The views expressed are those of the authors, not necessarily *

those of the Comptroller of the Currency, or the Department of the

Treasury. The authors thank James Barth, Peter von zur Muehlen, Thomas

Lutton, and Larry Mote whose comments considerably improved the paper.

The authors also benefited from discussions with participants in

seminars at the Office of the Comptroller of the Currency and the

Federal Reserve Board. 

John Leusner died on March 5, 1996. We mourn his tragic death **

and will miss our good friend. 

1

Solving an Empirical Puzzle in the Capital Asset Pricing Model*

John Leusner**

Graduate School of Business

The University of Chicago, Chicago, IL 60637

Jalal D. Akhavein

Graduate Student

Department of Economics

New York University, New York, NY 10003

The Wharton Financial Institutions Center 

University of Pennsylvania, Philadelphia, PA 19104

P.A.V.B. Swamy 

Senior Financial Economist

Office of the Comptroller of the Currency, Washington, DC 20219

January 31, 1996

ABSTRACT

A long standing puzzle in the Capital Asset Pricing Model (CAPM) has been the inability of

empirical work to validate it. Roll (1977) was the first to point out this problem, and recently, Fama and French

(1992, 1993) bolstered Roll’s original critique with additional empirical results. Does this mean the CAPM is

dead? This paper presents a new empirical approach to estimating the CAPM. This approach takes into

account the differences between observable and expected returns for risky assets and for the market portfolio of

all traded assets, as well as inherent nonlinearities and the effects of excluded variables. Using this approach,

we provide evidence that the CAPM is alive and well. 

KEY WORDS: The capital asset pricing model; An empirical puzzle; Effect of mismeasuring the market

portfolio; Omitted-variables bias; Unknown functional forms; Consistency of assumptions;

Concomitants. 
$

$

2

1. Introduction

The Capital Asset Pricing Model (CAPM) of Sharpe (1964), Lintner (1965), and Black (1972) in its

various formulations provides predictions for equilibrium expected returns on risky assets. More specifically,

one of its formulations states that the expected excess return over the risk-free interest rate of an asset (or a

group of assets) equals a coefficient, denoted by , times the (mean-variance efficient) market portfolio’s

expected excess return over the risk-free interest rate. This relatively straightforward relationship between

various rates of return is difficult to implement empirically because expected returns and the efficient market

portfolio are unobservable. Despite this formidable difficulty, a substantial number of tests have nonetheless

been performed, using a variety of ex-post values and proxies for the unobservable ex-ante variables.

Recognizing the seriousness of this situation quite early, Roll (1977) emphasized correctly that tests following

such an approach provide no evidence about the validity of the CAPM. The obvious reason is that ex-post

values and proxies are only approximations and therefore not the variables one should actually be using to test

the CAPM. The primary purpose of this paper is to provide a new approach to testing the CAPM that

overcomes this deficiency. 

Recently, Fama and French (1992, 1993) conducted extensive tests of the CAPM and found that the

relation between average stock return and is flat, and that average firm size and the ratio of book-to-market

equity do a good job capturing the cross-sectional variation in average stock returns. These findings suggest,

among other things, that a formal accounting of the effects of “excluded variables” may resurrect the CAPM.

This will be the central issue in this paper. 

According to Fama and French (1993), some questions that need to be addressed are: (i) how are the

size and book-to-market factors in returns driven by the stochastic behavior of earnings? (ii) how does

profitability, or any other fundamental, produce common variation in returns associated with size and book-tomarket equity that is not picked up by the market return? (iii) can specific fundamentals be identified as state

variables that lead to common a variation in returns that is independent of the market and carries a different

premium than general market risk? This paper attempts to answer these questions. 

In an interesting article, Black (1995) gives three theoretical explanations of the measured flat line
Erit & rft ' $it(ErMt &rft),

$

$

$

Erit

ErMt rft

$it

3

(1)

relating expected return and : (i) mismeasuring the market portfolio, (ii) restricted borrowing, and (iii)

reluctance to borrow. Even if such reasoning is correct, we have found that the relation between the observed

counterparts of expected return and is nonlinear. Finally, we shall provide some answers to questions posed

by Black (1995) concerning the future prospects of the CAPM: (i) will the line be flat in the future? (ii) will it

be steep as the CAPM says it should be? and (iii) will it be flatter, but not completely flat? 

