in the example above.

18.4

b̂1 − b̂2sb̂ −b̂Related tests

Similar tests in G * Power 3.0:
• Multiple Regression: Omnibus (R2 deviation fromzero).
• Correlation: Point biserial model

47

19

t test: Means - difference between
two dependent means (matchedpairs)The null hypothesis of this test is that the population meansµ x , µy of two matched samples x, y are identical. The sampling method leads to N pairs ( xi , yi ) of matched observations.
The null hypothesis that µ x = µy can be reformulated interms of the difference zi = xi − yi . The null hypothesis isthen given by µz = 0. The alternative hypothesis states thatµz has a value different from zero:
H0 :
H1 :

µz = 0µz ̸= 0.

If the sign of µz cannot be predicted a priori then a twosided test should be used. Otherwise use the one-sided test.

19.1

Effect size index
Figure 25: Effect size dialog to calculate effect size d from the parameters of two correlated random variables

The effect size index dz is defined as:
dz =.

|µ x − µy |
|µz |
= q
σz
σ2 + σ2 − 2ρ xy σx σyxy“Mean group 2 = 0.4”, “SD group 1 = 1”, and “SD group 2= 1”. Finally, we set the “Correlation between groups” to be
0.55. Pressing the "Calculate and transfer to main window"
button copies the resulting effect size dz = 0.421637 to themain window. We supply the remaining input values in themain window and press "Calculate"where µ x , µy denote the population means, σx and σy denote the standard deviation in either population, and ρ xy
denotes the correlation between the two random variables.
µz and σz are the population mean and standard deviationof the difference z.
A click on the Determine button to the left of the effectsize label in the main window opens the effect size drawer(see Fig. 25). You can use this drawer to calculate dz fromthe mean and standard deviation of the differences zi . Alternatively, you can calculate dz from the means µ x and µy
and the standard deviations σx and σy of the two randomvariables x, y as well as the correlation between the randomvariables x and y.

19.2• Select
Type of power analysis: Post hoc• Input
Tail(s): Two
Effect size dz: 0.421637
α err prob: 0.05
Total sample size: 50• Output
Noncentrality parameter δ: 2.981424
Critical t: 2.009575
df: 49Power (1- β): 0.832114

Options

This test has no options.

19.3

ExamplesLet us try to replicate the example in Cohen (1969, p. 48).
The effect of two teaching methods on algebra achievements are compared between 50 IQ matched pairs of pupils(i.e. 100 pupils). The effect size that should be detected isd = (m0 − m1 )/σ = 0.4. Note that this is the effect size
index representing differences between two independent
means (two groups). We want to use this effect size as abasis for a matched-pairs study. A sample estimate of thecorrelation between IQ-matched pairs in the population has
been calculated to be r = 0.55. We thus assume ρ xy = 0.55.
What is the power of a two-sided test at an α level of 0.05?
To compute the effect size dz we open the effect size
drawer and choose "from group parameters". We only know
the ratio d = (µ x − µy )∆m/σ = 0.4. We are thus free tochoose any values for the means and (equal) standard deviations that lead to this ratio. We set “Mean group 1 = 0”,

The computed power of 0.832114 is close to the value 0.84
estimated by Cohen using his tables. To estimate the increase in power due to the correlation between pairs (i.e.,due to the shifting from a two-group design to a matchedpairs design), we enter "Correlation between groups = 0" inthe effect size drawer. This leads to dz = 0.2828427. Repeating the above analysis with this effect size leads to a powerof only 0.500352.
How many subjects would we need to arrive at a powerof about 0.832114 in a two-group design? We click X-Y plot
for a range of values to open the Power Plot window.
Let us plot (on the y axis) the power (with markers anddisplaying the values in the plot) as a function of thetotal sample size. We want to plot just 1 graph with theerr prob set to 0.05 and effect size dz fixed at 0.2828427.
48

Figure 26: Power vs. sample size plot for the example.

Clicking Draw plot yields a graph in which we can see that
we would need about 110 pairs, that is, 220 subjects (see Fig
26 on page 49). Thus, by matching pupils we cut down therequired size of the sample by more than 50%.

19.4

Related tests

19.5

Implementation notes

The H0 distribution is the central Student t distributiont( N − 1, 0), the H1 distribution is the noncentral√Student
t distribution t( N − 1, δ), with noncentrality δ = d N.

19.6

Validation

The results were checked against the values produced by
GPower 2.0.

49

20

t test: Means - difference from con- 20.3 Examples
stant (one sample case)We want to test the null hypothesis that the population
mean is µ = µ0 = 10 against the alternative hypothesisthat µ = 15. The standard deviation in the population is estimated to be σ = 8. We enter these values in the effect size
dialog: Mean H0 = 10, Mean H1 = 15, SD σ = 8 to calculate
the effect size d = 0.625.
Next we want to know how many subjects it takes todetect the effect d = 0.625, given α = β = .05. We are onlyinterested in increases in the mean and thus choose a onetailed test.

The one-sample t test is used to determine whether the population mean µ equals some specified value µ0 . The data arefrom a random sample of size N drawn from a normally
distributed population. The true standard deviation in the
population is unknown and must be estimated from thedata. The null and alternate hypothesis of the t test state:
H0 :
H1 :

µ − µ0 = 0µ − µ0 ̸= 0.

• Select
Type of power analysis: A priori

The two-sided (“two tailed”) test should be used if there
is no restriction on the sign of the deviation from µ0 assumed in the alternate hypothesis. Otherwise use the onesided (“one tailed”) test .

20.1• Input
Tail(s): One
Effect size d: 0.625
α err prob: 0.05
Power (1-β err prob): 0.95

Effect size index

The effect size index d is defined as:• Output
Noncentrality parameter δ: 3.423266
Critical t: 1.699127
df: 29Total sample size: 30
Actual power: 0.955144µ − µ0d=
σwhere σ denotes the (unknown) standard deviation in the
population. Thus, if µ and µ0 deviate by one standard deviation then d = 1.
Cohen (1969, p. 38) defines the following conventional
values for d:

The results indicates that we need at least N = 30 subjects
to ensure a power > 0.95. The actual power achieved withthis N (0.955144) is slightly higher than the requested one.
Cohen (1969, p.59) calculates the sample size needed in a
two-tailed test that the departure from the population meanis at least 10% of the standard deviation, that is d = 0.1,
given α = 0.01 and β ≤ 0.1. The input and output values
for this analysis are:• small d = 0.2• medium d = 0.5• large d = 0.8
Pressing the Determine button on the left side of the effect size label opens the effect size drawer (see Fig. 27). You
can use this dialog to calculate d from µ, µ0 and the standard deviation σ.

• Select
Type of power analysis: A priori• Input
Tail(s): Two
Effect size d: 0.1
α err prob: 0.01
Power (1-β err prob): 0.90
• Output
Noncentrality parameter δ: 3.862642
Critical t: 2.579131
df: 1491
Total sample size: 1492
Actual power: 0.900169

Figure 27: Effect size dialog to calculate effect size d means andstandard deviation.

G * Power outputs a sample size of n = 1492 which isslightly higher than the value 1490 estimated by Cohen using his tables.

.

Options

20.4

Related tests

This test has no options.

20.5

Implementation notes

20.2

The H0 distribution is the central Student t distributiont( N − 1, 0), the H1 distribution is the noncentral√Student
t distribution t( N − 1, δ), with noncentrality δ = d N.
50

20.6

Validation

The results were checked against the values produced by
GPower 2.0.

51

21

t test: Means - difference between two independent means (twogroups)

The two-sample t test is used to determine if two population means µ1 , µ2 are equal. The data are two samplesof size n1 and n2 from two independent and normally distributed populations. The true standard deviations in thetwo populations are unknown and must be estimated fromthe data. The null and alternate hypothesis of this t test are:
H0 :
H1 :

µ1 − µ2 = 0µ1 − µ2 ̸= 0.

The two-sided test (“two tails”) should be used if there
is no restriction on the sign of the deviation assumed in
the alternate hypothesis. Otherwise use the one-sided test(“one tail”).

21.1

Effect size index
Figure 28: Effect size dialog to calculate effect size d from the parameters of two independent random variables.

The effect size index d is defined as:d=µ1 − µ2σ.

• medium d = 0.5

t distribution t( N − 2, δ), where N = n1 + n2 and δ thenoncentrality parameter, which is defined as:
r
n1 n2
δ=d
n1 + n2• large d = 0.8

21.6

Cohen (1969, p. 38) defines the following conventional
values for d:• small d = 0.2

Pressing the button Determine on the left side of the effect size label opens the effect size drawer (see Fig. 28). You
can use it to calculate d from the means and standard deviations in the two populations. The t-test assumes the variances in both populations to be equal. However, the test is
relatively robust against violations of this assumption if thesample sizes are equal (n1 = n2 ). In this case a mean σ′ may
be used as the common within-population σ (Cohen, 1969,p.42):sσ′ =The results were checked against the values produced by
GPower 2.0.

σ12 + σ22
2

In the case of substantially different sample sizes n1 ̸= n2
this correction should not be used because it may lead topower values that differ greatly from the true values (Cohen, 1969, p.42).

21.2

Options

This test has no options.

21.3

Examples21.4

Related tests

21.5

Implementation notes

Validation

The H0 distribution is the central Student t distributiont( N − 2, 0); the H1 distribution is the noncentral Student

52

22

Wilcoxon signed-rank test: Means - bound for the power that results if min ARE is chosen isdifference from constant (one sam- 0.7319.
ple case)
22.2 Related tests

This test is essentially identical to the Wilcoxon signed-rank
test for matched pairs described in chapter 23. The only difference is in the calculation of the effect size d (or the corresponding moment p1 ). The effect size used here is analogous to the one sample t-test.

Related tests in G * Power 3.0:
• Means: Difference from constant.
• Proportions: Sign test.
• Means: Wilcoxon rank sum test.

22.1

Examples22.3

We replicate some power values from table 4 given in Shieh,Jan, and Randles (2007) (see table 29).
To compare these values with the power computed by
G * Power , we choose the Lehmann method without continuity correction. For effect size d = 0.1 and a normal distribution as parent we get:

Validation

The results were checked against the results given in Shiehet al. (2007), analogous computations in PASS (Hintze, 2006)
and results produced by unifyPow O’Brien (1998). There
was complete correspondence with the values given in
O’Brien, while there were slight differences to those produced by PASS. The reason of these differences seems tobe that PASS truncates the weighted sample sizes to integer
values.

• Select
Type of power analysis: Post hoc• Input
Tail(s): One
Parent distribution: Normal
Effect size d: 0.1
α err prob: 0.05
Total sample size: 649• Output
critical z: 1.6448536
Power (1- β): 0.8000589
Moment p1: 0.5398278
Moment p2: 0.5562315
Moment p3: 0.3913924
For d = 0.8 and the Laplace parent we get• Select
Type of power analysis: Post hoc• Input
Tail(s): One
Parent distribution: Laplace
Effect size dz: 0.8
α err prob: 0.05
Total sample size: 11• Output
critical z: 1.6448536
Power (1- β): 0.8308341
Moment p1: 0.8387046
Moment p2: 0.8890997
Moment p3: 0.8206572
In both cases, the values computed with G * Power using the “Lehmann method without continuity correction”
correspond closely to those given in the table for the “Exact variance” method and both values correspond closely
to simulated power.
In the latter case we would get a power of 0.923 With the
A.R.E-method if the Laplace-parent is chosen. The lower
53

d=

0.1

0.2

0.4

0.6

0.8

1.0

Figure 29: A part of table 4 in Shieh et al. (2007) that compares nominal power values computed using different methods with simulated
power

54

23

Wilcoxon
signed-rank(matched pairs)

test:

0.5

Logistic

0.4

The Wilcoxon signed-rank test is a nonparametric alternative to the one sample t test. Its use is mainly motivated byuncertainty concerning the assumption of normality madein the t test.
The Wilcoxon signed-rank test can be used to test
whether a given distribution H is symmetric about zero.
The power routines implemented in G * Power refer to theimportant special case of a “shift model”, which states that
H is obtained by subtracting two symmetric distributions F
and G, where G is obtained by shifting F by an amount ∆:G ( x ) = F ( x − ∆) for all x. The relation of this shift modelto the one sample t test is obvious if we assume that F is the
fixed distribution with mean µ0 stated in the null hypothesis and G the distribution of the test group with mean µ.
Under this assumptions H ( x ) = F ( x ) − G ( x ) is symmetric
about zero under H0 , that is if ∆ = µ − µ0 = 0 or, equivalently F ( x ) = G ( x ), and asymmetric under H1 , that is if∆ ̸= 0.

Laplace
Normal

f(x)

0.3

0.2

0.1

0.0
-6

-4

-2

0

2

4

6

x

Figure 30: Densities of the Normal, Laplace, and Logistic distributions (with µ = 0, σ = 1)

Power of the Wilcoxon rank-sum test The signed-rank
test as described above is distribution free in the sense thatits validity does not depend on the specific form of the response distribution H. This distribution independence does
no longer hold, however, if one wants to estimate numerical values for the power of the test. The reason is that theeffect of a certain shift ∆ on the deviation from symmetry
and therefore the distribution of Vs depends on the specificform of F (and G). For power calculations it is therefore
necessary to specify the response distribution F. G * Power
provides three predefined continuous and symmetric response functions that differ with respect to kurtosis, thatis the “peakedness” of the distribution:

The Wilcoxon signed-rank test The signed-rank test isbased on ranks. Assume that a sample of size N is drawn
from a distribution H ( x ). To each sample value xi a rank
S between 1 and N is assigned that corresponds to the position of | xi | in a increasingly ordered list of all absolute
sample values. The general idea of the test is to calculate
the sum of the ranks assigned to positive sample values(x > 0) and the sum of the ranks assigned to negative sample values (x < 0) and to reject the hypothesis that H issymmetric if these two rank sums are clearly different.
The actual procedure is as follows: Since the rank sum ofnegative values is known if that of positive values is given,
it suffices to consider the rank sum Vs of positive values.
The positive ranks can be specified by a n-tupel (S1 , . . . , Sn ),where 0 ≤ n ≤ N. There are ( Nn ) possible n-tuples for agiven n. Since n can take on the values 0, 1, . . . , N, the total
number of possible choices for the S’s is: ∑iN=0 ( Ni ) = 2 N .
(We here assume a continuous distribution H for which theprobabilities for ties, that is the occurance of two identical| x |, is zero.) Therefore, if the null hypothesis is true, thenthe probability to observe a particular n and a certain ntuple is:P( N+ = n; S1 = s1 , . . . , Sn = sn ) = 1/2 N . To calculate the probability to observe (under H0 ) a particular positive rank sum Vs = S1 + . . . + Sn we just need to count the
number k of all tuples with rank sum Vs and to add theirprobabilities, thus P(Vs = v) = k/2n . Doing this for all
possible Vs between the minimal value 0 corresponding to
the case n = 0, and the maximal value N ( N + 1)/2, corresponding to the n = N tuple (1, 2, . . . , N ), gives the discrete
probability distribution of Vs under H0 . This distribution
is symmetric about N ( N + 1)/4. Referring to this probability distribution we choose in a one-sided test a criticalvalue c with P(Vs ≥ c) ≤ α and reject the null hypothesis
if a rank sum Vs > c is observed. With increasing sample
size the exact distribution converges rapidly to the normal
distribution with mean E(Vs ) = N ( N + 1)/4 and variance
Var (Vs ) = N ( N + 1)(2N + 1)/24.

