B.1 

Appendix B:

Ordinary Least Squares and 

Poisson Regression Models 

by 

Luc Anselin 

Arizona State University 

Tempe, AZ 

This note provides a brief description of the statistical background, estimators and 

model characteristics for a regression specification, estimated by means of both Ordinary 

Least Squares (OLS) and Poisson regression. 

 

Ordinary Least Squares Regression

With an assumption of normality for the regression error term, OLS also 

corresponds to Maximum Likelihood (ML) estimation. The note contains the statistical 

model and all expressions that are needed to carry out estimation and essential model 

diagnostics. Both concise matrix notation as well as more extensive full summation 

notation are employed, to provide a direct link to “loop” structures in the software code, 

except when full summation is too unwieldy (e.g., for matrix inverse). Some references 

are provided for general methodological descriptions. 

Statistical Issues 

The classical multivariate linear regression model stipulates a linear relationship 

between a dependent variable (also called a response variable) and a set of explanatory

variables (also called independent variables, or covariates). The relationship is stochastic, 

in the sense that the model is not exact, but subject to random variation, as expressed in 

an error term (also called disturbance term). 

Formally, for each observation i, the value of the dependent variable, Yi, is related 

to a sum of K explanatory variables, Xih, with h=1,...,K, each multiplied with a regression 

coefficient, βh, and the random error term, i: 

௜ߝ ൅ ௛ߚ௛௜ܺ ∑ ൌ௜ ܻ

௄

௛ୀଵ (B.1) 

Typically, the first explanatory variable is set equal to one, and referred to as the 

constant term. Its coefficient is referred to as the intercept, the other coefficients are 
B.2 

slopes. Using a constant term amounts to extracting a mean effect and is equivalent to 

using all variables as deviations from their mean. In practice, it is highly recommended to 

always include a constant term. 

In matrix notation, which summarizes all observations, i=1,...,N, into a single 

compact expression, an N by 1 vector of values for the dependent variable, y is related to 

an N by K matrix of values for the explanatory variables, X, a K by 1 vector of regression 

coefficients, β, and an N by 1 vector of random error terms,  : 

	ܻ ൌ ܺߚ ൅ ߳) B.2) 

This model stipulates that on average, when values are observed for the 

explanatory variables, X, the value for the dependent variable equals Xβ, or: 

 

ܧሺܻ|ܺሻ ൌ ܺߚ) B.3) 

where E[ | ] is the conditional expectation operator. This is referred to as a specification 

for the conditional mean, conditional because X must be observed. It is a theoretical 

model, built on many assumptions. In practice, one does not know the coefficient vector, 

β, nor is the error term observed. 

Estimation boils down to finding a “good” value for the β, with known statistical 

properties. The statistical properties depend on what is assumed in terms of the 

characteristics of the distribution of the unknown (and never observed) error term. To 

obtain a Maximum Likelihood estimator, the complete distribution must be specified, 

typically as a normal distribution, with mean zero and variance, σ

2

. The mean is set to 

zero to avoid systematic under- or over-prediction. The variance is an unknown 

characteristic of the distribution that must be estimated together with the coefficients, β. 

The estimate for β (Greek letter) will be referred to as b (Latin letter with bh as the 

estimate for the individual coefficient, βh). 

The estimator is the procedure followed to obtain an estimate, such as OLS, for 

bOLS, or ML, for bML. The residual of the regression is the difference between the 

observed value and the predicted value, typically referred to as e. For each observation, 

݁௜ ൌ ܻ௜ െ ∑ ܺ௜௛ߚ௛

௄

௛ୀଵ (B.4) 

or, in matrix notation, with ܻ෠ ൌ ܾܺ	as short hand for the vector of predicted values, 

݁ൌܻെܻ෠ (B.5) 
B.3 

Note that the residual is not the same as the error term, but only serves as an 

estimate for the error. What is of interest is not so much the individual residuals, but the 

properties of the (unknown) error distribution. Within the constraints of the model 

assumptions, some of the characteristics of the error distribution can be estimated from 

the residuals, such as the error variance, σ

2

, whose estimate is referred to as s2. 