First, the CAPM is modified to take into account the differences between expected and observable

returns and between the market portfolio and its proxy. In this modified model, is not required to be a

constant, but instead is permitted to vary. Second, the effects of excluded variables and departures from a linear

functional form are taken into account. Third, all the modifications are then expressed in terms of observable

variables. Finally, the coefficients on the observable regressors are modeled as stochastic functions of the

variables that Fama and French (1992) include in their test of the CAPM and find to have reliable power in

explaining a cross-section of average stock returns. Once this has been done, the resulting model is estimated

using data for 10 stock portfolios formed on the basis of both firm size and the ratio of book-to-market equity.

(This procedure of forming portfolios was originated by Fama and French (1993).)

The specific model to be estimated is developed in Section 2. The issue regarding what constitutes a

reasonable inference based upon this model is addressed in Section 3. A brief description of the data used to

estimate the model is presented in Section 4. Section 5 discusses the empirical results and their applications.

Section 6 contains the conclusions. 

2. Interpreting and Extending the CAPM

2.1 A brief description of the model

The CAPM may be expressed as

where is the (subjective) expected return on an asset (or a group of assets) an investor chooses to hold,

 is the (subjective) expected return on the mean-variance efficient market portfolio, is the risk-free

rate, i indexes assets or groups of assets, t indexes time, and is equal to the ratio of the covariance
rit ' Erit % vit,

rMt ' ErMt % vMt,

rit & rft ' $

(

it(rMt & rft) % vit,

rit rMt cov(rit, rMt), rMt F

2

M

$i

$i

Erit ErMt cov(rit, rMt) F

2

M

$it

rit rMt vit vMt

$

(

it ' $it[1 &

vMt

rMt & rft

]

vit vMt $it

4

(2)

(3)

(4)

between and , denoted by and the variance of , denoted by . (Alternative

definitions of are provided in Ingersoll (1987, pp. 92, 124, and 134) and Constantinides (1989).) The time

variability of this variance and covariance implies that is time varying. It is important to note that as in the

case of and , both , and are the moments of a subjective distribution.

Homogeneous expectations or beliefs are not assumed here by allowing the subjective distributions to be

different for each and every investor. 

2.2 Some problems with the CAPM 

A difficulty with empirically testing whether is significantly different from zero in equation (1),

from a statistical standpoint, is that it represents a statement about expected returns, which are not observable.

To transform the relationship into observable variables for testing purposes, we introduce the following two

equations relating observable returns to expected returns:

where and are the observable returns, and and are random variables. These latter variables

will be distributed with zero means only if the data-generating processes and subjective processes of returns

possess the same means. Substituting equations (2) and (3) into equation (1) yields 

where . 

Although equation (4) is not expressed in the form of an errors-in-variables model, it reduces to such

a model if the means of and are zero and is a constant. Models possessing these properties have

been extensively studied in statistics and econometrics literature (see Lehmann (1983, pp. 450-451)). As it

turns out, estimation of equation (4) when it is not restricted as an errors-in-variables model is relatively

straightforward, as will be shown in the next section. Furthermore, the now classic “Roll’s (1977) critique” of
rit & rft ' $

(

it(rMt & rft) % vit % j

m

j'1

.ijtxijt,

vMt

vMt

vMt

rMt

ErMt vMt

$its

5

(5)

tests of the CAPM noted earlier does not apply to the estimation of equation (4) because accounts for any

differences between the “true” (unobservable) market portfolio and the particular portfolio that is chosen as a

proxy. Further, is permitted to have a non-zero and time-varying mean to cover situations where these

differences are systematic and time varying. The presence of in equation (4) makes the effects of

mismeasurements noted by Black (1995) explicit, although equation (3) indicates that is a good proxy for

 only if the mean of is zero. 

Even if the mismeasurement issue is resolved, equation (4) may nonetheless still be criticized insofar

as important regressors are excluded. For example, no asset is perfectly liquid because all trades required to

convert assets into cash involve some transaction cost. As a result, investors may choose to hold more liquid

assets with lower transaction costs than otherwise. If so, an illiquidity premium should be taken into account. 

This can be done by allowing for trading costs to enter the right-hand side of equation (4) (see Amihud and

Mendelson (1986)). Other potentially important variables, excluded from equation (4), are discussed below. 