• Normal distribution N (µ, σ2 )2
2
1p( x ) = √ e−( x−µ) /(2σ )2π• Laplace or Double Exponential distribution:p( x ) =1 −| x|e2• Logistic distributionp( x ) =e− x(1 + e − x )2

Scaled and/or shifted versions of the Laplace and Logistic
densities that can be calculated by applying the transformation 1a p(( x − b)/a), a > 0, are again probability densities
and are referred to by the same name.
Approaches to the power analysis G * Power implements
two different methods to estimate the power for the signedrank Wilcoxon test: A) The asymptotic relative efficiency(A.R.E.) method that defines power relative to the one sample t test, and B) a normal approximation to the power proposed by Lehmann (1975, pp. 164-166). We describe the general idea of both methods in turn. More specific information
can be found in the implementation section below.
55

Var (Vs )• A.R.E-method: The A.R.E method assumes the shiftmodel described in the introduction. It relates normal
approximations to the power of the one-sample t-test(Lehmann, 1975, Eq. (4.44), p. 172) and the Wilcoxon
test for a specified H (Lehmann, 1975, Eq. (4.15), p.
160). If for a model with fixed H and ∆ the sample
size N are required to achieve a specified power for
the Wilcoxon signed-rank test and a samples size N ′ isrequired in the t test to achieve the same power, thenthe ratio N ′ /N is called the efficiency of the Wilcoxon
signed-rank test relative to the one-sample t test. The
limiting efficiency as sample size N tends to infinity
is called the asymptotic relative efficiency (A.R.E. or
Pitman efficiency) of the Wilcoxon signed rank test relative to the t test. It is given by (Hettmansperger, 1984,
p. 71):

2 
12σH

+∞
Z

2
H 2 ( x )dx  = 12

−∞

+∞
Z


x2 H ( x )dx 

−∞

+∞
Z

The value p1 is easy to interpret: If H is continuousand shifted by an amount ∆ > 0 to larger values, thenp1 is the probability to observe a negative value. For anull shift (no treatment effect, ∆ = 0, i.e. H symmetric
about zero) we get p1 = 1/2.

2
H 2 ( x )dx −∞

If c denotes the critical value of a level α test and Φ
the CDF of the standard normal distribution, then thenormal approximation of the power of the (one-sided)
test is given by:#"c − a − E(Vs )pΠ( H ) ≈ 1 − ΦVar (Vs )where a = 0.5 if a continuity correction is applied, anda = 0 otherwise.
The formulas for p1 , p2 , p3 for the predefined distributions are given in the implementation section.

23.1

Effect size index

The conventional values proposed by Cohen (1969, p. 38)
for the t-test are applicable. He defines the following conventional values for d:

If H is a normal distribution, then the A.R.E. is 3/π ≈0.955. This shows, that the efficiency of the Wilcoxon
test relative to the t test is rather high even if the assumption of normality made in the t test is true. It
can be shown that the minimal A.R.E. (for H withfinite variance) is 0.864. For non-normal distributions
the Wilcoxon test can be much more efficient than thet test. The A.R.E.s for some specific distributions are
given in the implementation notes. To estimate thepower of the Wilcoxon test for a given H with the
A.R.E. method one basically scales the sample size withthe corresponding A.R.E. value and then performs theprocedure for the t test for two independent means.

• small d = 0.2• medium d = 0.5• large d = 0.8
Pressing the button Determine on the left side of the effect size label opens the effect size dialog (see Fig. 31). You
can use this dialog to calculate d from the means and acommon standard deviations in the two populations.
If N1 = N2 but σ1 ̸= σ2 you may use a mean σ′ as common within-population σ (Cohen, 1969, p.42):s
σ12 + σ22σ′ =2

• Lehmann method: The computation of the power requires the distribution of Vs for the non-null case, thatis for cases where H is not symmetric about zero. The
Lehmann method uses the fact thatIf N1 ̸= N2 you should not use this correction, since thismay lead to power values that differ greatly from the true
values (Cohen, 1969, p.42).

Vs − E(Vs )√VarVs
tends to the standard normal distribution as N tends
to infinity for any fixed distributions H for which
0 < P( X < 0) < 1. The problem is then to compute
expectation and variance of Vs . These values dependon three “moments” p1 , p2 , p3 , which are defined as:

23.2

Options

This test has several options (see Fig. 32). First, you
may choose between the A.R.E-method and the Lehmannmethod. Given the A.R.E method you may either choose
the value for k (see implementation notes) implicitly via
the corresponding parent distribution or specify it manually. Analogously, you may specify the three moments bychoosing the distribution or “manually”. There are further
options to choose between two different types of effect size
parameter (d vs.Moment p1), the type of distribution shownin the plot window, whether both d and p2 are given in the
output, and finally, whether to use a continuity-correction.

– p1 = P ( X < 0).
– p2 = P ( X + Y > 0).
– p3 = P( X + Y > 0 and X + Z > 0).
where X, Y, Z are independent random variables with
distribution H. The expectation and variance are givenas:=N ( N − 1)( N − 2)( p3 − p21 )+ N ( N − 1)[2( p1 − p2 )2+3p2 (1 − p2 )]/2 + N p1 (1 − p1 )Note, that the A.R.E. of the Wilcoxon signed-rank test
to the one-sample t test is identical to the A.R.E of the
Wilcoxon rank-sum test to the two-sample t test (if H =F; for the meaning of F see the documentation of the
Wilcoxon rank sum test).

E(Vs )=N ( N − 1) p2 /2 + N p1
56

23.4

Related tests

Related tests in G * Power 3.0:
• Means: Difference from contrast.
• Proportions: Sign test.
• Means: Wilcoxon rank sum test.

23.5

Implementation notes

The H0 distribution is the central Student t distributiont( Nk − 2, 0); the H1 distribution is the noncentral Student t
distribution t( Nk − 2, δ), where the noncentrality parameterδ is given by:s
δ=d

N1 N2 kN1 + N2

The parameter k represents the asymptotic relative efficiency vs. correspondig t tests (Lehmann, 1975, p. 371ff)
and depends in the following way on the parent distribution:
Figure 31: Effect size dialog to calculate effect size d from the parameters of two matched random variables.

ParentUniform:
Normal:Logistic:
Laplace:min ARE:.

23.3

ExamplesWe replicate the example in O’Brien (2002, p. 141):
The question is, whether women report eating different
amounts of food on pre- vs. post-menstrual days? The null
hypothesis is that there is no difference, that is 50% report
to eat more in pre-menstrual days and 50% to eat more in
post-menstrual days. In the alternative hypothesis the assumption is that 80% of the women report to eat more in
the pre-menstrual days.
We want to know the power of a two-sided test under theassumptions that the responses are distributed according toa Laplace parent distribution, that N = 10, and α = 0.05.
We use the effect size dialog zu compute dz. In the options
we select the Lehmann method, the use of predefined parent distributions, the effect size measure d and to use thecontinuity correction.

Value of k (ARE)
1.03/pi
π 2 /93/2
0.864

Min ARE is a limiting case that gives a theoretic al minimum of the power for the Wilcoxon-Mann-Whitney test.

23.6

Validation

The results were checked against the values produced
by PASS (Hintze, 2006) and those produced by unifyPow
O’Brien (1998). There was complete correspondence withthe values given in O’Brien, while there were slight differences to those produced by PASS. The reason of thesedifferences seems to be that PASS truncates the weighted
sample sizes to integer values.

• Select
Type of power analysis: Post hoc• Input
Tail(s): Two
Parent distribution: Laplace
Effect size dz: 1.13842
α err prob: 0.05
Total sample size: 10• Output
critical z: 1.9599640
Power (1- β): 0.8534958
Moment p1: 0.9000531
Moment p2: 0.9478560
Moment p3: 0.9122280

57

Figure 32: Options for Wilcoxon tests

58

24

Wilcoxon-Mann-Whitney test of adifference between two independent means

0.5

Logistic

0.4

Laplace

The Wilcoxon-Mann-Whitney (WMW) test (or U-test) is a
nonparametric alternative to the two-group t test. Its use ismainly motivated by uncertainty concerning the assumption of normality made in the t test.
It refers to a general two sample model, in which F andG characterize response distributions under two different
conditions. The null effects hypothesis states F = G, while
the alternative is F ̸= G. The power routines implemented
in G * Power refer to the important special case of a “shiftmodel”, which states that G is obtained by shifting F by
an amount ∆: G ( x ) = F ( x − ∆) for all x. The shift model
expresses the assumption that the treatment adds a certain
amout ∆ to the response x (Additivity).

Normal

f(x)

0.3

0.2

0.1

0.0
-6

-4

-2

0

2

4

6

x

Figure 33: Densities of the Normal, Laplace, and Logistic distributions (with µ = 0, σ = 1)

The WMW test The WMW-test is based on ranks. Assume m subjects are randomly assigned to control group
A and n other subjects to treatment group B. After the experiment, all N = n + m subjects together are ranked according to a some response measure, i.e. each subject is
assigned a unique number between 1 and N. The generalidea of the test is to calculate the sum of the ranks assigned
to subjects in either group and to reject the “no effect hypothesis” if the rank sums in the two groups are clearly
different. The actual procedure is as follows: Since the mranks of control group A are known if those of treatmentgroup B are given, it suffices to consider the n ranks ofgroup B. They can be specified by a n-tupel (S1 , . . . , Sn ).
There are ( Nn ) possible n-tuples. Therefore, if the null hypothesis is true, then the probability to observe a particularn-tupel is:P(S1 = s1 , . . . , Sn = sn ) = 1/( Nn ). To calculate
the probability to observe (under H0 ) a particular rank sum
Ws = S1 + . . . + Sn we just need to count the number k of
all tuples with rank sum Ws and to add their probabilities,
thus P(Ws = w) = k/( Nn ). Doing this for all possible Ws between the minimal value n(n + 1)/2 corresponding to thetuple (1, 2, . . . , n), and the maximal value n(1 + 2m + n)/2,
corresponding to the tuple ( N − n + 1, N − n + 2, . . . , N ),gives the discrete probability distribution of Ws under H0 .
This distribution is symmetric about n( N + 1)/2. Referringto this probability distribution we choose a critical value
c with P(Ws ≥ c) ≤ α and reject the null hypothesis if
a rank sum Ws > c is observed. With increasing sample
sizes the exact distribution converges rapidly to the normal
distribution with mean E(Ws ) = n( N + 1)/2 and variance
Var (Ws ) = mn( N + 1)/12.
A drawback of using Ws is that the minimal value of Ws
depends on n. It is often more convenient to subtract theminimal value and use WXY = Ws − n(n + 1)/2 instead.
The statistic WXY (also known as the Mann-Whitney statistic) can also be derived from a slightly different perspective:
If X1 , . . . , Xm and Y1 , . . . , Yn denote the observations in thecontrol and treatment group, respectively, than WXY is thenumber of pairs (Xi , Yj ) with Xi < Yj . The approximating
normal distribution of WXY has mean E(WXY ) = mn/2 andVar (WXY ) = Var (Ws ) = mn( N + 1)/12.

Power of the WMW test The WMW test as describedabove is distribution free in the sense that its validity does
not depend on the specific form of the response distribution F. This distribution independence does not longer
hold, however, if one wants to estimate numerical values
for the power of the test. The reason is that the effect of acertain shift ∆ on the rank distribution depends on the specific form of F (and G). For power calculations it is therefore
necessary to specify the response distribution F. G * Power
provides three predefined continuous and symmetric response functions that differ with respect to kurtosis, thatis the “peakedness” of the distribution:• Normal distribution N (µ, σ2 )2
2
1p( x ) = √ e−( x−µ) /(2σ )2π• Laplace or Double Exponential distribution:p( x ) =1 −| x|e2• Logistic distributionp( x ) =e− x(1 + e − x )2

Scaled and/or shifted versions of the Laplace and Logistic
densities that can be calculated by applying the transformation 1a p(( x − b)/a), a > 0, are again probability densities
and are referred to by the same name.
Approaches to the power analysis G * Power implements
two different methods to estimate the power for the WMW
test: A) The asymptotic relative efficiency (A.R.E.) methodthat defines power relative to the two groups t test, and B) a
normal approximation to the power proposed by Lehmann(1975, pp. 69-71). We describe the general idea of both methods in turn. More specific information can be found in theimplementation section below.
59values, then p1 is the probability to observe a lower
value in the control condition than in the test condition. For a null shift (no treatment effect, ∆ = 0) we get
p1 = 1/2.

• A.R.E-method: The A.R.E method assumes the shiftmodel described in the introduction. It relates normal
approximations to the power of the t-test (Lehmann,
1975, Eq. (2.42), p. 78) and the Wilcoxon test for a specified F (Lehmann, 1975, Eq. (2.29), p. 72). If for a modelwith fixed F and ∆ the sample sizes m = n are required to achieve a specified power for the Wilcoxon
test and samples sizes m′ = n′ are required in the t test
to achieve the same power, then the ratio n′ /n is called
the efficiency of the Wilcoxon test relatve to the t test.
The limiting efficiency as sample sizes m and n tend
to infinity is called the asymptotic relative efficiency(A.R.E. or Pitman efficiency) of the Wilcoxon test relative to the t test. It is given by (Hettmansperger, 1984,
p. 71):
22

12σF2 +∞Z

F2 ( x )dx  = 12+∞Z

x2 F ( x )dx 

−∞

−∞

+∞
Z

If c denotes the critical value of a level α test and Φ
the CDF of the standard normal distribution, then thenormal approximation of the power of the (one-sided)
test is given by:#"c − a − mnp1
Π( F, G ) ≈ 1 − Φ p
Var (WXY )where a = 0.5 if a continuity correction is applied, anda = 0 otherwise.
The formulas for p1 , p2 , p3 for the predefined distributions are given in the implementation section.