Because the model has a random component, the observed y are random as well, 

and any “statistic” computed using these observed data will be random too. Therefore, 

the estimates b will have a distribution, intimately linked to the assumed distribution for 

the error term. When the error is taken to be normally distributed, the regression 

coefficient will also follow a normal distribution. Statistical inference (significance tests) 

can be carried out once the characteristics (parameters) of that distribution have been 

obtained (they are never known, but must be estimated from the data as well). An 

important result is that OLS is unbiased. In other words, the mean of the distribution of 

the estimate b is β, the true, but unknown, coefficient, such that “on average,” the 

estimation is on target. Also, the variance of the distribution of b is directly related to the 

variance of the error term (and the values for the X). It can be computed by replacing σ

2

by its estimate, s2

. 

An extensive discussion of the linear regression model can be found in most texts 

on linear modeling, multivariate statistics, or econometrics, for example, Rao (1973), 

Greene (2000), or Wooldridge (2002). 

Ordinary Least Squares Estimator 

In its most basic form, OLS is simply a fitting mechanism, based on minimizing 

the sum of squared residuals or residual sum of squares (RSS). Formally, bOLS is the 

vector of parameter values that minimizes 

ܴܵܵ ൌ ∑ ݁௜

ଶ ൌ ∑ ሺܻ௜ െ ∑ ܺ௜௝ܾ௛ሻ ௄ ଶ

௛ୀଵ

ே

௜ୀଵ

ே

௜ୀଵ (B.6) 

or, in matrix notation, 

ܴܵܵ ൌ ݁ᇱ

݁ ൌ ሺܻ െ ܾܺሻᇱሺܻ െ ܾܺሻ (B.7) 

The solution to this minimization problem is given by the so-called normal 

equations, a system of K equations of the form: 
B.4 

∑ ሺܻ௜ െ ∑ ܺ௜௛ܾ௛ሻܺ௜௛ ൌ 0 ௄

௛ୀଵ

ே

௜ୀଵ (B.8) 

for h=1 to K, or, in matrix notation, 

ܺᇱ

ሺܻ െ ܾܺሻ ൌ 0 (B.9) 

ܺᇱ

ܾܺ ൌ 	 ܺᇱܻ (B.10) 

The solution to this system of equations yields the familiar matrix expression for 

bOLS: 

ܾை௅ௌ ൌ ሺܺᇱܺሻିଵܺᇱܻ (B.11) 

An estimate for the error variance follows as 

ݏை௅ௌ

ଶ ൌ ∑ ሺ௒೔ି∑ ௑೔ೕ௕ೀಽೄ,೓ሻ ಼ మ ೓సభ

ಿ

೔సభ

ேି௄ (B.12) 

or, in matrix notation, 

ݏை௅ௌ

ଶ ൌ ௘ᇲ௘

ேି௄ (B.13) 

It can be shown that when the X are exogenous

1

 only the assumption that E[]=0 is 

needed to show that the OLS estimator is unbiased. With the additional assumption of a 

fixed error variance, s

2

, OLS is also most efficient, in the sense of having the smallest 

variance among all other linear and unbiased estimators. This is referred to as the BLUE 

(Best Linear Unbiased Estimator) property of OLS. Note, that in order to obtain these 

properties, no additional assumptions need to be made about the distribution of the error 

term. However, to carry out statistical inference, such as significance tests, this is 

insufficient, and further characteristics of the error distribution need to be specified (such 

as assuming a normal distribution) or asymptotic assumptions need to be invoked in the 

form of laws of large numbers (typically yielding a normal distribution). 

 

1

 In practice, this means that each explanatory variable must be uncorrelated with the error term. The 

easiest way to ensure this is to assume that the X are fixed. But even when they are not, this property 

holds, as long as the randomness in X and  are not related. In other words, knowing something about 

the value of an explanatory variable should not provide any information about the error term. 

Formally, this means that X and  must be orthogonal, or E[X’]=0. Failure of this assumption will lead 

to so-called simultaneous equation bias. 
B.5 

Maximum Likelihood Estimator 

When the error terms are assumed to be independently distributed as normal 

random variables, OLS turns out to be equivalent to ML. 