Another issue in the CAPM is whether investors face only one risk arising from uncertainty about the

future values of assets. In all likelihood, investors face many sources of risk, as shown by Merton’s (1973)

inter-temporal asset pricing model. In such instances, investors would supplement the market portfolio with

additional positions in hedge portfolios to offset these risks. This results in separate and risk premiums

for every significant source of risk that investors try to hedge. Equation (4), therefore, should be extended to

account for the effects of extra-market hedging transactions on equilibrium rates of return. Such an expanded

version of equation (4) would recognize the multidimensional nature of risk and thereby show that some

important regressors are necessarily excluded from equation (4). 

2.3 A generalization of the CAPM 

Including previously excluded regressors in equation (4) is not trivial because the functional form of

the relationship between them and the dependent variable is unknown. This difficulty is resolved in principle by

modifying equation (4) as follows:
xijt ' R0ijt % R1ijt(rMt & rft), j ' 1,2,...,m,

xijt .ijt

rit & rft

xijt

xijt

(rMt&rft)

R0ijt xijt (rMt & rft)

(rMt & rft) xijt

R0ijt R0ijt R1ijt

xijt (rMt & rft)

R0ijt R1ijt

6

(6)

where the represent excluded variables, the denote their coefficients, and m denotes the number of

excluded variables. Since m cannot be known with certainty, one may assume without restricting it to be equal

to a specific number that the regressors of equation (5) form a sufficient set in the sense that they exactly

determine the values of in all periods. 

As a general rule, by allowing all the coefficients in a linear equation to be different for each and every

observation, the equation is permitted to pass through every data point and hence it coincides, for certain

variations in the coefficients, with the actual process generating the data on its dependent variable. Because of

this rule, equation (5) provides the only reliable way to capture unknown functional forms without relying upon

strong prior information. One may assume that the coefficients of equation (5) are constants only when this

equation is known with certainty to be linear. In contrast, with varying coefficients, equation (5) is truly

nonlinear. 

Obviously, equation (5) cannot be empirically estimated if the data on the are not available.

What is not so obvious is that when the are not observable, one cannot prove they are uncorrelated with

 (see Pratt and Schlaifer (1984)). An approach for resolving this problem is to avoid making such

uncorrelatedness assumptions and assume instead that 

where is the portion of remaining after the effect of the variable has been removed.

Accordingly, even if the variable is correlated with the , it can nonetheless be uncorrelated

with the remainders, . Also, for certain variations in and , equation (6) exactly coincides with

the true relationship between the and , if such a relationship exists. Once again, however,

one cannot assume the and are constants unless equation (6) is known with certainty to be linear.

Equation (6) should be recognized as an auxiliary equation, a linear form of which has been used to analyze the

effects of excluded variables in the econometrics literature (see Greene (1993, pp. 245-247)). Since this

equation does not impose any constraints on the coefficients of equation (5), it does not prevent the latter

equation from coinciding with the true relationship between the variables. Substituting equation (6) into

equation (5) yields 
rit & rft ' (0it % (1it(rMt & rft),

(0it ' (vit % j

m

j'1

.ijtR0ijt) (1it ' ($

(

it % j

m

j'1

.ijtR1ijt).

(1it

$it &

$itvMt

rMt & rft

j

m

j'1

.ijtR1ijt

(1it (0it

(0it (1it

(0it (1it

(1it

vMt

$it (1it (rMt & rft) $

(

it

(0it (1it .ijt

(0it (1it (rMt & rft)

.ijt

The main difference between this approach and the previous 1

approaches to estimating the CAPM lies in equations (3)-(6) which do not

appear in the latter approaches (see Jagannathan and Wang (1996) and the

previous studies by Harvey, Shanken and others referred to therein). The

linear least squares residuals introduced in these previous studies

cannot represent measurement errors and the effects of omitted variables

(see Pratt and Schlaifer (1984, p. 11-12)). 

7

(7)

where and 

The coefficient, , has a relatively straightforward economic interpretation. It consists of three

parts, the “true” beta, , of the CAPM, a mismeasurement effect, , and an omitted-variables

bias, . More so than the “true” beta, the omitted-variables bias changes over time because the set

of excluded variables undoubtedly changes quite frequently, lending further real-world, economic plausibility

to time variability of . Similarly, the connection of with the intercepts of equations (5) and (6)

clarifies its real-world origin.