F2 ( x )dx 

−∞

24.1

If F is a normal distribution, then the A.R.E. is 3/π ≈0.955. This shows, that the efficiency of the Wilcoxon
test relative to the t test is rather high even if the assumption of normality made in the t test is true. It
can be shown that the minimal A.R.E. (for F with finite variance) is 0.864. For non-normal distributions
the Wilcoxon test can be much more efficient than
the t test. The A.R.E.s for some specific distributions
are given in the implementation notes. To estimate thepower of the Wilcoxon test for a given F with the A.R.E.
method one basically scales the sample size with thecorresponding A.R.E. value and then performs the procedure for the t test for two independent means.

Effect size index

You may choose between the A.R.E. and the LehmannMethod in the options section.
A.R.E. method
defined as:

In the A.R.E. method the effect size d is∆
µ1 − µ2=σσwhere µ1 , µ2 are the means of the response functions F and
G and σ the standard deviation of the response distribution
F.
In addition to the effect size, you have to specify the
A.R.E. for F. If you want to use the predefined distributions,
you may choose the Normal, Logistic, or Laplace transformation or the minimal A.R.E. From this selection G * Power
determines the A.R.E. automatically. Alternatively, you canchoose the option to determine the A.R.E. value by hand.
In this case you must first calculate the A.R.E. for your F
using the formula given above.
The conventional values proposed by Cohen (1969, p. 38)
for the t-test are applicable. He defines the following conventional values for d:d=• Lehmann method: The computation of the power requires the distribution of WXY for the non-null caseF ̸= G. The Lehmann method uses the fact that
WXY − E(WXY )√VarWXYtends to the standard normal distribution as m and ntend to infinity for any fixed distributions F and G for
X and Y for which 0 < P( X < Y ) < 1. The problem
is then to compute expectation and variance of WXY .
These values depend on three “moments” p1 , p2 , p3 ,which are defined as:• small d = 0.2• medium d = 0.5• large d = 0.8– p1 = P ( X < Y ).

Pressing the button Determine on the left side of the effect size label opens the effect size dialog (see Fig. 34). You
can use this dialog to calculate d from the means and acommon standard deviations in the two populations.
If N1 = N2 but σ1 ̸= σ2 you may use a mean σ′ as common within-population σ (Cohen, 1969, p.42):s
σ12 + σ22σ′ =2

– p2 = P( X < Y and X < Y ′ ).
– p3 = P( X < Y and X ′ < Y ).
Note that p2 = p3 for symmetric F in the shift model.
The expectation and variance are given as:
E(WXY )

= mnp1
Var (WXY ) = mnp1 (1 − p1 ) . . .
+mn(n − 1)( p2 − p21 ) . . .

If N1 ̸= N2 you should not use this correction, since thismay lead to power values that differ greatly from the true
values (Cohen, 1969, p.42).

+nm(m − 1)( p3 − p21 )The value p1 is easy to interpret: If the response distribution G of the treatment group is shifted to larger
60butions, the effect size measure d and to use the continuity
correction.
• Select
Type of power analysis: Post hoc• Input
Tail(s): Two
Parent distribution: Laplace
Effect size d: 0.375
α err prob: 0.05
Sample size group 1: 67
Sample size group 2: 134• Output
critical z: 1.9599640
Power (1- β): 0.8472103
Moment p1: 0.6277817
Moment p2: 0.6277817
The computed power of 0.847 corresponds exactly to thevalue given in O’Brien for this case. Changing the parent
distribution to normal reduces the power to 0.68.

Figure 34: Effect size dialog to calculate effect size d from the parameters of two independent random variables with equal variance.

Example 2 This replicates the example described in
O’Brien (2002, p. 134ff) and illustrates the specification of
the effect size via the first moment p1 . To enable this input
method we select in the options the Lehmann method, theuse of predefined parent distributions, the effect size measure p1 and to use the continuity correction.
The first moment is defined as the probability that a random Y from the first group is larger than a random Y fromthe second group. The parent distribution is assumed to be“logistic” and we choose p1 = 0.6 relative to this parent distribution. An increase in SD in the second group can also
be specified. Here a value of rel. SD = 1.15 is used.

.
Lehmann method If the Lehmann method is chosen thenthe effect size is given by the first moment p1 and the relative SD. If one of the three pre-defined parent distributions
is chosen, then the remaining moments p2 and p3 are computed automatically, otherwise they must also be given.

24.2

Options

This test has several options (see Fig. 32). First, you
may choose between the A.R.E-method and the Lehmannmethod. Given the A.R.E method you may either choose
the value for k (see implementation notes) implicitly via thecorresponding parent distribution or specify it manually.
Analogously, you may specify the three moments by choosing the distribution or by entering all three values manually.
There are further options to choose between two different
types of effect size parameter (d vs.Moment p1 ), the typeof distribution shown in the plot window, whether both dand p2 are given in the output, and finally, whether to use
a continuity-correction.

24.3• Select
Type of power analysis: Post hoc• Input
Tail(s): One
Parent distribution: Logistic
Relative SD: 1.15
Moment p1: 0.6
α err prob: 0.05
Sample size group 1: 67
Sample size group 2: 134• Output
critical z: 1.6448536
Power (1- β): 0.6461276
Effect size d: 0.2911838
Moment p2: 0.4251206

ExamplesExample 1 We replicate the example in O’Brien (2002, p.
130ff): The question is, whether the injury rates in two typesof shoes differ. The null hypothesis is that there is no difference. The means 0.3 and 0.27 and a common standard
deviation sd = 0.08 are assumed in group 1 and 2, respectively.
We want to know the power of a two-sided test under theassumptions that the responses are distributed according toa Laplace parent distribution, that N1 = 67, N2 = 134 (i.e.
N1 + N2 = 201) and α = 0.05. We use the effect size dialog
to compute the effect size d = 0.375. In the options we selectthe Lehmann method, the use of predefined parent distri-The computed power 0.646 is exactly the value given in
O’Brien.
If the parent distribution is changed to “Laplace” then avery similar power 0.647 is computed. This deviates considerably from the value 0.755 given in O’Brien for this case in
his table. The reason is that G * Power always uses a single
parent, that is, if the parent is changed to “Laplace” then thegiven p1 is interpreted relative to this parent. For the table
in O’Brien, however, the effect size was always computed
61

24.6

from the Logistic parent. To replicate this behavior, one canchange to the effect size d input mode, use the effect sized given in the output for the logistic parent (d = 0.291183),and choose Laplace as the new parent:

The results were checked against the values produced
by PASS (Hintze, 2006) and those produced by unifyPow
O’Brien (1998). There was complete correspondence withthe values given in O’Brien, while there were slight differences to those produced by PASS. The reason of thesedifferences seems to be that PASS truncates the weighted
sample sizes to integer values.

• Select
Type of power analysis: Post hoc• Input
Tail(s): One
Parent distribution: Laplace
Relative SD: 1.15
Effect size d: 0.2911838
α err prob: 0.05
Sample size group 1: 67
Sample size group 2: 134• Output
critical z: 1.6448536
Power (1- β): 0.7551564
Moment p1: 0.6005708
Moment p2: 0.4405953

24.4

Related tests

24.5

Implementation notes

The procedure uses the The H0 distribution is the central
Student t distribution t( Nk − 2, 0); the H1 distribution is
the noncentral Student t distribution t( Nk − 2, δ), where thenoncentrality parameter δ is given by:s
N1 N2 kδ=d
N1 + N2
The parameter k represents the asymptotic relative efficiency vs. correspondig t tests (Lehmann, 1975, p. 371ff)
and depends in the following way on the parent distribution:
Parent Value of k (ARE)
Uniform: 1.0
Normal: 3/pi
Logistic: π 2 /9
Laplace: 3/2
min ARE: 0.864
Min ARE is a limiting case that gives a theoretical minimumof the power for the Wilcoxon-Mann-Whitney test.
24.5.1

Conversion from effect size d to p1

Given effect size d the first moment p1 is• Normal distribution: √ 
p1 = Φ d/ 2

• Logistic distribution:
p1 =√with u = dπ/ 3

Validation

1 − (1 + u) exp(−u)(1 − exp(−u))2• Laplace distribution
p1 = 1 − exp(−u)(1 + u/2)/2√
with u = d 26225t test: Generic caseWith generic t tests you can perform power analyses forany test that depends on the t distribution. All parametersof the noncentral t-distribution can be manipulated independently. Note that with Generic t-Tests you cannot doa priori power analyses, the reason being that there is no
definite association between N and df (the degrees of freedom). You need to tell G * Power the values of both N anddf explicitly.

25.1

Effect size index

In the generic case, the noncentrality parameter δ of thenoncentral t distribution may be interpreted as effect size.

25.2

Options

This test has no options.

25.3

ExamplesTo illustrate the usage we calculate with the generic t test
the power of a one sample t test. We assume N = 25,µ0 = 0, µ1 = 1, and σ = 4 and get the effect sized = (µ0 − µ1 )/σ √= (0 − 1)/4 √= −0.25, the noncentrality
parameter δ = d N = −0.25 25 = −1.25, and the degrees of freedom d f = N − 1 = 24. We choose a post hoc
analysis and a two-sided test. As result we get the power(1 − β) = 0.224525; this is exactly the same value we getfrom the specialized routine for this case in G * Power .

25.4

Related tests

Related tests in G * Power are:• Correlation: Point biserial model• Means: Difference between two independent means(two groups)• Means: Difference between two dependent means(matched pairs)• Means: Difference from constant (one sample case)

25.5

Implementation notes

One distribution is fixed to the central Student t distribution t(d f ). The other distribution is a noncentral t distribution t(d f , δ) with noncentrality parameter δ.
In an a priori analysis the specified α determines the critical tc . In the case of a one-sided test the critical value tc
has the same sign as the noncentrality parameter, otherwise
there are two critical values t1c = −t2c .

25.6

Validation

The results were checked against the values produced by
GPower 2.0.

63χ2 test: Variance - difference fromconstant (one sample case)26• Input
Tail(s): One
Ratio var1/var0: 0.6666667
α err prob: 0.05
Power (1- β): 0.80

This procedure allows power analyses for the test that the
population variance σ2 of a normally distributed randomvariable has the specific value σ02 . The null and (two-sided)
alternative hypothesis of this test are:
H0 :
H1 :

• Output
Lower critical χ2 : 60.391478
Upper critical χ2 : 60.391478
Df: 80Total sample size: 81
Actual power : 0.803686

σ − σ0 = 0σ − σ0 ̸= 0.

The two-sided test (“two tails”) should be used if there
is no restriction on the sign of the deviation assumed in
the alternative hypothesis. Otherwise use the one-sided test(“one tail”).

26.1

The output shows that using a one-sided test we need atleast 81 subjects in order to achieve the desired level of the
α and β error. To apply the test, we would estimate thevariance s2 from the sample of size N. The one-sided test
would be significant at α = 0.05 if the statistic x = ( N − 1) ·s2 /σ02 were lower than the critical value 60.39.
By setting “Tail(s) = Two”, we can easily check that a twosided test under the same conditions would have requireda much larger sample size, namely N = 103, to achieve theerror criteria.

Effect size index

The ratio σ2 /σ02 of the variance assumed in H1 to the baseline variance is used as effect size measure. This ratio is 1
if H0 is true, that is, if both variances are identical. In ana priori analysis a ratio close or even identical to 1 would
imply an exceedingly large sample size. Thus, G * Powerprohibits inputs in the range [0.999, 1.001] in this case.
Pressing the button Determine on the left side of the effect size label in the main window opens the effect size
drawer (see Fig. 35) that may be used to calculate the ratio from the two variances. Insert the baseline variance σ02
in the field variance V0 and the alternate variance in thefield variance V1.

26.4

Related tests

Similar tests in G * Power 3.0:
• Variance: Test of equality (two sample case).

26.5

Implementation notes

It is assumed that the population is normally distributed
and that the mean is not known in advance but estimatedfrom a sample of size N. Under these assumptions theH0-distribution of s2 ( N − 1)/σ02 is the central χ2 distribution with N − 1 degrees of freedom (χ2N −1 ), and the H1distribution is the same central χ2 distribution scaled withthe variance ratio, that is, (σ2 /σ02 ) · χ2N −1 .

26.6

Figure 35: Effect size drawer to calculate variance ratios.

26.2

The correctness of the results were checked against values
produced by PASS (Hintze, 2006) and in a monte carlo simulation.

Options

This test has no options.

26.3

Validation

Examples

We want to test whether the variance in a given population
is clearly lower than σ0 = 1.5. In this application we use“σ2 is less than 1” as criterion for “clearly lower”. Inserting variance V0 = 1.5 and variance V1 = 1 in the effectsize drawer we calculate as effect size a Ratio var1/var0
of 0.6666667.
How many subjects are needed to achieve the error levels
α = 0.05 and β = 0.2 in this test? This question can be
answered by using the following settings in G * Power :• Select
Type of power analysis: A priori

64

27

z test: Correlation - inequality of
two independent Pearson r’s

We have two data sets one using A with N1 = 51, and asecond using B with N2 = 206. What is the power of a twosided test for a difference between these correlations, if we
set α = 0.05?
We use the effect size drawer to calculate the effect size q
from both correlations. Setting correlation coefficient
r1 = 0.75 and correlation coefficient r2 = 0.88 yields
q = −0.4028126. The input and outputs are as follows:This procedure refers to tests of hypotheses concerning differences between two independent population correlation
coefficients. The null hypothesis states that both correlation
coefficients are identical ρ1 = ρ2 . The (two-sided) alternative hypothesis is that the correlation coefficients are different: ρ1 ̸= ρ2 :H0 : ρ1 − ρ2 = 0
H1 : ρ1 − ρ2 ̸= 0.

• Select
Type of power analysis: Post hoc• Input
Tail(s): two
Effect size q: -0.4028126
α err prob: 0.05
Sample size: 260
Sample size: 51

If the direction of the deviation ρ1 − ρ2 cannot be predicted a priori, a two-sided (’two-tailed’) test should be
used. Otherwise use a one-sided test.

27.1

Effect size index

The effect size index q is defined as a difference between
two ’Fisher z’-transformed correlation coefficients: q = z1 −z2 , with z1 = ln((1 + r1 )/(1 − r1 ))/2, z2 = ln((1 + r2 )/(1 −r2 ))/2. G * Power requires q to lie in the interval [−10, 10].
Cohen (1969, p. 109ff) defines the following effect size
conventions for q:• Output
Critical z: 60.391478
Upper critical χ2 : -1.959964
Power (1- β): 0.726352• small q = 0.1

The output shows that the power is about 0.726. This isvery close to the result 0.72 given in Example 4.3 in Cohen(1988, p. 131), which uses the same input values. The small
deviations are due to rounding errors in Cohen’s analysis.
If we instead assume N1 = N2 , how many subjects do
we then need to achieve the same power? To answer thisquestion we use an a priori analysis with the power calculated above as input and an Allocation ratio N2/N1 =1 to enforce equal sample sizes. All other parameters arechosen identical to that used in the above case. The result
is, that we now need 84 cases in each group. Thus choosing
equal sample sizes reduces the totally needed sample size
considerably, from (260+51) = 311 to 168 (84 +84).