Maximum Likelihood estimation proceeds as follows. First, consider the density 

for a single error term: 

ߝ~ܰሺ0, ߪଶሻ (B.14) 

or 

f([]i|s2)= 1

2Aσ2

e-(1/2)(i

2/σ2) (B.15) 

 

A subtle, but important, point is that the error itself is not observed, but only the 

“data” (y and X) are. We move from a model for the error, expressed in unobservables, to 

a model that contains observables and the regression parameter by means of a standard 

“transformation of random variables” procedure. Since Yi is a linear function of ε it will 

also be normally distributed. Its density is obtained as the product of the density of ε and 

the “Jacobian” of the transformation, using i = yi – xiβ (with xi as the i-th row in the X 

matrix). As it turns out, the Jacobian is one, so that 

 

f([yi |β]i|s2)= 1

2Aσ2

e-(1/2)((yi-xiβ)

2/σ2) (B.16) 

 

The likelihood function is the joint density of all the observations, given a value for 

the parameters β and σ

2

. Since independence is assumed, this is simply the product of the 

individual densities from equation B.16. The log-likelihood is then the log of this product, 

or the sum of the logs of the individual densities. The contribution to the log likelihood of 

each observation follows from equation B.16: 

݋ܮ݃ሺ݂ሺܻ௜|ߚ, ߪଶሻ ൌ ܮ ௜ൌ െ0.5 logሺ2ߨሻ െ 0.5 logሺߪଶሻ െ 0.5ሺ௒೔ି௑೔ఉሻమ

ఙమ ሻ (B.17) 

 The full log-likelihood follows as: 

ܮ ൌ ∑ ܮ ௜ൌ െ ே

ଶ

ே

௜ୀଵ logሺ2ߨሻ െ ே

ଶ logሺߪଶሻ െ ఙమଶ ∑ ሺܻ௜ െ ܺ௜ߚ ே

௜ୀଵ ሻଶ (B.18) 
B.6 

or, in matrix notation, 

ܮൌെ ே

ଶ logሺ2ߨሻ െ ேଶ logሺߪଶሻ െ ఙమ

ଶ ሺܻ െ ܺߚሻᇱ

ሺܻ െ ܺߚሻ (B.19) 

 

A Maximum Likelihood estimator for the parameters in the model finds the values 

for β and σ

2

 that yield the highest value for equation B.19. It turns out that minimizing 

the residual sum of squares (or, least squares), the last term in equations B.18 and B.19, is 

equivalent to maximizing the log-likelihood. More formally, the solution to the 

maximization problem is found from the first-order conditions (setting the first partial 

derivatives of the log-likelihood to zero), which yield the OLS estimator for b and 

௅ெݏ

ଶ ൌ ∑ ௘೔

మ

ே

ே

௜ୀଵ (B.20) 

or, in matrix notation, 

௅ெݏ

ଶ ൌ ௘ᇲ௘

ே (B.21) 

Inference 

With estimates for the parameters in hand, the missing piece is a measure for the 

precision of these estimates, which can then be used in significance tests, such as t-tests 

and F-tests. The estimated variance-covariance matrix for the regression coefficients is 

ܸܽݎሺܾሻ ൌ ݏଶሺܺᇱܺሻିଵ (B.22) 

where s

2

 is either sOLS

2

 or sML

2

. The diagonal elements of this matrix are the variance 

terms, and their square root the standard error. Note that the estimated variance using 

sML

2 will always be smaller than that based on the use of sOLS2

. This may be spurious, 

since the ML estimates are based on asymptotic considerations (with a “conceptual” 

sample size approaching infinity), whereas the OLS estimates use a “degrees of freedom” 

(N-K) correction. In large samples, the distinction between OLS and ML disappears (for 

very large N as N and N-K will be very close). 

Typically, interest focuses on whether a particular population coefficient (the 

unknown bh) is different from zero, or, in other words, whether the matching variable 

contributes to the regression. Formally, this is a test on the null hypothesis that bh = 0. 

This leads to a t test statistic as the ratio of the estimate over its standard error (the square 

root of the h,h element in the variance-covariance matrix), or 
B.7 

೓௕ ൌ ݐ

ට௦మሺ௑ᇲ௑ሻ೓೓

షభ

 (B.23) 

This test statistic follows a Student t distribution with N-K degrees of freedom. If, 

according to this reference distribution, the probability that a value equal to or larger than 

the t-value (for a one-sided test) occurs is very small, the null hypothesis will be rejected 

and the coefficient deemed “significant.”2

 

Note that when sML

2

 is used as the estimate for s2, the t-test is referred to as an 

“asymptotic” t-test. In practice, this is a standard normal variate. Hence, instead of 

comparing the t test statistic to a Student t distribution, its probability should be evaluated 

from the standard normal density. 