The preceding discussion proposes the introduction of varying coefficients and auxiliary equations

into the estimation procedure as an important approach to dealing with unknown functional forms and the

effects of omitted variables. Equation (7) provides a useful formulation that does not suffer from various

specification errors when testing the CAPM, and it avoids such serious errors by not relying on any definitions

of and other than those provided by equation (7).1

3. Econometric Underpinnings of the Extended CAPM 

 Estimation of equation (7) requires specific stochastic assumptions about and . The

permissible set of assumptions is, however, restricted. For example, one cannot assume that is a constant

because doing so would contradict the assumption that is a random variable, even ignoring any variations

in and any omitted-variable bias. In addition, the fact that depends on via , and

 and are functions of the common set of time-varying coefficients prohibits one from assuming

that the variables , , and are uncorrelated with one another. (Remember that the

nonlinearities involved in equation (5) cannot be captured without the time-varying .) In other words, one
F

2

i ' j

n

j'1 j

n

k'1

jÖk

wijtwiktcov(rijt, rikt) % j

n

j'1

w

2

ijtvar(rijt),

F

2

i ' var((1it(rMt & rft)) % var((0it) % 2cov((0it, (1it(rMt & rft)).

E[(rit & rft)|(rMt & rft)] ' " % $(rMt & rft) " $

(0it (1it

(0it (1it

(0it (1it

rit

rit ' j

n

j'1

wijtrijt rijt

rijt rft wijt

rit

var(rijt) rijt cov(rijt, rikt)

rijt rikt.

rit

F

2

i

(rMt & rft)

8

(8)

(9)

cannot assume that , where and are constants,

without contradicting the definitions of and . This argument lies at the heart of Roll’s (1977)

criticism of earlier tests of the CAPM. In principle, generalizing the set of assumptions about and 

can help in this respect. The reason is that general assumptions are more likely to encompass true assumptions

as special cases than more restrictive assumptions. We shall proceed therefore by weakening the assumptions

about and . 

Suppose that refers to the rate of return on the i-th group or portfolio of assets. Then

, where is the rate of return on the j-th asset included in the i-th portfolio, one of the

 is equal to , and is the proportion of the investor’s total budgets allocated to the j-th asset. It

follows that the variance of is 

where denotes the variance of and denotes the covariance between

 and 

The variance of can also be obtained directly from equation (7). It is 

Equation (9) eases the computational burden compared to equation (8) because the latter equation

involves a large number of variances and covariances that may be time varying and cannot be estimated unless

one knows how they vary over time even if the data on all n securities in equation (8) are available. However, it

does not have the advantage of parsimony (in terms of a preference for a model with fewer parameters and in

all other respects almost as good as other competing models) if these two equations yield different values for

the same variance . One should, therefore, consider the conditions under which the same values would be

obtained. Suppose that the third term on the right-hand side of equation (9) is zero. Suppose also that the

covariances and variances given in equation (8) are attributable to and other specific variables,
zi6,t&1

(

it

' Azit % ,

it

,

,

it

' M,

i,t&1

% ait,

n 6 4,

(

it

((0it, (1it)

) A [Bkj]

zit

(1, zi1,t&1, zi2,t&1, zi3,t&1, zi4,t&1, zi5,t&1, zi6,t&1)

)

zi1,t&1

zi2,t&1

zi3,t&1

zi4,t&1

zi5,t&1

zi6,t&1,

it

(,0it, ,1it)

)

The variable proxies for the January effect which is 2

explained in Bodie, Kane and Marcus (1993, pp. 380-381). 

9

(10)

(11)

respectively. In this case, the first and second terms on the right-hand side of equation (9) can be equal to those

on the right-hand side of equation (8), respectively. If, in addition, the first and second terms on the right-hand

side of equation (8) tend to nonzero and zero, respectively, as then the first and second terms on the

right-hand side of equation (9) can be defined to be the systematic (or nondiversifiable) and nonsystematic (or

diversifiable) risk components of the portfolio variance, respectively (see Swamy, Lutton and Tavlas (1995)).

These definitions are more comprehensive than the corresponding definitions found in the finance literature

because, as has been shown above, equation (7) captures all sources of risk whereas equation (1) captures only

one such source. 