• medium q = 0.3• large q = 0.5
Pressing the button Determine on the left side of the effect size label opens the effect size dialog:

27.4• Correlations: Difference from constant (one samplecase)

It can be used to calculate q from two correlations coefficients.
If q and r2 are given, then the formula r1 = ( a − 1)/( a +1), with a = exp[2q + ln((1 + r2 )/(1 − r2))] may be used to
calculate r1 .
G * Power outputs critical values for the zc distribution. To transform these values to critical values qc related
p to the effect size measure q use the formula: qc =zc ( N1 + N2 − 6)/(( N1 − 3)( N2 − 3)) (see Cohen, 1969, p.
135).

27.2

• Correlations: Point biserial model

27.5

Implementation notes

The H0-distribution is the standard normal distribution N (0, 1), the H1-distribution is normal distribution
N (q/s, 1), where q denotes the effect size as defined above,N
p1 and N2 the sample size in each groups, and s =1/( N1 − 3) + 1/( N2 − 3).

Options

27.6

This test has no options.

27.3

Related tests

Validation

The results were checked against the table in Cohen (1969,
chap. 4).

ExamplesAssume it is know that test A correlates with a criterionwith r1 = 0.75 and that we want to test whether an alternative test B correlates higher, say at least with r2 = 0.88.
65

28

z test: Correlation - inequality of
two dependent Pearson r’s

The null hypothesis in the special case of a ‘common index’ states that ρ a,b = ρ a,c . The (two-sided) alternative hypothesis is that these correlation coefficients are different:
ρ a,b ̸= ρ a,c :H0 : ρ a,b − ρ a,c = 0
H1 : ρ a,b − ρ a,c ̸= 0.

This procedure provides power analyses for tests of the
hypothesis that two dependent Pearson correlation coefficients ρ a,b and ρc,d are identical. The corresponding test
statistics Z1⋆ and Z2⋆ were proposed by Dunn and Clark(1969) and are described in Eqns. (11) and (12) in Steiger(1980) (for more details on these tests, see implementation
notes below).
Two correlation coefficients ρ a,b and ρc,d are dependent,
if at least one of the four possible correlation coefficients
ρ a,c , ρ a,d , ρb,c and ρb,d between other pairs of the four data
sets a, b, c, d is non-zero. Thus, in the general case wherea, b, c, and d are different data sets, we do not only have toconsider the two correlations under scrutiny but also four
additional correlations.
In the special case, where two of the data sets are identical, the two correlations are obviously always dependent,
because at least one of the four additional correlations mentioned above is exactly 1. Two other of the four additionalcorrelations are identical to the two correlations under test.
Thus, there remains only one additional correlation that canbe freely specified. In this special case we denote the H0
correlation ρ a,b , the H1 correlation ρ a,c and the additionalcorrelation ρb,c .
It is convenient to describe these two cases by two corresponding correlations matrices (that may be sub-matrices ofa larger correlations matrix): A 4 × 4-matrix in the generalcase of four different data sets (‘no common index’):1
ρ a,b ρ a,c ρ a,d ρ a,b1
ρb,c ρb,d ,
C1 = 
 ρ a,c ρb,c1
ρc,d 
ρ a,d ρb,d ρc,d1

Here, G * Power refers to the test Z1⋆ described in Eq. (11)in Steiger (1980).
If the direction of the deviation ρ a,b − ρc,d (or ρ a,b − ρ a,cin the ’common index’ case) cannot be predicted a priori, a
two-sided (’two-tailed’) test should be used. Otherwise use
a one-sided test.

28.1

Effect size index

In this procedure the correlation coefficient assumed under
H1 is used as effect size, that is ρc,d in the general case of‘no common index’ and ρ a,c in the special case of a ’common
index’.
To fully specify the effect size, the following additional
inputs are required:• ρ a,b , the correlation coefficient assumed under H0, and• all other relevant correlation coefficients that specify
the dependency between the correlations assumed in
H0 and H1: ρb,c in the ’common index’ case, andρ a,c , ρ a,d , ρb,c , ρb,d in the general case of ’no common index’.
G * Power requires the correlations assumed under H0 and
H1 to lie in the interval [−1 + ε, 1 − ε], with ε = 10−6 , andthe additional correlations to lie in the interval [−1, 1]. Ina priori analyses zero effect sizes are not allowed, because
this would imply an infinite sample size. In this case theadditional restriction |ρ a,b − ρc,d | > 10−6 (or |ρ a,b − ρ a,c | >10−6 ) holds.
Why do we not use q, the effect size proposed by Cohen(1988) for the case of two independent correlations? The effect size q is defined as a difference between two ’Fisherz’-transformed correlation coefficients: q = z1 − z2 , with
z1 = ln((1 + r1 )/(1 − r1 ))/2, z2 = ln((1 + r2 )/(1 − r2 ))/2.
The choice of q as effect size is sensible for tests of independent correlations, because in this case the power of thetest does not depend on the absolute value of the correlation coefficient assumed under H0, but only on the difference q between the transformed correlations under H0 and
H1. This is no longer true for dependent correlations andwe therefore used the effect size described above. (See theimplementation section for a more thorough discussion ofthese issues.)
Although the power is not strictly independent of thevalue of the correlation coefficient under H0, the deviations
are usually relatively small and it may therefore be convenient to use the definition of q to specify the correlation
under H1 for a given correlation under H0. In this way, one
can relate to the effect size conventions for q defined by
Cohen (1969, p. 109ff) for independent correlations:and a 3 × 3 matrix in the special case, where one of the data
sets is identical in both correlations (‘common index’):1
ρ a,b ρ a,c1
ρb,c 
C2 =  ρ a,b
ρ a,c ρb,c1
Note: The values for ρ x,y in matrices C1 and C2 cannot bechosen arbitrarily between -1 and 1. This is easily illustrated
by considering the matrix C2 : It should be obvious that wecannot, for instance, choose ρ a,b = −ρ a,c and ρb,c = 1.0,
because the latter choice implies that the other two correlations are identical. It is, however, generally not that easy to
decide whether a given matrix is a valid correlation matrix.
In more complex cases the following formal criterion can beused: A given symmetric matrix is a valid correlation matrix, if and only if the matrix is positive semi-definite, thatis, if all eigenvalues of the matrix are non-negative.
The null hypothesis in the general case with ’no common
index’ states that ρ a,b = ρc,d . The (two-sided) alternative
hypothesis is that these correlation coefficients are different:
ρ a,b ̸= ρc,d :H0 : ρ a,b − ρc,d = 0
H1 : ρ a,b − ρc,d ̸= 0.

• small q = 0.1• medium q = 0.3

Here, G * Power refers to the test Z2⋆ described in Eq. (12)in Steiger (1980).

• large q = 0.5
66

Corr ρ_bd: 0.8

The effect size drawer, which can be opened by pressing
the button Determine on the left side of the effect size label,
can be used to do this calculation:• Output
Critical z: 1.644854
Sample Size: 886
Actual Power: 0.800093
We find that the sample size in each group needs to be
N = 886. How large would N be, if we instead assume
that ρ1,4 and ρ2,3 are independent, that is, that ρ ac = ρ ad =
ρbc = ρbd = 0? To calculate this value, we may set the corresponding input fields to 0, or, alternatively, use the procedure for independent correlations with an allocation ration
N2/N1 = 1. In either case, we find a considerably larger
sample size of N = 1183 per data set (i.e. we correlate data
vectors of length N = 1183). This shows that the powerof the test increases considerably if we take dependencies
between correlations coefficients into account.
If we try to change the correlation ρbd from 0.8 to 0.9,
G * Power shows an error message stating: ‘The correlation
matrix is not valid, that is, not positive semi-definite’. This
indicates that the matrix C p with ρ3,4 changed to 0.9 is nota possible correlation matrix.

The dialog performs the following transformation: r2 =( a − 1)/( a + 1), with a = exp[−2q + ln((1 + r1 )/(1 − r1 ))],
where q, r1 and r2 denote the effect size q, and the correlations coefficient assumed under H0 and H1, respectively.

28.2

Options

This test has no options.

28.3

Examples28.3.2

We assume the following correlation matrix in the population:1
ρ1,2 ρ1,3 ρ1,4
 ρ1,21
ρ2,3 ρ2,4 
Cp = 
 ρ1,3 ρ2,3
1
ρ3,4 
ρ1,4 ρ2,4 ρ3,41

1
0.5 0.4 0.1
 0.5
1
0.2 −0.4 =  0.4 0.21
0.8 
0.1 −0.4 0.8
128.3.1

Special case: Common index

Assuming again the population correlation matrix C p
shown above, we want to do an a priori analysis for thetest whether ρ1,3 = ρ2,3 or ρ1,3 > ρ2,3 holds. With respect tothe notation used in G * Power we have the following identities: a = 3 (the common index), b = 1 (the index of thesecond data set entering in the correlation assumed underH0, here ρ a,b = ρ3,1 = ρ1,3 ), and c = 2 (the index of theremaining data set).
Thus, we have: H0 correlation ρ a,b = ρ3,1 = ρ1,3 = 0.4, H1
correlation ρ a,c = ρ3,2 = ρ2,3 = 0.2, and ρb,c = ρ1,2 = 0.5
For this effect size we want to calculate how large oursample size needs to be, in order to achieve the error levels
α = 0.05 and β = 0.2. We choose the procedure ‘Correlations: Two independent Pearson r’s (common index)’ andset:

General case: No common index

We want to perform an a priori analysis for a one-sided test
whether ρ1,4 = ρ2,3 or ρ1,4 < ρ2,3 holds.
With respect to the notation used in G * Power we havethe following identities a = 1, b = 4, c = 2, d = 3. Thus wet
get: H0 correlation ρ a,b = ρ1,4 = 0.1, H1 correlation ρc,d =ρ2,3 = 0.2, and ρ a,c = ρ1,2 = 0.5, ρ a,d = ρ1,3 = 0.4, ρb,c =
ρ4,2 = ρ2,4 = −0.4, ρb,d = ρ4,3 = ρ3,4 = 0.8.
We want to know how large our samples need to be inorder to achieve the error levels α = 0.05 and β = 0.2.
We choose the procedure ‘Correlations: Two independent
Pearson r’s (no common index)’ and set:

• Select
Type of power analysis: A priori• Input
Tail(s): one
H1 corr ρ_ac: 0.2
α err prob: 0.05
Power (1-β err prob): 0.8
H0 Corr ρ_ab: 0.4
Corr ρ_bc: 0.5• Select
Type of power analysis: A priori• Output
Critical z: 1.644854
Sample Size: 144
Actual Power: 0.801161• Input
Tail(s): one
H1 corr ρ_cd: 0.2
α err prob: 0.05
Power (1-β err prob): 0.8
H0 Corr ρ_ab: 0.1
Corr ρ_ac: 0.5
Corr ρ_ad: 0.4
Corr ρ_bc: -0.4

The answer is that we need sample sizes of 144 in eachgroup (i.e. the correlations are calculated between data vectors of length N = 144).

67

28.3.3

leads to a N × K data matrix, and pair-wise correlation of
all columns to a K × K correlation matrix. By drawing M
samples of size N one can compute M such correlation ma2 of the samtrices, and one can determine the variances σa,b
ple of M correlation coefficients r a,b , and the covariances
σa,b;c,d between samples of size M of two different correlations r a,b , rc,d . For M → ∞, the elements of Ψ, which denotes
the asymptotic variance-covariance matrix of the correlations times N, are given by [see Eqns (1) and (2) in Steiger(1980)]:

Sensitivity analyses

We now assume a scenario that is identical to that describedabove, with the exception that ρb,c = −0.6. We want toknow, what the minimal H1 correlation ρ a,c is that we candetect with α = 0.05 and β = 0.2 given a sample size
N = 144. In this sensitivity analysis, we have in general
two possible solutions, namely one for ρ a,c ≤ ρ a,b and one
for ρ a,c ≥ ρ a,b . The relevant settings for the former case are:• Select
Type of power analysis: Sensitivity

Ψ a,b;a,b
Ψ a,b;c,d• Input
Tail(s): one
Effect direction: ρ_ac ≤ ρ_ab
α err prob: 0.05
Power (1-β err prob): 0.8
Sample Size: 144
H0 Corr ρ_ab: 0.4
Corr ρ_bc: -0.6

Ψ a,b;a,c

The result is that the error levels are as requested or lower
if the H1 correlation ρ a,c is equal to or lower than 0.047702.
We now try to find the corresponding H1 correlation thatis larger than ρ a,b = 0.4. To this end, we change the effectsize direction in the settings shown above, that is, we choose
ρ a,c ≥ ρ a,b . In this case, however, G * Power shows an error
message, indicating that no solution was found. The reason
is, that there is no H1 correlation ρ a,c ≥ ρ a,b that leads toa valid (i.e. positive semi-definite) correlation matrix andsimultaneously ensures the requested error levels. To indicate a missing result the output for the H1 correlation is set
to the nonsensical value 2.
G * Power checks in both the ‘common index’ case andthe general case with no common index, whether the correlation matrix is valid and outputs a H1 correlation of 2 if
no solution is found. This is also true in the XY-plot if theH1 correlation is the dependent variable: Combinations of
input values for which no solution can be found show up
with a nonsensical value of 2.
Using the effect size dialog

28.4

Related tests

BackgroundNσa,b;a,c=

ρb,c (1 − ρ2a,b − ρ2a,c ) −−ρ a,b ρ a,c (1 − ρ2a,c − ρ2a,b − ρ2b,c )/2(13)

c a,b;a,b= ( N − 3)σza,b ;za,b = 1

c a,b;c,d= ( N − 3)σza,b ;zc,d =Ψ a,b;c,d(1 − ρ2a,b )(1 − ρ2c,d )c a,b;a,c= ( N − 3)σza,b ;za,c =Ψ a,b;a,c(1 − ρ2a,b )(1 − ρ2a,c )28.5.2

Test statistics

The test statistics proposed by Dunn and Clark (1969) are[see Eqns (12) and (13) in Steiger (1980)]:r
2 − 2s a,b;a,c
Z1⋆ = (z a,b − z a,c )/N−3r
2 − 2s a,b;c,dZ2⋆ = (z a,b − zc,d )/N−3where s a,b;a,c and s a,b;c,d denote sample estimates of the covariances c a,b;a,c and c a,b;c,d between the transformed correlations, respectively.
Note: 1. The SD of the difference of z a,b − z a,c given in
the denominator of the formula for Z1⋆ depends on thevalue of ρ a,b and ρ a,c ; the same holds analogously for Z2⋆. 2.
The only difference between Z2⋆ and the z-statistic used forindependent correlations is that in the latter the covariances a,b;c,d is assumed to be zero.