A second important null hypothesis pertains to all the coefficients taken together 

(other than the intercept). This is a test on the significance of the regression as a whole, or 

a test on the null hypothesis that, jointly, bh = 0, for h=2,...,K (note that there are K-1 

hypotheses). The F test statistic for this test is constructed by comparing the residual sum 

of squares (RSS) in the regression to that obtained without a model. The latter is referred 

to as the “constrained” (i.e., with all the β except the constant term set to zero) residual 

sum of squares (RSSC). It is computed as the sum of squares of the yi in deviation from 

the mean, or 

ܴܵܵ஼ ൌ ∑ ሺܻ௜ െ ܻതሻ ே ଶ

௜ୀଵ (B.24) 

where ܻത ൌ ܻ௜/ܰ. The F statistic then follows as: 

ൌ ܨ

ሺೃೄೄ಴షೃೄೄሻ

ሺ಼షభሻ

ೃೄೄ

ሺಿష಼ሻ

 (B.25) 

It is distributed as an F-variate with K-1,N-K degrees of freedom. 

Model Fit 

The most common measure of fit of the regression is the R2

, which is closely 

related to the F-test. The R2 departs from a decomposition of the total sum of squares, or 

the RSSC from equation BError! Reference source not found., into the “explained” sum of 

 

2

 Any notion of significance is always with respect to a given p-value, or Type I error. The Type I 

error is the chance of making a wrong decision, i.e., of rejecting the null hypothesis when in fact it 

is true. 
B.8 

squares (the sum of squares of predicted values, in deviations from the mean), and the 

residual sum of squares, RSS. 

The R2

 is a measure of how much of this decomposition is due to the “model.” It 

is easily computed as:3

 

 R2=1-RSS/RSSC (B.26) 

In general, the model with the highest R2

 is considered best. However, this may 

be misleading since it is always possible to increase the R2

 by adding another explanatory 

variable, irrespective of whether this variable contributes “significantly.” The adjusted R2

(Ra

2

) provides a better guide that compensates for “over-fitting” the data by correcting for 

the number of variables included in the model. It is computed by rescaling the numerator 

and denominator in equation B.26, as 

ܴ௔

ଶ ൌ1െ

ೃೄೄ

ሺಿష಼ሻ

ೃೄೄ಴

ሺಿషభሻ

 (B.27) 

For very large data sets, this rescaling will have negligible effect and the R2 and Ra

2

 will 

be virtually the same. 

When OLS is viewed as a ML estimator, an alternative measure of fit is the value 

of the maximized log-likelihood. This is obtained by substituting the estimates bML and 

sML

2

 into expression B.18 or B.19. With e = y - XbML as the residual vector and sML

2

 = 

e'e/N, the log-likelihood can be written in a simpler form: 

ܮൌെ ே

ଶ logሺ2ߨሻ െ ேଶ log ቀ௘ᇲ௘ே ቁ െ 0.5 ௘ᇲ௘ே ሺ݁ᇱ

݁ሻ (B.28) 

ܮൌെ ே

ଶ logሺ2ߨሻ െ ேଶ െ ேଶ log	ሺ௘ᇲ௘ே ሻ (B.29) 

Note that the only term that changes with the model fit is the last one, the 

logarithm of the average residual sum of squares. Therefore, the constant part is not 

always reported. To retain comparability with other models (e.g., spatial regression 

models), it is important to be consistent in this reporting. The model with the highest

maximized log-likelihood is considered to be best, even though the likelihood, as such, is 

technically not a measure of fit. 

 

3

 When the regression specification does not contain a constant term, the value obtained for the R2

using equation 26 will be incorrect. This is because the constant term forces the residuals to have 

mean zero. Without a constant term, the RSS must be computed in the same waysameway as *in 

equation Error! Reference source not found. by subtracting the average residual ê = Σ ei/N. 
B.9 

Similar to the Ra

2

, there exist several corrections of the maximized log-likelihood 

to take into account potential over-fitting. The better-known measures are the Akaike 

Information Criterion (AIC) and the Bayesian Information Criterion (BIC)/Schwartz 

Criterion (SC), familiar in the literature on Bayesian statistics. They are easily constructed 

from the maximized log-likelihood. They are, respectively: 

 (.30B (ܭ2 ൅ ܮ2െ ൌ ܥܫܣ

 (.31B (ሻܰሺ݃݋݈ܭ ൅ ܮ2െ ൌ ܥ/ܵܥܫܤ

The model with the lowest information criterion value is considered to be best. 