Using a specific model for testing the CAPM, following Fama and French (1993), the third term on

the right-hand side of equation (9) can be zero under the following general Assumptions I and II:

Assumption I. The coefficients of equation (7) satisfy the stochastic equation 

where denotes the two element column vector ; denotes the 2×7 matrix , k = 0, 1,

j = 0, 1, ..., 6; denotes the seven element column vector

,

 = the log of average size over all firms in the i-th portfolio (a firm’s size is equal to its market equity,

ME = a stock’s price times shares outstanding, for June of year t-1), 

 = the average of book-to-market ratio over all firms in the i-th portfolio ( a firm’s book-to-market ratio

is equal to its book equity, BE = book value of its common equity as measured by Fama and French (1993, p.

11), for the fiscal year ending in calendar year t-1, divided by its market equity, ME, in December of t-1), 

 = the dividend price ratio (dividend/price) for the S&P 500,

 = the default premium (Moody’s Baa bond rate minus Moody’s Aaa bond rate),

 = the yield on the 10-year Treasury bill minus the 1-year Treasury bill rate, 

 = a dummy variable that is 1 in January and 0 in other months , and denotes the two element 2

column vector that 

satisfies the stochastic difference equation 
|N11| ' 0 or 1, *01a

' *10a' *11a' 0, Bkj Ö 0 for k ' 0, 1 and j Ö 0.

M diag(N00, N11) &1 < N00, N11 < 1

ait ' (a0it, a1it)

) F

2

a)a

' F

2

a[*kja],

vit vMt

zit

zit

zit

(rMt & rft) ,

it

zit

(rMt & rft) (0it

zit

(1it

M )a,1it

,0it zit (rMt & rft)

zit

The conjunction of Assumptions I and II and restrictions (12) is 3

weaker than the assumptions given in Greene (1993, pp. 375-9) for the

consistency of instrumental variable estimators (see Pratt and Schlaifer

(1988, pp. 47-48)). 

10

(12)

where denotes the 2×2 diagonal matrix with and

 is distributed with mean zero and variance-covariance matrix k =

0, 1, j = 0, 1.

Note that Assumption I permits the and of equations (2) and (3) to have nonzero and timevarying means. Fama and French (1992, 1993) found that the current values of have reliable power to

explain the cross-section of average returns, even though their chosen variables do not appear directly in the

CAPM. They also found that stock risks are multidimensional and the elements of proxy for different

dimensions of risk. As discussed earlier, the coefficients of equation (7) capture the multidimensional nature of

risk. For all these reasons, equation (10), relating the coefficients of equation (7) to is an appropriate

specification. 

Assumption II. The are independent of the given a value of .

This assumption is weaker than the assumption that the are independent of the (see Dawid

(1979, p. 5)) and both of Assumptions I and II are weaker than the assumptions made by Jagannathan and

Wang, Harvey, and others in testing the CAPM. 

The variables, denoted by , are called ‘concomitants’ in Pratt and Schlaifer (1988) and form a

sufficient set of regressors for equation (10) if they completely explain all the variation in . Algebraically,

this condition can be expressed as

The implication of these conditions on and is that the distribution of is degenerate. If it is

degenerate and if the expectation of , given the values of and , is zero, as implied by

Assumptions I and II, then the third term on the right-hand side of equation (9) is zero given a value of ,

and model (7) reduces to a regression model with first-order autoregressive errors. In this case, the usual

consistency proofs apply to Swamy, Mehta and Singamsetti’s (1995) parameter estimators of model (7).3
E $it&$it

vMt

(rMt&rft)

%j

m

j'1

.ijtR1ijt *zit ' B10%j

6

j'1

B1jzij,t&1%E(,1it|zit) .

Bkj ' 0 for k ' 0, 1 and j Ö 0.

E($it|zit) E($it|zit) ' B10

$it zit

E($it|zit) ' B10

B10 %

1

T j

T

t'1 j

6

j'1

B1jzij,t&1B10

E(

1

T j

T

t'1

&$it

vMt

(rMt&rft)

|zit)

1

T j

T

t'1

E(,1it|zit)

1

T j

T

t'1

(1it B10 %

1

T j

T

t'1 j

6

j'1

B1jzij,t&1

1

T j

T

t'1

(1it

E(

1

T j

T

t'1

(1it|zit) E(

1

T j

T

t'1

&$it

vMt

(rMt&rft)

|zit)

1

T j

T

t'1

E(,1it|zit)

T 6 4

M )a

M )a

Equation (7) is estimated under Assumptions I and II, using a 4

computer program developed by I-Lok Chang and Stephen Taubman. This

program uses an algorithm developed by Chang, Hallahan and Swamy (1992)

and is based on a methodology introduced by Swamy and Tinsley (1980).