• Correlations: Two independent Pearson r’s (two sample case)

28.5.1=then the elements of the variance-covariance matrix times(N-3) of the transformed raw correlations are [see Eqs. (9)(11) in Steiger (1980)]:• Correlations: Point biserial model

Implementation notes(12)

If the raw sample correlations r a,b are transformed by the
Fisher r-to-z transform:
1 + r a,b
1
z a,b = ln
2
1 − r a,b

• Correlations: Difference from constant (one sample
case)

28.5

(11)

When two correlations have an index in common, the expression given in Eq (12) simplifies to [see Eq (3) in Steiger(1980)]:• Output
Critical z: -1.644854
H1 corr ρ_ac: 0.047702

28.3.42= Nσa,b= (1 − ρ2a,b )2= Nσa,b;b,c= [(ρ a,c − ρ a,b ρb,c )(ρb,d − ρb,c ρc,d ) + . . .
(ρ a,d − ρ a,c ρc,d )(ρb,c − ρ a,b ρ a,c ) + . . .
(ρ a,c − ρ a,d ρc,d )(ρb,d − ρ a,b ρ a,d ) + . . .
(ρ a,d − ρ a,b ρb,d )(ρb,c − ρb,d ρc,d )]/2

Let X1 , . . . , XK denote multinormally distributed randomvariables with mean vector µ and covariance matrix C.
A sample of size N from this K-dimensional distribution
6828.5.3

Central and noncentral distributions in power calculations

For the special case with a ‘common index’ the H0distribution is the standard normal distribution N (0, 1) andthe H1 distribution the normal distribution N (m1 , s1 ), withq
(2 − 2c0 )/( N − 3),s0 =with c0 = c a,b;a,c for H0, i.e. ρ a,c = ρ a,bm1
s1= (z a,b − z a,c )/s0q
(2 − 2c1 )/( N − 3) /s0 ,=with c1 = c a,b;a,c for H1.

In the general case ‘without a common index’ the H0distribution is also the standard normal distribution N (0, 1)
and the H1 distribution the normal distribution N (m1 , s1 ),withq(2 − 2c0 )/( N − 3),s0 =with c0 = c a,b;c,d for H0, i.e. ρc,d = ρ a,bm1
s1= (z a,b − zc,d )/s0q
(2 − 2c1 )/( N − 3) /s0 ,=with c1 = c a,b;a,c for H1.

28.6

Validation

The results were checked against Monte-Carlo simulations.

69

29

Z test: Multiple Logistic Regression• R2 other X.
In models with more than one covariate, the influenceof the other covariates X2 , . . . , X p on the power of thetest can be taken into account by using a correction factor. This factor depends on the proportion R2 = ρ21·23...pof the variance of X1 explained by the regression relationship with X2 , . . . , X p . If N is the sample size considering X1 alone, then the sample size in a setting withadditional covariates is: N ′ = N/(1 − R2 ). This correction for the influence of other covariates has been
proposed by Hsieh, Bloch, and Larsen (1998). R2 mustlie in the interval [0, 1].

A logistic regression model describes the relationship between a binary response variable Y (with Y = 0 and Y = 1
denoting non-occurance and occurance of an event, respectively) and one or more independent variables (covariates
or predictors) Xi . The variables Xi are themselves randomvariables with probability density function f X ( x ) (or probability distribution f X ( x ) for discrete X).
In a simple logistic regression with one covariate X theassumption is that the probability of an event P = Pr (Y =1) depends on X in the following way:P=1
e β0 + β1 x=
1 + e β0 + β1 x
1 + e−( β0 + β1 x)• X distribution:
Distribution of the Xi . There a 7 options:

For β 1 ̸= 0 and continuous X this formula describes asmooth S-shaped transition of the probability for Y = 1
from 0 to 1 (β 1 > 0) or from 1 to 0 (β 1 < 0) with increasingx. This transition gets steeper with increasing β 1 . Rearranging the formula leads to: log( P/(1 − P)) = β 0 + β 1 X. Thisshows that the logarithm of the odds P/(1 − P), also calleda logit, on the left side of the equation is linear in X. Here,β 1 is the slope of this linear relationship.
The interesting question is whether covariate Xi is related
to Y or not. Thus, in a simple logistic regression model, the nulland alternative hypothesis for a two-sided test are:
H0 :
H1 :

1. Binomial [P(k ) = ( Nk )π k (1 − π ) N −k , where k is
the number of successes (X = 1) in N trials of
a Bernoulli process with probability of success π,
0<π<1]
2. Exponential [ f ( x ) = (1/λ)e−1/λ , exponential distribution with parameter λ > 0]√
3. Lognormal [ f ( x ) = 1/( xσ 2π ) exp[−(ln x −µ)2 /(2σ2 )], lognormal distribution with parameters µ and σ > 0.]√
4. Normal [ f ( x )=1/(σ 2π ) exp[−( x −µ)2 /(2σ2 )], normal distribution with parameters µ and σ > 0)β1 = 0β 1 ̸= 0.

The procedures implemented in G * Power for this case estimates the power of the Wald test. The standard normally
distributed test statistic of the Wald test is:
z= q

β̂ 1
var( β̂ 1 )/N

5. Poisson (P( X = k) = (λk /k!)e−λ , Poisson distribution with parameter λ > 0)
6. Uniform ( f ( x ) = 1/(b − a) for a ≤ x ≤ b, f ( x ) =0 otherwise, continuous uniform distribution in
the interval [ a, b], a < b)β̂ 1=SE( β 1 )7. Manual (Allows to manually specify the varianceof β̂ under H0 and H1)

where β̂ 1 is the maximum likelihood estimator for parameter β 1 and var( β̂ 1 ) the variance of this estimate.
In a multiple logistic regression model log( P/(1 − P)) =β 0 + β 1 x1 + · · · + β p x p the effect of a specific covariate in
the presence of other covariates is tested. In this case thenull hypothesis is H0 : [ β 1 , β 2 , . . . , β p ] = [0, β 2 , . . . , β p ] andthe alternative H1 : [ β 1 , β 2 , . . . , β p ] = [ β̂, β 2 , . . . , β p ], whereβ̂ ̸= 0.

29.1

G * Power provides two different types of procedure
to calculate power: An enumeration procedure andlarge sample approximations. The Manual mode is
only available in the large sample procedures.

29.2

Options

Input mode You can choose between two input modes forthe effect size: The effect size may be given by either specifying the two probabilities p1 , p2 defined above, or insteadby specifying p1 and the odds ratio OR.

Effect size index

In the simple logistic model the effect of X on Y is given by
the size of the parameter β 1 . Let p1 denote the probabilityof an event under H0, that is exp( β 0 ) = p1 /(1 − p1 ), andp2 the probability of an event under H1 at X = 1, that isexp( β 0 + β 1 ) = p2 /(1 − p2 ). Then exp( β 0 + β 1 )/ exp( β 0 ) =exp( β 1 ) = [ p2 /(1 − p2 )]/[ p1/(1 − p1 )] := odds ratio OR,
which implies β 1 = log[OR].
Given the probability p1 (input field Pr(Y=1|X=1) H0)
the effect size is specified either directly by p2 (input field
Pr(Y=1|X=1) H1) or optionally by the odds ratio (OR) (input field Odds ratio). Setting p2 = p1 or equivalently
OR = 1 implies β 1 = 0 and thus an effect size of zero.
An effect size of zero must not be used in a priori analyses.
Besides these values the following additional inputs areneededProcedure G * Power provides two different types of procedure to estimate power. An "enumeration procedure" proposed by Lyles, Lin, and Williamson (2007) and large sample approximations. The enumeration procedure seems toprovide reasonable accurate results over a wide range ofsituations, but it can be rather slow and may need large
amounts of memory. The large sample approximations aremuch faster. Results of Monte-Carlo simulations indicatethat the accuracy of the procedures proposed by Demidenko (2007) and Hsieh et al. (1998) are comparable to thatof the enumeration procedure for N > 200. The procedurebase on the work of Demidenko (2007) is more general and70slightly more accurate than that proposed by Hsieh et al.
(1998). We thus recommend to use the procedure proposed
by Demidenko (2007) as standard procedure. The enumeration procedure of Lyles et al. (2007) may be used to validate
the results (if the sample size is not too large). It must also
be used, if one wants to compute the power for likelihood
ratio tests.

recommend to check the dependence of power on effect sizein the plot window.
Covariates with a Lognormal distribution are especiallyproblematic, because this distribution may have a very longtail (large positive skew) for larger values of m and s andmay thus easily lead to numerical problems. In version 3.1.8
the numerical stability of the procedure has been considerably improved. In addition the power with highly skewed
distributions may behave in an unintuitive manner and you
should therefore check such cases carefully.

1. The enumeration procedure provides power analyses for the Wald-test and the Likelihood ratio test.
The general idea is to construct an exemplary data
set with weights that represent response probabilities
given the assumed values of the parameters of the Xdistribution. Then a fit procedure for the generalized
linear model is used to estimate the variance of the regression weights (for Wald tests) or the likelihood ratio
under H0 and H1 (for likelihood ratio tests). The size ofthe exemplary data set increases with N and the enumeration procedure may thus be rather slow (and mayneed large amounts of computer memory) for large
sample sizes. The procedure is especially slow for analysis types other then "post hoc", which internally callthe power routine several times. By specifying a threshold sample size N you can restrict the use of the enumeration procedure to sample sizes < N. For sample
sizes ≥ N the large sample approximation selected in
the option dialog is used. Note: If a computation takes
too long you can abort it by pressing the ESC key.

29.4

We first consider a model with a single predictor X, which
is normally distributed with m = 0 and σ = 1. We assume that the event rate under H0 is p1 = 0.5 and that theevent rate under H1 is p2 = 0.6 for X = 1. The odds ratio is then OR = (0.6/0.4)/(0.5/0.5) = 1.5, and we have
β 1 = log(OR) ≈ 0.405. We want to estimate the sample size
necessary to achieve in a two-sided test with α = 0.05 a
power of at least 0.95. We want to specify the effect size interms of the odds ratio. When using the procedure of Hsieh
et al. (1998) the input and output is as follows:• SelectStatistical test: Logistic Regression
Type of power analysis: A priori• Options:
Effect size input mode: Odds ratio
Procedure: Hsieh et al. (1998)

2. G * Power provides two different large sample approximations for a Wald-type test. Both rely on the asymptotic normal distribution of the maximum likelihood
estimator for parameter β 1 and are related to themethod described by Whittemore (1981). The accuracyof these approximation increases with sample size, but
the deviation from the true power may be quite noticeable for small and moderate sample sizes. This isespecially true for X-distributions that are not symmetric about the mean, i.e. the lognormal, exponential,
and poisson distribution, and the binomial distribution with π ̸= 1/2.The approach of Hsieh et al. (1998)
is restricted to binary covariates and covariates withstandard normal distribution. The approach based on
Demidenko (2007) is more general and usually moreaccurate and is recommended as standard procedure.
For this test, a variance correction option can be selected that compensates for variance distortions that
may occur in skewed X distributions (see implementation notes). If the Hsieh procedure is selected, the program automatically switches to the procedure of Demidenko if a distribution other than the standard normal
or the binomial distribution is selected.

29.3

Examples• Input
Tail(s): Two
Odds ratio: 1.5
Pr(Y=1) H0: 0.5
α err prob: 0.05
Power (1-β err prob): 0.95
R2 other X: 0
X distribution: Normal
X parm µ: 0
X parm σ: 1
• Output
Critical z: 1.959964
Total sample size: 317
Actual power: 0.950486
The results indicate that the necessary sample size is 317.
This result replicates the value in Table II in Hsieh et al.
(1998) for the same scenario. Using the other large sample
approximation proposed by Demidenko (2007) we instead
get N = 337 with variance correction and N = 355 without.
In the enumeration procedure proposed by Lyles et al.
(2007) the χ2 -statistic is used and the output is

Possible problems• Output
Noncentrality parameter λ: 13.029675
Critical χ2 :3.841459
Df: 1
Total sample size: 358
Actual power: 0.950498

As illustrated in Fig. 36, the power of the test does not always increase monotonically with effect size, and the maximum attainable power is sometimes less than 1. In particular, this implies that in a sensitivity analysis the requested
power cannot always be reached. From version 3.1.8 on
G * Power returns in these cases the effect size which maximizes the power in the selected direction (output field "‘Actual power"’). For an overview about possible problems, we71Thus, this routine estimates the minimum sample size inthis case to be N = 358.
In a Monte-Carlo simulation of the Wald test in the above
scenario with 50000 independent cases we found a mean
power of 0.940, 0.953, 0.962, and 0.963 for samples sizes 317,
337, 355, and 358, respectively. This indicates that in thiscase the method based on Demidenko (2007) with variancecorrection yields the best approximation.
We now assume that we have additional covariates andestimate the squared multiple correlation with these others
covariates to be R2 = 0.1. All other conditions are identical.
The only change we need to make is to set the input fieldR2 other X to 0.1. Under this condition the necessary sample
size increases from 337 to a value of 395 when using theprocedure of Demidenko (2007) with variance correction.
As an example for a model with one binary covariate X
we choose the values of the fourth example in Table I in
Hsieh et al. (1998). That is, we assume that the event rate
under H0 is p1 = 0.05, and the event rate under H0 with
X = 1 is p2 = 0.1. We further assume a balanced design(π = 0.5) with equal sample frequencies for X = 0 and
X = 1. Again we want to estimate the sample size necessary
to achieve in a two-sided test with α = 0.05 a power of atleast 0.95. We want to specify the effect size directly in termsof p1 and p2 :sample size further to 2368 (in both cases the Demidenko
procedure with variance correction was used). These examples demonstrate the fact that a balanced design requires asmaller sample size than an unbalanced design, and a low
prevalence rate requires a smaller sample size than a highprevalence rate (Hsieh et al., 1998, p. 1625).