Poisson Regression

 Next, the Poisson regression model is examined. 

Likelihood Function 

In the Poisson regression model, the dependent variable for observation i (with 

i=1,...,N), Yi is modeled as a Poisson random variate with a mean 8i that is specified as a 

function of a K by 1 (column) vector of explanatory variables xi, and a matching vector of 

parameters β. The probability of observing yi is expressed as: 

ܲ݋ݎܾሺܻ௜ሻ ൌ ௘షഊఒ೔௒೔

௒೔! (B.32) 

The conditional mean of yi, given observations on xi is specified as an exponential 

function of x: 

ܧሺܻ௜|ܺ௜ሻ ൌ ݈௜ ൌ ݁௑೔

ᇲ

௕ (B.33) 

where xi' is a row vector. Equivalently, this is sometimes referred to as a loglinear model, 

since 

lnሺ݈௜ሻ ൌ ܺ௜

ᇱ

ܾ (B.34) 

Note that the mean in B.33 is nonlinear, which means that the effect of a change in 

Xi will depend not only on β (as in the classical linear regression), but also on the value of 

Xi. Also, in the Poisson model, the mean equals the variance (equidispersion) so that there 

is no need to separately estimate the latter. 
B.10 

There is a fundamental difference between a classical linear regression model and 

the specification for the conditional mean in the Poisson regression model in that the latter 

does not contain a random error term (in its “pure” form). Consequently, unlike the 

approach taken for the linear regression, the log-likelihood is not derived from the joint 

density of the random errors, but from the distribution for dependent variable itself, using 

B.32. Also, there is no need to estimate a residual variance, as in the classical regression 

model. 

Assuming independence among the count variables (e.g., excluding spatial 

correlation), the log-likelihood for the Poisson regression model follows as: 

௜ܺ௜ܻ ∑ ൌ ܮ

௕ െ ݁௑೔

ᇲ

௕ െ ݈ܻ݊௜! ே

௜ୀଵ (B.35) 

Note that the third term is a constant and does not change with the parameter values. 

Some programs may not include this term in what is reported as the log-likelihood. Also, 

it is not needed in a Likelihood Ratio test, since it will cancel out. 

The first order conditions, ∂L/∂β=0, yield a system of K equations (one for each β) 

of the form: 



i=1

N

 (yi-exi'b)xi=0 (B.36) 

Note how this takes the usual form of an orthogonality condition between the “residuals” 

(yi-exi'b) and the explanatory variables, Xi. This also has the side effect that when X

contains a constant term, the sum of the predicted values, exi'b equals the sum of the 

observed counts.4

 The system B.36 is nonlinear in β and does not have an analytical 

solution. It is typically solved using the Newton-Raphson method (see below). 

 

 Once the estimates of β are obtained, they can be substituted into the loglikelihood (equation B.36) to compute the value of the maximum log-likelihood. This can 

then be inserted in the AIC and BIC information criteria in the usual way. 

 

4

 A different way of stating this property is to note that the sum of the residuals equals zero. As for 

the classical linear regression model, this is not guaranteed without a constant term in the 

regression. 
B.11 

 Predicted Values and Residuals 

The predicted value, ܻప

෡, is the conditional mean or the average number of events, 

given the Xi. This is also denoted as 8i and is typically not an integer number whereas the 

observed value Yi is a count. The use of the exponential function guarantees that the 

predicted value is non-negative. Specifically: 

పߣ

෡ ൌ ݁௑೔

ᇲ

௕෠

 (B.37) 

The “residuals” are simply the difference between observed and predicted: 

݁௜ ൌ ܻ௜ െ ݁௘೉೔

ᇲ್෡

ൌ ܻ௜ െ ߣప

෡ (B.38) 

 Note that, unlike the case for the classical regression model, these residuals are not 

needed to compute estimates for error variance (since there is no error term in the model). 

Estimation Steps 

The well known Newton-Raphson procedure proceeds iteratively. Starting from a 

set of estimates β^

t the next value is obtained as: 

β^

t+1=β^t - H^ t

-1g^

t (B.39) 

where g^

t is the first partial derivative of the log-likelihood, evaluated at β^t and H^ t is the 

Hessian, or second partial derivative, also evaluated at β^

t. 