For further discussion of this methodology, see Swamy, Mehta and

Singamsetti (1996). 

11

(13)

(14)

Under Assumptions I and II, 

If the sum of the second and third terms on the left-hand side of this equation is equal to the sum of the second

and third terms on its right-hand side and is a constant, then . Thus,

Assumptions I and II can aid in the estimation of the conditional mean of in the original CAPM, given .

Even when , 

the conditional mean is preferable to as a measure of the risks inherent in the

i-th asset (or group of assets) if is equal to , since

equation (5), unlike equation (1), covers all sources of risk. For large T, the mean 

 will be equal to if converges in probability to

 and if both and tend to zero as

.

Estimation of equation (7) under Assumptions I and II is performed with and without the restrictions

that4

These restrictions, when imposed, eliminate the time-varying z’s from equation (10). 

Note that Assumption I does not permit the restrictions on and given in (12) to be exactly

satisfied. They can only be nearly satisfied if the estimates of and are equal to the boundary values
(

it

' Azit % μi% ,

it

μi' (μ0i, μ1i)

)

rit&rft ' [1 (rMt&rft)]

B00 B01  ̨ B06

B10 B11  ̨ B16

1

zi1,t&1

!

zi6,t&1

% [1 (rMt&rft)]

,0it

,1it

.

|N11| ' .99, *01a

' *10a' *11a' .0001

A

rit & rft

rMt & rft

$

zit (rMt & rit)

If time-series-cross-section data are used to estimate equation 5

(7), then equation (10) may be changed to , where

 is a constant through time; it is an attribute of the ith asset (or group of assets) which is unaccounted for by the included 

12

(15)

. If these restrictions are nearly satisfied when the

restrictions on given in (14) are not imposed than when they are imposed, then one can conclude that the

regressors included in equation (10) are appropriate in the sense that they adequately explain the variation in

the coefficients of equation (7). 

Equations (7) and (10) jointly describe the time-series model being estimated to explain the excess

returns in Section 5 below. Thus, substituting equation (10) into equation (7):

This equation is not linear and contains an error term that is both heteroscedastic and serially correlated. The

explanatory variables in this equation are the excess market return , the six concomitants, z,

(introduced in Assumption I that includes firm size and the ratio of book-to-market equity), and the interactions

between the excess market return and each of the concomitants. Used alone without these interactions, the

concomitants may not have adequate power to explain stock returns because of the multidimensional nature of

stock risks. Previous tests of the CAPM neglect to consider these interactions. They also use two-pass

regressions (see Bodie, Kane and Marcus (1993, Chapter 11) for a survey of these tests). In principle, applying

one-pass regression to (15) is superior to two-pass regressions, even when the second-pass regression

overcomes the measurement error problem created by the estimates. Under Assumptions I and II, the

conditional mean of the dependent variable of equation (7), given the values of and , is equal

to the first term on the right-hand side of equation (15). For empirical estimation of equation (15), any one of

three data sets (time-series data, cross-section data, and time-series-cross-section data) may be used, although

Assumptions I and II are well suited only to time-series data (see Swamy and Tavlas (1995)).5
μi

(B00, B10)

) B00

B10

Bkj ' 0 for k ' 0, 1 and j Ö 0, N11 ' 0, *01a' *10a' *11a' 0, *00a' 1,

Bkj ' 0 for k ' 0, 1 and j Ö 0, M ' 0, *01a' *10a' *11a' 0, *00a' 1.

rit & rft

B01zi1,t&1% B02zi2,t&1rit & rft

,

it

rit & rft

variables but varies across i. When equation (7) is estimated separately

for different i, the vector gets absorbed into . So and

 in equation (15) are implicitly allowed to vary across i. 

13

(16)

(17)

Now suppose that Assumptions I and II and the restrictions given by (12) hold. It follows that: (i)

equation (15) explains how firm size and the ratio of book-to-market equity influence the excess returns,

, which, in turn, influence the stochastic behavior of earnings, (ii) the sum

 measures variation in excess returns, , associated with firm size and

BE/ME that is not captured by the market return, and (iii) the can be identified as state variables that lead

to common variation in the excess returns, , that is independent of the market and thus carries a

different premium than general market risk. Note that (i)-(iii) are directly responsive to issues (i)-(iii) raised by

Fama and French (1993) and restated in the Introduction. 