29.5

Related tests• Poisson regression

29.6

Implementation notes

29.6.1

Enumeration procedure

The procedures for the Wald- and Likelihood ratio tests are
implemented exactly as described in Lyles et al. (2007).
29.6.2Large sample approximations

The large sample procedures for the univariate case areboth related to the approach outlined in Whittemore (1981).
The correction for additional covariates has been proposedby Hsieh et al. (1998). As large sample approximations they
get more accurate for larger sample sizes.
Demidenko-procedure In the procedure based on Demidenko (2007, 2008), the H0 distribution is the standard normal distribution N (0, 1), the H1 distribution the normal distribution N (m1 , s1 ) with:q
N (1 − R2 )/v1 β 1
(14)
m1 =
q
s1 =
( av0 + (1 − a)v1 )/v1
(15)

• SelectStatistical test: Logistic Regression
Type of power analysis: A priori• Options:
Effect size input mode: Two probabilities• Input
Tail(s): Two
Pr(Y=1|X=1) H1: 0.1
Pr(Y=1|X=1) H0: 0.05
α err prob: 0.05
Power (1-β err prob): 0.95
R2 other X: 0
X Distribution: Binomial
X parm π: 0.5

where N denotes the sample size, R2 the squared multiple correlation coefficient of the covariate of interest on
the other covariates, and v1 the variance of β̂ 1 under H1,∗
whereas v0 is the
R variance of β̂ 1 for H0 with b0 = ln(µ/(1 −µ)), with µ = f X ( x ) exp(b0 + b1 x )/(1 + exp(b0 + b1 x )dx.
For the procedure without variance correction a = 0, that is,
s1 = 1. In the procedure with variance correction a = 0.75
for the lognormal distribution, a = 0.85
√ for the binomial
distribution, and a = 1, that is s1 = v0 /v1 , for all other
distributions.
The motivation of the above setting is as follows: Under
H0 the Wald-statistic has asymptotically a standard normal distribution, whereas under H1, the asymptotic normal√
distribution of the test statistic has mean β 1 /se( β̂ 1 ) =β 1 / v1 /N, and standard deviation 1. The variance for finite n is, however, biased(the degree depending on thepX-distribution) and ( av0 + (1 − a)v1 )/v1 , which is close
to 1 for symmetric X distributions but deviate from 1 forskewed ones, gives often a much better estimate of the actual standard deviation of the distribution of β̂ 1 under H1.
This was confirmed in extensive Monte-Carlo simulations
with a wide range of parameters and X distributions.
The procedure uses the result that the (m + 1) maximum
likelihood estimators β 0 , β i , . . . , β m are asymptotically normally distributed, where the variance-covariance matrix is• Output
Critical z: 1.959964
Total sample size: 1437
Actual power: 0.950068
According to these results the necessary sample size is 1437.
This replicates the sample size given in Table I in Hsieh
et al. (1998) for this example. This result is confirmed by
the procedure proposed by Demidenko (2007) with variance correction. The procedure without variance correction
and the procedure of Lyles et al. (2007) for the Wald-testyield N = 1498. In Monte-Carlo simulations of the Wald
test (50000 independent cases) we found mean power values of 0.953, and 0.961 for sample sizes 1437, and 1498,
respectively. According to these results, the procedure of
Demidenko (2007) with variance correction again yields thebest power estimate for the tested scenario.
Changing just the parameter of the binomial distribution(the prevalence rate) to a lower value π = 0.2, increases the
sample size to a value of 2158. Changing π to an equallyunbalanced but higher value, 0.8, increases the required72

given by the inverse of the (m + 1) × (m + 1) Fisher information matrix I. The (i, j)th element of I is given by"#∂2 log L
Iij = − E
∂β i ∂β j=

NE[ Xi X jrounding errors. Further checks were made against the corresponding routine in PASS Hintze (2006) and we usuallyfound complete correspondence. For multiple logistic regression models with R2 other X > 0, however, our values
deviated slightly from the result of PASS. We believe that
our results are correct. There are some indications that thereason for these deviations is that PASS internally rounds
or truncates sample sizes to integer values.
To validate the procedures of Demidenko (2007) andLyles et al. (2007) we conducted Monte-Carlo simulationsof Wald tests for a range of scenarios. In each case 150000
independent cases were used. This large number of cases isnecessary to get about 3 digits precision. In our experience,the common praxis to use only 5000, 2000 or even 1000 independent cases in simulations (Hsieh, 1989; Lyles et al.,2007; Shieh, 2001) may lead to rather imprecise and thus
misleading power estimates.
Table (2) shows the errors in the power estimatesfor different procedures. The labels "Dem(c)" and "Dem"denote the procedure of Demidenko (2007) with andwithout variance correction, the labels "LLW(W)" and"LLW(L)" the procedure of Lyles et al. (2007) for theWald test and the likelihood ratio test, respectively. All
six predefined distributions were tested (the parameters
are given in the table head). The following 6 combinations of Pr(Y=1|X=1) H0 and sample size were used:(0.5,200),(0.2,300),(0.1,400),(0.05,600),(0.02,1000). These values were fully crossed with four odds ratios (1.3, 1.5, 1.7,2.0), and two alpha values (0.01, 0.05). Max and mean errors
were calculated for all power values < 0.999. The results
show that the precision of the procedures depend on the X
distribution. The procedure of Demidenko (2007) with thevariance correction proposed here, predicted the simulated
power values best.

exp( β 0 + β 1 Xi + . . . + β m Xm )1 + exp( β 0 + β 1 Xi + . . . + β m Xm ))2

Thus, in the case of one continuous predictor, I is a 4 × 4
matrix with elements:
I00
I10 = I01
I11

=
=
=

Z ∞
−∞

Z ∞

−∞

Z ∞

−∞

GX ( x )dx
xGX ( x )dx
x2 GX ( x )dx

with
GX ( x ) : = f X ( x )exp( β 0 + β 1 x ),(1 + exp( β 0 + β 1 x ))2

where f X ( x ) is the PDF of the X distribution (for discrete
predictors, the integrals must be replaced by correspondingsums). The element M11 of the inverse of this matrix (M =I −1 ), that is the variance of β 1 , is given by: M11 = Var ( β) =2 ). In G * Power , numerical integration is
I00 /( I00 I11 − I01
used to compute these integrals.
To estimate the variance of β̂ 1 under H1, the parameterβ 0 and β 1 in the equations for Iij are chosen as implied by
the input, that is β 0 = log[ p1 /(1 − p1 )], β 1 = log[OR]. To
estimate the variance under H0, one chooses β 1 = 0 and
β 0 = β∗0 , where β∗0 is chosen as defined above.
Hsieh et al.-procedure The procedures proposed in Hsieh
et al. (1998) are used. The samples size formula for continuous, normally distributed covariates is [Eqn (1) in Hsieh etal. (1998)]:(z1−α/2 + z1− β )2N=
p1 (1 − p1) β̂2where β̂ = log([ p1/(1 − p1)]/[ p2/(1 − p2)]) is the tested
effect size, and p1 , p2 are the event rates at the mean of X
and one SD above the mean, respectively.
For binary covariates the sample size formula is [Eqn (2)in Hsieh et al. (1998)]:hN=z 1− αppqB + z1− βpp1 q1 + p2 q2 (1 − B)/Bi2

( p1 − p2 )2 (1 − B )with qi := 1 − pi , q := (1 − p), where p = (1 − B) p1 + Bp2
is the overall event rate, B is the proportion of the sample
with X = 1, and p1 , p2 are the event rates at X = 0 and
X = 1 respectively.

29.7

Validation

To check the correct implementation of the procedure proposed by Hsieh et al. (1998), we replicated all examples presented in Tables I and II in Hsieh et al. (1998). The single
deviation found for the first example in Table I on p. 1626
(sample size of 1281 instead of 1282) is probably due to73

1
Simulation (50000 samples)
Demidenko (2007) with var.corr.
Demidenko (2007)

0.9
0.8

Power 1−β

0.7
0.6
0.5
0.4
0.3
0.2
0.1
0

0

0.05

0.1

0.15

0.2
0.25
Pr(Y=1|X=1)H1

0.3

0.35

0.4

Figure 36: Results of a simulation study investigating the power as a function of effect size for Logistic regression with covariate X ∼Lognormal(0,1), p1 = 0.2, N = 100, α = 0.05, one-sided. The plot demonstrates that the method of Demidenko (2007), especially if
combined with variance correction, predicts the power values in the simulation rather well. Note also that here the power does notincrease monotonically with effect size (on the left side from Pr (Y = 1| X = 0) = p1 = 0.2) and that the maximum attainable power
may be restricted to values clearly below one.

procedureDem(c)
Dem(c)Dem
Dem
LLW(W)LLW(W)LLW(L)LLW(L)

tails
1
2
1
2
1
2
1
2

bin(0.3)
0.01320.01250.01400.01450.01440.01450.01740.0197procedureDem(c)
Dem(c)Dem
Dem
LLW(W)LLW(W)LLW(L)LLW(L)

tails
1
2
1
2
1
2
1
2

binomial
0.00450.00640.00520.00490.00520.00490.00740.0079max error
lognorm(0,1)
0.03090.03400.12730.14140.12670.1407
0.09460.1483mean error
exp
lognorm
0.01130.01200.01370.01560.02580.04650.02650.05210.0246
0.04690.02530.05220.01740.01260.02380.0209exp(1)
0.02790.03260.08790.09290.08780.09270.07900.0828norm(0,1)
0.01250.01490.03140.03580.03460.03990.01420.0155poisson(1)0.01850.01990.04720.05540.04480.05410.03590.0424uni(0,1)0.01030.01010.01090.0106
0.03150.02590.02830.0232

norm
0.00490.00610.00690.00800.00780.00920.00400.0047poisson0.00720.00830.01550.01540.01310.01390.01000.0131

uni
0.00310.00390.00350.00450.01110.00700.01150.0073

Table 1: Results of simulation. Shown are the maximum and mean error in power for different procedures (see text).

74

30

Z test: Poisson Regression

the base rate is assumed if X is increased by one unit,
this value is set to (100+10)/100 = 1.1.

A Poisson regression model describes the relationship between a Poisson distributed response variable Y (a count)
and one or more independent variables (covariates or predictors) Xi , which are themselves random variables withprobability density f X ( x ).
The probability of y events during a fixed ‘exposure time’t is:e−λt (λt)yPr (Y = y|λ, t) =y!

An input of exp( β 1 ) = 1 corresponds to "no effect" andmust not be used in a priori calculations.
• Base rate exp(β0).
This is the mean event rate assumed under H0. It must
be greater than 0.
• Mean exposure.
This is the time unit during which the events arecounted. It must be greater than 0.

It is assumed that the parameter λ of the Poisson distribution, the mean incidence rate of an event during exposure
time t, is a function of the Xi ’s. In the Poisson regression
model considered here, λ depends on the covariates Xi inthe following way:

• R2 other X.
In models with more than one covariate, the influenceof the other covariates X2 , . . . , X p on the power of thetest can be taken into account by using a correction factor. This factor depends on the proportion R2 = ρ21·23...pof the variance of X1 explained by the regression relationship with X2 , . . . , X p . If N is the sample size considering X1 alone, then the sample size in a setting withadditional covariates is: N ′ = N/(1 − R2 ). This correction for the influence of other covariates is identicalto that proposed by Hsieh et al. (1998) for the logistic
regression.

λ = exp( β 0 + β 1 X1 + β 2 X2 + · · · + β m Xm )where β 0 , . . . , β m denote regression coefficients that are estimated from the data Frome (1986).
In a simple Poisson regression with just one covariate
X1 , the procedure implemented in G * Power estimates thepower of the Wald test, applied to decide whether covariate
X1 has an influence on the event rate or not. That is, thenull and alternative hypothesis for a two-sided test are:
H0 :
H1 :

In line with the interpretation of R2 as squared correlation it must lie in the interval [0, 1].

β1 = 0β 1 ̸= 0.

• X distribution:

The standard normally distributed test statistic of theWald test is:
z= q

β̂ 1
var( β̂ 1 )/N

Distribution of the Xi . There a 7 options:
1. Binomial [P(k ) = ( Nk )π k (1 − π ) N −k , where k is
the number of successes (X = 1) in N trials of
a Bernoulli process with probability of success π,
0<π<1]β̂ 1=
SE( β 1 )2. Exponential [ f ( x ) = (1/λ)e−1/λ , exponential distribution with parameter λ > 0]√
3. Lognormal [ f ( x ) = 1/( xσ 2π ) exp[−(ln x −µ)2 /(2σ2 )], lognormal distribution with parameters µ and σ > 0.]√
4. Normal [ f ( x )=1/(σ 2π ) exp[−( x −µ)2 /(2σ2 )], normal distribution with parameters µ and σ > 0)

where β̂ 1 is the maximum likelihood estimator for parameter β 1 and var( β̂ 1 ) the variance of this estimate.
In a multiple Poisson regression model: λ = exp( β 0 +β 1 X1 + · · · + β m Xm ), m > 1, the effect of a specific covariate
in the presence of other covariates is tested. In this case thenull and alternative hypotheses are:
H0 :
H1 :

[ β 1 , β 2 , . . . , β m ] = [0, β 2 , . . . , β m ][ β 1 , β 2 , . . . , β m ] = [ β∗1 , β 2 , . . . , β m ]5. Poisson (P( X = k) = (λk /k!)e−λ , Poisson distribution with parameter λ > 0)

where β∗1 > 0.

30.1

6. Uniform ( f ( x ) = 1/(b − a) for a ≤ x ≤ b, f ( x ) =0 otherwise, continuous uniform distribution in
the interval [ a, b], a < b)

Effect size index

The effect size is specified by the ratio R of λ under H1 toλ under H0:
R=

7. Manual (Allows to manually specify the varianceof β̂ under H0 and H1)

exp( β 0 + β 1 X1 )= exp β 1 X1exp( β 0 )G * Power provides two different procedures—
an enumeration procedure and a large sample
approximation—to calculate power. The manualmode is only available in the large sample procedure.

.
The following additional inputs are needed• Exp(β1).
This is the value of the λ-ratio R defined above for X =1, that is the relative increase of the event rate overthe base event rate exp( β 0 ) assumed under H0, if X isincreased one unit. If, for instance, a 10% increase over
75

30.2

Options

ance distortions that may occur in skewed X distributions (see implementation notes).

G * Power provides two different types of procedure to estimate power. An "enumeration procedure" proposed by
Lyles et al. (2007) and large sample approximations. The
enumeration procedure seems to provide reasonable accurate results over a wide range of situations, but it can berather slow and may need large amounts of memory. The
large sample approximations are much faster. Results ofMonte-Carlo simulations indicate that the accuracy of theprocedure based on the work of Demidenko (2007) is comparable to that of the enumeration procedure for N > 200,
whereas errors of the procedure proposed by Signorini(1991) can be quite large. We thus recommend to use theprocedure based on Demidenko (2007) as the standard procedure. The enumeration procedure of Lyles et al. (2007)
may be used for small sample sizes and to validate the results of the large sample procedure using an a priori analysis (if the sample size is not too large). It must also beused, if one wants to compute the power for likelihood ratio tests. The procedure of Signorini (1991) is problematicand should not be used; it is only included to allow checksof published results referring to this widespread procedure.