In the Poisson regression model, 

g= 

i=1

N

 xi(yi-8i ^ ) (B.40) 

 H= - 

i=1

N

8i ^ xixi' (B.41) 

In practice, one can proceed along the following lines. 
B.12 

1. Set initial values for parameters, say b0[h], for h=1,...,K. One can set 

ܾ଴ሾ1ሿ ൌ ݕത, the overall average count as the constant term, and the other 

ܾ଴[h]=0, for h=2,...,K. 

2. Compute predicted values for each i, the value of 8i ^ =exi'b0. 

3. Compute gradient, g, using the starting values. Note that g[h] is a K by 1 

vector. Each element of this vector is the difference between: 

ܱ௜ ൌ ∑ ܺ௜௛ܻ௜

ே

௜ୀଵ (B.42) 

௜ߣ௛௜ܺ ∑ ൌ௜ ܲ

ே

௜ୀଵ (B.43) 

݃௜ ൌ ܱ௜ െ ܲ௜ (B.44) 

Note that B.42 does not contain any unknown parameters and needs only to be 

computed once (provided there is sufficient storage). As the Newton-Raphson iterations 

proceed, the values of g will become very small. 

4. Compute the Hessian, H, using the starting values. H is a K by K matrix 

(B.41) that needs to be inverted at each iteration in B.39. It is not the X'X

of the classical model, but rather more like X' Σ X, where Σ is a diagonal 

matrix. One way to implement this is to multiply each row of the X matrix 

by 8^

i, e.g., xs[i][h]=x[i][h]*sqrt(8^[i]), where xs is the new matrix (X*

), i

is the observation (row) and h the column of X. The Hessian then becomes 

the cross product of the new matrices, or, H=X*' X*

. This needs to be done 

in each iteration. There is no need to take a negative since the negative in 

B.41 and in B.39 cancel. 

5. Update the estimate for the b[h], say b1[h] is obtained using the updating 

equation B.39 except that the product H-1g is added to the initial value. In 

general, for iteration t, the new estimates are obtained as bt+1. After 

checking for convergence, the old bt is set to bt+1 and inserted in the 

computation of the predicted values, in step 2 above. 

6. Convergence. Stop the iterations when the difference between bt+1 and bt

becomes below some tolerance level. A commonly used criterion is the 

norm of the difference vector or Σh (bt+1[h] - bt[h])2

. When the norm is 

below a preset level, stop the iterations and report the last bt as the result. 
B.13 

The reason for not using bt+1 is that the latter would require an extra 

computation of the Hessian needed for inference. 

Inference 

 The asymptotic variance matrix is the inverse Hessian obtained at the last iteration 

(i.e., using bt). The variance of the estimates are the diagonal elements, the standard errors 

their square roots. The asymptotic t-test is constructed in the usual way, as the ratio of the 

estimate over its standard error. The only difference with the classic linear regression 

case is that the p-values must be looked up in a standard normal distribution, not a 

Student t distribution. 

Likelihood Ratio Test 

 A simple test on the overall fit of the model, as an analogue to the F-test in the 

classical regression model is a Likelihood Ratio test on the “slopes”. The model with only 

the intercept is nothing but the mean of the counts, or 

ߣூ ൌ ܻതœ (B.45) 

with ܻത ൌ ∑ ܻ௜/ܰ ே

௜ୀଵ . 

 

 The corresponding log-likelihood is: 

ܮோ ൌ െܻܰത ൅ lnሺܻതሻ ሺ∑ ܻ௜ሻ െ ∑ lnሺܻ௜ሻ ! ே

௜ୀଵ

ே

௜ୀଵ (B.46) 

where the R stands for the “restricted” model, as opposed to the “unrestricted” model with 

K-1 slope parameters. The last term in B.46 can be dropped, as long as it is also dropped 

in the calculation of the maximized likelihood (B.35) for the unrestricted model (LU), 

using li=exi'bt. The Likelihood Ratio test is then: 

 LR=2(LU-LR), (B.47) 

and follows a P2

 distribution with K-1 degrees of freedom. 
B.14 

References

Gentle, J. E. (1998). Numerical Linear Algebra for Applications in Statistics. SpringerVerlag, New York, NY. 

Gentleman, W. M. (1974). Algorithm AS 75: Basic procedures for large, sparse or 

weighted linear least problems. Applied Statistics, 23: 448–454. 

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