30.3

ExamplesWe replicate the example given on page 449 in Signorini(1991). The number of infections Y of swimmers (X = 1)
vs. non-swimmers (X=0) during a swimming season (exposure time = 1) is tested. The infection rate is modeled
as a Poisson distributed random variable. X is assumed tobe binomially distributed with π = 0.5 (equal numbers ofswimmers and non-swimmers are sampled). The base rate,that is, the infection rate in non-swimmers, is estimated tobe 0.85. The significance level is α = 0.05. We want to know
the sample size needed to detect a 30% or greater increase
in infection rate with a power of 0.95. A 30% increase implies a relative rate of 1.3 ([100%+30%]/100%).
We first choose to use the procedure of Signorini (1991)• SelectStatistical test: Regression: Poisson Regression
Type of power analysis: A priori• Input
Tail(s): One
Exp(β1): 1.3
α err prob: 0.05
Power (1-β err prob): 0.95
Base rate exp(β0): 0.85
Mean exposure: 1.0
R2 other X: 0
X distribution: Binomial
X parm π: 0.5

1. The enumeration procedure provides power analyses for the Wald-test and the Likelihood ratio test.
The general idea is to construct an exemplary data
set with weights that represent response probabilities
given the assumed values of the parameters of the Xdistribution. Then a fit procedure for the generalized
linear model is used to estimate the variance of the regression weights (for Wald tests) or the likelihood ratio
under H0 and H1 (for likelihood ratio tests). The size ofthe exemplary data set increases with N and the enumeration procedure may thus be rather slow (and mayneed large amounts of computer memory) for large
sample sizes. The procedure is especially slow for analysis types other then "post hoc", which internally callthe power routine several times. By specifying a threshold sample size N you can restrict the use of the enumeration procedure to sample sizes < N. For sample
sizes ≥ N the large sample approximation selected in
the option dialog is used. Note: If a computation takes
too long you can abort it by pressing the ESC key.

• Output
Critical z: 1.644854
Total sample size: 697
Actual power: 0.950121
The result N = 697 replicates the value given in Signorini(1991). The other procedures yield N = 649 (Demidenko(2007) with variance correction, and Lyles et al. (2007) for
Likelihood-Ratio tests), N = 655 (Demidenko (2007) without variance correction, and Lyles et al. (2007) for Wald
tests). In Monte-Carlo simulations of the Wald test for this
scenario with 150000 independent cases per test a mean
power of 0.94997, 0.95183, and 0.96207 was found for sample sizes of 649, 655, and 697, respectively. This simulation
results suggest that in this specific case the procedure basedon Demidenko (2007) with variance correction gives themost accurate estimate.
We now assume that we have additional covariates andestimate the squared multiple correlation with these others
covariates to be R2 = 0.1. All other conditions are identical.
The only change needed is to set the input field R2 other
X to 0.1. Under this condition the necessary sample size
increases to a value of 774 (when we use the procedure of
Signorini (1991)).

2. G * Power provides two different large sample approximations for a Wald-type test. Both rely on the asymptotic normal distribution of the maximum likelihood
estimator β̂. The accuracy of these approximation increases with sample size, but the deviation from thetrue power may be quite noticeable for small andmoderate sample sizes. This is especially true for Xdistributions that are not symmetric about the mean,i.e. the lognormal, exponential, and poisson distribution, and the binomial distribution with π ̸= 1/2.
The procedure proposed by Signorini (1991) and variants of it Shieh (2001, 2005) use the "null variance formula" which is not correct for the test statistic assumed
here (and that is used in existing software) Demidenko(2007, 2008). The other procedure which is based on
the work of Demidenko (2007) on logistic regression isusually more accurate. For this test, a variance correction option can be selected that compensates for vari-Comparison between procedures To compare the accuracy of the Wald test procedures we replicated a numberof test cases presented in table III in Lyles et al. (2007) andconducted several additional tests for X-distributions not76

considered in Lyles et al. (2007). In all cases a two-sided
test with N = 200, β 0 = 0.5, α = 0.05 is assumed.
If we use the procedure of Demidenko (2007), then thecomplete G * Power input and output for the test in the firstrow of table (2) below would be:

Whittemore (1981). They get more accurate for larger sample sizes. The correction for additional covariates has been
proposed by Hsieh et al. (1998).
The H0 distribution is the standard normal distribution N (0, 1), the H1 distribution the normal distribution
N (m1 , s1 ) with:q(16)
m1 = β 1 N (1 − R2 ) · t/v0ps1 =v1 /v0(17)• SelectStatistical test: Regression: Poisson Regression
Type of power analysis: Post hoc• Input
Tail(s): TwoExp(β1): =exp(-0.1)
α err prob: 0.05
Total sample size: 200
Base rate exp(β0): =exp(0.5)
Mean exposure: 1.0
R2 other X: 0
X distribution: Normal
X parm µ: 0
X parm σ: 1in the Signorini (1991) procedure, andq
m1 = β 1 N (1 − R2 ) · t/v1
s1

= s

When using the enumeration procedure, the input is exactly the same, but in this case the test is based on the χ2 distribution, which leads to a different output.
• Output
Noncentrality parameter λ: 3.254068
Critical χ2 : 3.841459
Df: 1Power (1-β err prob): 0.438076
The rows in table (2) show power values for different test
scenarios. The column "Dist. X" indicates the distributionof the predictor X and the parameters used, the column"β 1 " contains the values of the regression weight β 1 underH1, "Sim LLW" contains the simulation results reported by
Lyles et al. (2007) for the test (if available), "Sim" contains
the results of a simulation done by us with considerable
more cases (150000 instead of 2000). The following columns
contain the results of different procedures: "LLW" = Lyles etal. (2007), "Demi" = Demidenko (2007), "Demi(c)" = Demidenko (2007) with variance correction, and "Signorini" Signorini (1991).

I00

I11

• Logistic regression

30.5.1

Enumeration procedure

=
=

Z ∞
−∞

Z ∞

−∞

Z ∞

−∞

f ( x ) exp( β 0 + β 1 x )dx
f ( x ) x exp( β 0 + β 1 x )dx
f ( x ) x2 exp( β 0 + β 1 x )dx

where f ( x ) is the PDF of the X distribution (for discrete
predictors the integrals must be replaced by corresponding sums). The element M11 of the inverse of this matrix (M = I −1 ), that is the variance of β 1 , is given by:
2 ). In G * Power , numeriM11 = Var ( β 1 ) = I00 /( I00 I11 − I01
cal integration is used to compute these integrals.
To estimate the variance of β̂ 1 under H1, the parameter β 0
and β 1 in the equations for Iij are chosen as specified in the
input. To estimate the variance under H0, one chooses β 1 =0 and β 0 = β⋆ . In the procedures proposed by Signorini(1991) β⋆ = β 0 . In the procedure with variance correction,β⋆ is chosen as defined above.

The procedures for the Wald- and Likelihood ratio tests are
implemented exactly as described in Lyles et al. (2007).
30.5.2=

Related tests

Implementation notes(19)

Thus, in the case of one continuous predictor I is a 4 × 4
matrix with elements:

I10 = I01

30.5

(18)

in the procedure based on Demidenko (2007). In these equations t denotes the mean exposure time, N the sample size,
R2 the squared multiple correlation coefficient of the covariate of interest on the other covariates, and v0 and v1
the variance of β̂ 1 under H0 and H1, respectively. Without variancecorrection s⋆ = 1, with variance correctionqs⋆ = ( av0⋆ + (1 − a)v1 )/v1 , where v0⋆ is the variance unR
der H0 for β⋆0 = log(µ), with µ = f X ( x ) exp( β 0 + β 1 x )dx.
For the lognormal distribution a = 0.75; in all other cases
a = 1. With variance correction, the value of s⋆ is often
close to 1, but deviates from 1 for X-distributions that are
not symmetric about the mean. Simulations showed that
this compensates for variance inflation/deflation in the distribution of β̂ 1 under H1 that occurs with such distributions
finite sample sizes.
The large sample approximations use the result that the(m + 1) maximum likelihood estimators β 0 , β i , . . . , β m areasymptotically (multivariate) normal distributed, where thevariance-covariance matrix is given by the inverse of the(m + 1) × (m + 1) Fisher information matrix I. The (i, j)thelement of I is given by"#∂2 log L
Iij = − E
= NE[ Xi X j e β0 + β1 Xi +...+ β m Xm ]∂β i ∂β j• Output
Critical z: -1.959964
Power (1-β err prob): 0.444593

30.4

⋆

Large sample approximations

The large sample procedures for the univariate case use thegeneral approach of Signorini (1991), which is based on theapproximate method for the logistic regression described in77Dist. X
N(0,1)
N(0,1)
N(0,1)
Unif(0,1)
Unif(0,1)
Unif(0,1)
Logn(0,1)
Logn(0,1)
Logn(0,1)
Poiss(0.5)
Poiss(0.5)
Bin(0.2)
Bin(0.2)
Exp(3)
Exp(3)

β1
-0.10
-0.15
-0.20
-0.20
-0.40
-0.80
-0.05
-0.10
-0.15
-0.20
-0.40
-0.20
-0.40
-0.40
-0.70

Sim LLW
0.449
0.796
0.950
0.167
0.478
0.923
0.275
0.750
0.947
-

Sim
0.440
0.777
0.952
0.167
0.472
0.928
0.298
0.748
0.947
0.614
0.986
0.254
0.723
0.524
0.905

LLW
0.438
0.774
0.953
0.169
0.474
0.928
0.305
0.690
0.890
0.599
0.971
0.268
0.692
0.511
0.868

Demi
0.445
0.782
0.956
0.169
0.475
0.928
0.320
0.695
0.890
0.603
0.972
0.268
0.692
0.518
0.871

Demi(c)
0.445
0.782
0.956
0.169
0.474
0.932
0.291
0.746
0.955
0.613
0.990
0.254
0.716
0.521
0.919

Signorini
0.443
0.779
0.954
0.195
0.549
0.966
0.501
0.892
0.996
0.701
0.992
0.321
0.788
0.649
0.952

Table 2: Power for different test scenarios as estimated with different procedures (see text)

5. Uniform distribution with parameters (u, v) corresponding to the interval borders.

The integrals given above happen to be proportional toderivatives of the moment generating function g( x ) :=E(etX ) of the X distribution, which is often known in closed
form. Given g( x ) and its derivatives the variance of β is calculated as:
Var ( β) =g( x )

1
g( β)
·
′′
′
2
g( β) g ( β) − g ( β) exp( β 0 )With this definition, the variance of β̂ 1 under H0 is v0 =lim Var ( β) and the variance of β̂ 1 under H1 is v1 ==e xv − e xu
x (v − u)v0 exp( β 0 )= 12/(u − v)2v1 exp( β 0 )=hx

β31 (hu − hv )(u − v)
h2u + h2v − hu+v (2 + β21 (u − v)2 )= exp( β 1 x )

β →0Var ( β 1 ).
The moment generating function for the lognormal distribution does not exist. For the other five predefined distributions this leads to:

In the manual mode you have to calculate the variances
v0 and v1 yourself. To illustrate the necessary steps, weshow the calculation for the exponential distribution (given
above) in more detail:1. Binomial distribution with parameter πg( β)= (1 − β/λ)−1 =g′ ( β)=g′′ ( β)=Var ( β)/ exp( β 0 )== (1 − π + πe x )nv0 exp( β 0 ) = π/(1 − π )v1 exp( β 0 ) = 1/(1 − π ) + 1/(π exp( β 1 ))g( x )

2. Exponential distribution with parameter λg( x )= (1 − x/λ)−12v0 exp( β 0 )= λv1 exp( β 0 )= (λ − β 1 )3 /λ= exp(µx +( λ − β )3λv0 exp( β 0 ) = Var (0) exp( β 0 )= λ2
v1 exp( β 0 ) = Var ( β 1 ) exp( β 0 )

σ2 x2
)
2

v0 exp( β 0 )= 1/σ2v1 exp( β 0 )= exp(−[ β 1 µ + ( β 1 σ)2 /2])λ( λ − β )22λ( λ − β )3g( β)′′g ( β ) g ( β ) − g ′ ( x )2=

3. Normal distribution with parameters (µ, σ)g( x )λλ−β= (λ − β 1 )3 /λNote, however, that sensitivity analyses in which the effectsize is estimated, are not possible in the manual mode. Thisis so, because in this case the argument β of the function
Var ( β) is not constant under H1 but the target of the search.

4. Poisson distribution with parameter λ= exp(λ(e x − 1))v0 exp( β 0 ) = 1/λv1 exp( β 0 ) = exp(− β 1 + λ − exp( β 1 )λ)/λ30.6g( x )

Validation

The procedure proposed by Signorini (1991) was checkedagainst the corresponding routine in PASS (Hintze, 2006)
and we found perfect agreement in one-sided tests. (Theresults of PASS 06 are wrong for normally distributed X
78

with σ ̸= 1; this error has been corrected in the newest
version PASS 2008). In two-sided tests we found small deviations from the results calculated with PASS. The reason
for these deviations is that the (usually small) contribution
to the power from the distant tail is ignored in PASS but
not in G * Power . Given the generally low accuracy of the
Signorini (1991) procedure, these small deviations are of nopractical consequence.
The results of the other two procedures were comparedto the results of Monte-Carlo simulations and with eachother. The results given in table (2) are quite representativeof these more extensive test. These results indicate that thedeviation of the computed power values may deviate fromthe true values by ±0.05.

79

31

Z test: Tetrachoric Correlation

31.0.1

BackgroundThe exact computation of the tetrachoric correlation coefficient is difficult. One reason is of a computational nature(see implementation notes). A more principal problem is,however, that frequency data are discrete, which implies
that the estimation of a cell probability can be no more accurate than 1/(2N). The inaccuracies in estimating the truecorrelation ρ are especially severe when there are cell frequencies less than 5. In these cases caution is necessary ininterpreting the estimated r. For a more thorough discussion of these issues see Brown and Benedetti (1977) and
Bonett and Price (2005).

In the model underlying tetrachoric correlation it is assumed that the frequency data in a 2 × 2-table stem from dichotimizing two continuous random variables X and Y thatare bivariate normally distributed with mean m = (0, 0)
and covariance matrix:


1 ρ
Σ=
.
ρ 1
The value ρ in this (normalized) covariance matrix is thetetrachoric correlation coefficient. The frequency data inthe table depend on the criterion used to dichotomize themarginal distributions of X and Y and on the value of thetetrachoric correlation coefficient.
From a theoretical perspective, a specific scenario is completely characterized by a 2 × 2 probability matrix:
Y=0
Y=1

X=0
p11
p21
p ⋆1

X=1
p12
p22
p ⋆2

31.0.2The implemented routines estimate the power of a test thatthe tetrachoric correlation ρ has a fixed value ρ0 . That is, thenull and alternative hypothesis for a two-sided test are:
H0 :
H1 :

Z zy Z z x

−∞ −∞

31.1

Φ( x, y, r )dxdy

Y=0
Y=1

X=1
b
d
b+d

Effect size index

The correlation coefficient assumed under H1 (H1 corr ρ)
is used as effect size.
The following additional inputs are needed to fully specify the effect size
• H0 corr ρ.

where Φ( x, y, r ) denotes the density of the bivariate normal
distribution. It should be obvious that the upper limits z x
and zy are the values at which the variables X and Y aredichotomized.
Observed frequency data are assumed to be random samples from this theoretical distribution. Thus, it is assumed
that random vectors ( xi , yi ) have been drawn from the bivariate normal distribution described above that are afterwards assigned to one of the four cells according to the‘column’ criterion xi ≤ z x vs xi > z x and the ‘row’ criterion
yi ≤ zy vs yi > zy . Given a frequency table:
X=0
a
c
a+c

ρ − ρ0 = 0ρ − ρ0 ̸= 0.

The hypotheses are identical for both the exact and the approximation mode.
In the power procedures the use of the Wald test statistic:W = (r − ρ0 )/se0 (r ) is assumed, where se0 (r ) is the standard error computed at ρ = ρ0 . As will be illustrated in theexample section, the output of G * Power may be also be
used to perform the statistical test.

p 1⋆
p 2⋆
1

where the marginal probabilities are regarded as fixed. The
whole matrix is completely specified, if the marginal probabilities p⋆2 = Pr ( X = 1), p2⋆ = Pr (Y = 1) and the tableprobability p11 are given. If z1 and z2 denote the quantilesof the standard normal distribution corresponding to themarginal probabilities p⋆1 and p1⋆ , that is Φ(z x ) = p⋆1 andΦ(zy ) = p1⋆ , then p1,1 is the CDF of the bivariate normal
distribution described above, with the upper limits z x and
zy :
p11 =

Testing the tetrachoric correlation coefficientThis is the tetrachoric correlation coefficient assumed
under H0. An input H1 corr ρ = H0 corr ρ corresponds to "no effect" and must not be used in a priori
calculations.
• Marginal prop x.
This is the marginal probability that X > z x (i.e. p∗2 )• Marginal prop y.
This is the marginal probability that Y > zy (i.e. p2∗ )The correlations must lie in the interval ] − 1, 1[, and theprobabilities in the interval ]0, 1[.

a+b
c+d
N

Effect size calculation The effect size dialog may be used
to determine H1 corr ρ in two different ways:

the central task in testing specific hypotheses about tetrachoric correlation coefficients is to estimate the correlation
coefficient and its standard error from this frequency data.
Two approaches have be proposed. One is to estimate theexact correlation coefficient (e.g. Brown & Benedetti, 1977),
the other to use simple approximations ρ∗ of ρ that are easier to compute (e.g. Bonett & Price, 2005). G * Power provides power analyses for both approaches. (See, however,
the implementation notes for a qualification of the term ‘exact’ used to distinguish between both approaches.)• A first possibility is to specify for each cell of the2 × 2 table the probability of the corresponding event
assumed under H1. Pressing the "Calculate" button
calculates the exact (Correlation ρ) and approximate(Approx. correlation ρ∗ ) tetrachoric correlation coefficient, and the marginal probabilities Marginal prob
x = p12 + p22 , and Marginal prob y = p21 + p22 . The
exact correlation coefficient is used as H1 corr ρ. (seeleft panel in Fig. 37).
80

Note: The four cell probabilities must sum to 1. It therefore suffices to specify three of them explicitly. If you
leave one of the four cells empty, G * Power computes
the fourth value as: (1 - sum of three p).

We now want to know, how many subjects we need to aachieve a power of 0.95 in a one-sided test of the H0 thatρ = 0 vs. the H1 ρ = 0.24, given the same marginal probabilities and α = 0.05.
Clicking on ‘Calculate and transfer to main window’
copies the computed H1 corr ρ = 0.2399846 and themarginal probabilities p x = 0.602 and py = 0.582 to thecorresponding input fields in the main window. The complete input and output is as follows:• A second possibility is to compute a confidence interval for the tetrachoric correlation in the populationfrom the results of a previous investigation, and tochoose a value from this interval as H1 corr ρ. In thiscase you specify four observed frequencies, the relative position 0 < k < 1 inside the confidence interval(0, 0.5, 1 corresponding to the left, central, and rightposition, respectively), and the confidence level (1 − α)of the confidence interval (see right panel in Fig. 37).

• SelectStatistical test: Correlation: Tetrachoric modelType of power analysis: A priori• Input
Tail(s): One
H1 corr ρ: 0.2399846
α err prob: 0.05
Power (1-β err prob):0.95
H0 corr ρ: 0
Marginal prob x: 0.6019313
Marginal prob y: 0.5815451

From this data G * Power computes the total sample
size N = f 11 + f 12 + f 21 + f 22 and estimates the cell
probabilities pij by: pij = ( f ij + 0.5)/( N + 2). These areused to compute the sample correlation coefficient r,
the estimated marginal probabilities, the borders ( L, R)
of the (1 − α) confidence interval for the population
correlation coefficient ρ, and the standard error of r.
The value L + ( R − L) ∗ k is used as H1 corr ρ. The
computed correlation coefficient, the confidence interval and the standard error of r depend on the choice forthe exact Brown and Benedetti (1977) vs. the approximate Bonett and Price (2005) computation mode, madein the option dialog. In the exact mode, the labels of theoutput fields are Correlation r, C.I. ρ lwr, C.I. ρ
upr, and Std. error of r, in the approximate mode
an asterisk ∗ is appended after r and ρ.

• Output
Critical z: 1.644854
Total sample size: 463
Actual power: 0.950370
H1 corr ρ: 0.239985
H0 corr ρ: 0.0
Critical r lwr: 0.122484
Critical r upr: 0.122484
Std err r: 0.074465

Clicking on the button Calculate and transfer tomain window copies the values given in H1 corr ρ, Marginprob x, Margin prob y, and - in frequency mode - Total
sample size to the corresponding input fields in the main
window.

31.2

This shows that we need at least a sample size of 463 inthis case (the Actual power output field shows the power
for a sample size rounded to an integer value).
The output also contains the values for ρ under H0 andH1 used in the internal computation procedure. In the exact computation mode a deviation from the input values
would indicate that the internal estimation procedure did
not work correctly for the input values (this should only occur at extreme values of r or marginal probabilities). In theapproximate mode, the output values correspond to the rvalues resulting from the approximation formula.
The remaining outputs show the critical value(s) for r under H0: In the Wald test assumed here, z = (r − ρ0 )/se0 (r )is approximately standard normally distributed under H0.
The critical values of r under H0 are given (a) as a quantile z1−α/2 of the standard normal distribution, and (b) in
the form of critical correlation coefficients r and standard
error se0 (r ). (In one-sided tests, the single critical value isreported twice in Critical r lwr and Critical r upr).
In the example given above, the standard error of r under
H0 is 0.074465, and the critical value for r is 0.122484. Thus,(r − ρ0 )/se(r ) = (0.122484 − 0)/0.074465 = 1.64485 = z1−α
as expected.

Options

You can choose between the exact approach in which theprocedure proposed by Brown and Benedetti (1977) isused and the approximation suggested by Bonett and Price(2005).

31.3

ExamplesTo illustrate the application of the procedure we refer to example 1 in Bonett and Price (2005): The Yes or No answersof 930 respondents to two questions in a personality inventory are recorded in a 2 × 2-table with the following result:f 11 = 203, f 12 = 186, f 21 = 167, f 22 = 374.
First we use the effect size dialog to compute from thesedata the confidence interval for the tetrachoric correlationin the population. We choose in the effect size drawer,
From C.I. calculated from observed freq. Next, we insert the above values in the corresponding fields and press
Calculate. Using the ‘exact’ computation mode (selectedin the Options dialog in the main window), we get an estimated correlation r = 0.334, a standard error of r = 0.0482,
and the 95% confidence interval [0.240, 0.429] for the population ρ. We choose the left border of the C.I. (i.e. relative
position 0, corresponding to 0.240) as the value of the tetrachoric correlation coefficient ρ under H0.

Using G * Power to perform the statistical test of H0
G * Power may also be used to perform the statistical testof H0. Assume that we want to test the H0: ρ = ρ0 = 0.4
vs the two-sided alternative H1: ρ ̸= 0.4 for α = 0.05. Assume further that we observed the frequencies f 11 = 120,81

Figure 37: Effect size drawer to calculate H1 ρ, and the marginal probabilities (see text).

The critical values for r ∗ given in the output section of themain window are [0.233, 0.541] and the standard error forr ∗ is 0.0788. Note: To compute the p-Value in the approximate mode, we should use H0 corr ρ∗ given in the output and not H0 corr ρ specified in the input. Accordingly,
using the following input in the G * Power calculator z= (0.509-0.397)/0.0788; 1-normcdf(z,0,1) yields p =0.0776, a value very close to that given above for the exact
mode.

f 12 = 45, f 21 = 56, and f 22 = 89. To perform the test we
first use the option "From C.I. calculated from observed
freq" in the effect size dialog to compute from the observed frequencies the correlation coefficient r and the estimated marginal probabilities. In the exact mode we findr = 0.513, "Est. marginal prob x" = 0.433, and "Est. marginal
prob y" = 0.468. In the main window we then choose a"post hoc" analysis. Clicking on "Calculate and transfer tomain window" in the effect size dialog copies the values
for marginal x, marginal y, and the sample size 310 to themain window. We now set "H0 corr ρ" to 0.4 and "α err
prob" to 0.05. After clicking on "Calculate" in the main window, the output section shows the critical values for thecorrelation coefficient([0.244, 0.555]) and the standard error under H0 (0.079366). These values show that the test is
not significant for the chosen α-level, because the observedr = 0.513 lies inside the interval [0.244, 0.555]. We thenuse the G * Power calculator to compute the p-value. Inserting z = (0.513-0.4)/0.0794; 1-normcdf(z,0,1) andclicking on the "Calculate" button yields p = 0.077.
If we instead want to use the approximate mode, we firstchoose this option in the "Options" dialog and then proceed in essentially the same way. In this case we find avery similar value for the correlation coefficient r ∗ = 0.509.

31.4

Related tests

31.5

Implementation notes

Given ρ and the marginal probabilties p x and py , the following procedures are used to calculate the value of ρ (inthe exact mode) or ρ∗ (in the approximate mode) and to
estimate the standard error of r and r ∗.
31.5.1

Exact mode

In the exact mode the algorithms proposed by Brown and
Benedetti (1977) are used to calculate r and to estimate thestandard error s(r ). Note that the latter is not the expected82where c = (1 − | p1∗ − p∗1 |/5 − (1/2 − pm )2 )/2, with p1∗ =p11 + p12 , p∗1 = p11 + p21 , pm = smallest marginal proportion, and ω = p11 p22 /( p12 p21 ).
The same formulas are used to compute an estimator r ∗from frequency data f ij . The only difference is that estimates p̂ij = ( f ij + 0.5)/N of the true probabilities are used.

standard error σ(r )! To compute σ (r ) would require to enumerate all possible tables Ti for the given N. If p( Ti ) and ri
denote the probability and the correlation coefficient of table i, then σ2 (r ) = ∑i (ri − ρ)2 p( Ti ) (see Brown & Benedetti,
1977, p. 349) for details]. The number of possible tables increases rapidly with N and it is therefore in general computationally too expensive to compute this exact value. Thus,‘exact’ does not mean, that the exact standard error is usedin the power calculations.
In the exact mode it is not necessary to estimate r in orderto calculate power, because it is already given in the input.
We nevertheless report the r calculated by the routine in
the output to indicate possible limitations in the precisionof the routine for |r | near 1. Thus, if the r’s reported in the
output section deviate markedly from those given in the
input, all results should be interpreted with care.
To estimate s(r ) the formula based on asymptotic theory
proposed by Pearson in 1901 is used:s (r )=

Confidence Interval The 100 ∗ (1 − α) confidence interval
for r ∗ is computed as follows:
CI = [cos(π/(1 + Lĉ )), cos(π/(1 + U ĉ ))],
where

= exp(ln ω̂ + zα/2 s(ln ω̂ ))U = exp(ln ω̂ − zα/2 s(ln ω̂ ))
 
1/2
1
1
1
1
1
+
+
+
s(ln ω̂ ) =N p̂11
p̂12
p̂21
p̂22
L

1
{( a + d)(b + d)/4+
N 3/2 ϕ(z x , zy , r )and zα/2 is the α/2 quartile of the standard normal distribution.

+( a + c)(b + d)Φ22 + ( a + b)(c + d)Φ21 ++2( ad − bc)Φ1 Φ2 − ( ab − cd)Φ2 −−( ac − bd)Φ1 }Asymptotic standard error for r ∗given by:

1/2

or with respect to cell probabilities:s (r )=∗s (r ) = k1{( p11 + p22 )( p12 + p22 )/4+
N 1/2 ϕ(z x , zy , r )31.5.3z x − rzy
− 0.5,(1 − r2 )1/2zy − rz x
ϕ− 0.5, and(1 − r2 )1/2
1×2π (1 − r2 )1/2"#
z2x − 2rz x zy + z2y )× exp −.
2(1 − r 2 )Φ(z x , zy , r )

=

m1
s1

sin[π/(1 + w)](1 + w )2

Power calculation= (ρ − ρ0 )/sr0= sr1 /sr0

The values sr0 and sr1 denote the standard error (approximate or exact) under H0 and H1.

31.6

Validation

The correctness of the procedures to calc r, s(r ) and r ∗, s(r ∗)was checked by reproducing the examples in Brown and
Benedetti (1977) and Bonett and Price (2005), respectively.
The soundness of the power routines were checked byMonte-Carlo simulations, in which we found good agreement between simulated and predicted power.

Brown and Benedetti (1977) show that this approximationis quite good if the minimal cell frequency is at least 5 (seetables 1 and 2 in Brown et al.).
31.5.2

1/2

The H0 distribution is the standard normal distribution
N (0, 1). The H1 distribution is the normal distribution withmean N (m1 , s1 ), where:

where

=

1
1
1
1
+
+
+
p̂11
p̂12
p̂21
p̂22

where w = ω̂ ĉ .

−( p11 p21 − p12 p22 )Φ1 }1/2

Φ2k = π ĉw

+2( p11 p22 − p12 p21 )Φ1 Φ2 −−( p11 p12 − p21 p22 )Φ2 −

= ϕ

1
N

with

+( p11 + p21 )( p12 + p22 )Φ22 ++( p11 + p12 )( p21 + p22 )Φ21 +Φ1

The standard error isApproximation mode

Bonett and Price (2005) propose the following approximations.
Correlation coefficient Their approximation ρ∗ of thetetrachoric correlation coefficient is:ρ∗ = cos(π/(1 + ω c )),83

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85