It is useful here to consider variations of the model proposed above. Clearly, the conjunction of the

model given by equation (7) and Assumptions I and II is false if it cannot explain and predict the underlying

phenomenon better than the following inconsistent or restrictive alternatives introduced earlier, 

or,

Restriction (16) implies that equation (7) is a fixed-coefficients model with first-order autoregressive (AR(1))

errors, while Restriction (17) implies that equation (7) is a fixed-coefficients model with white-noise errors. 

It is useful to digress for a moment to the subject of model validation based on forecast comparisons.

A rationale for this type of comparison is provided by the cross-validation approach--which consists of splitting

the data sample into two subsamples. The choice of a model, including any necessary estimation, is then based

on one subsample and its performance is assessed by measuring its prediction against the other subsample. The

premise of this approach is that the validity of statistical estimates should be judged by data different from those

used to derive the estimates (see Mosteller and Tukey (1977, pp. 36-40)). Friedman and Schwartz (1991, p.
(i) root mean&square error ' RMSE '

1

F j

F

s'1

(Rˆ

i,T%s

& Ri,T%s)

2

Rit ' rit & rft.

Rit

Rˆ

it

Rˆ

it

14

(18)

47) also indicate that a persuasive test of a model must be based on data not used in its estimation.

Furthermore, formal hypothesis tests of a model on the data that are used to choose its numerical coefficients

are almost certain to overestimate performance: the use of statistical tests leads to false models with probability

1 if both the null and alternative hypotheses considered for these tests are false, as shown by Swamy and Tavlas

(1995, p. 171 and footnote 7). That this is a problem in the present case follows from the lack of any guarantee

that either a null or an alternative hypothesis will be true if the inconsistent restrictions (14) or (16) or (17) are

necessary parts of the maintained hypothesis. Conversely, a hypothesis is true if it is broad enough to cover the

true model as a special case. This is the motivation for extending the CAPM: to make it broad enough so that

there is a better chance of encompassing the true model as a special case. Accordingly, in Section 5 below,

model (7) and Assumptions I and II with or without restrictions (14) or (16) or (17) are evaluated based upon

forecast comparisons. 

Model (7), Assumptions I and II, and the three sets of restrictions can be combined as conjunctions,

listed here in decreasing order of generality regarding the restrictiveness of assumptions. 

Conjunction I: model (7), Assumptions I and II. 

Conjunction II: model (7), Assumptions I and II, and set (14).

Conjunction III: model (7), Assumptions I and II, and set (16).

Conjunction IV: model (7), Assumptions I and II, and set (17). 

The reason for considering Conjunctions II-IV--even though they are inconsistent--is to examine how they

perform in explanation and prediction relative to Conjunction I. Doing this is especially useful for

understanding earlier empirical work leading to the CAPM puzzle. 

The accuracy of the model, or its validity, is determined as follows. Let After

estimating the models defined by Conjunctions I-IV, forecasts of the out-of-sample values of are generated

from each of the estimated models. Let these forecasts be denoted by . Then two formulas are used to

measure the accuracy of ,
(ii) mean absolute error ' MAE '

1

F j

F

s'1

|Rˆ

i,T%s

& Ri,T%s|,

rit

rft rMt

rft rMt

zi1,t&1zi2,t&1

zi3,t&1, zi4,t&1, zi5,t&1

15

(19)

and 

where F is the number of periods being forecasted and T is the terminal date of the estimation period. 

4. Data

The are the monthly value-weighted stock returns on each of 10 portfolios that are formed

following Fama and French’s (1993, p. 11) procedure: “Each year t from 1963 to ... [1993] NYSE quintile

breakpoints for size (ME, ...), measured at the end of June, are used to allocate NYSE, Amex, and NASDAQ

stocks to five size quintiles. Similarly, NYSE quintile breakpoints for BE/ME are used to allocate NYSE,

Amex, and NASDAQ stocks to five book-to-market equity quintiles.” The 10 portfolios are formed as the

intersections of the five-firm size and the lowest- and highest-BE/ME quintiles, denoted nsibje, i = 1, 2, ..., 5

and j = 1, 5. For example, the ns1b1e portfolio contains the stocks in the smallest-ME quintile that are also in

the lowest-BE/ME quintile, and the ns5b5e portfolio contains the biggest-ME stocks that also have the highest

values of BE/ME.

The proxies for and are the same as those employed by Fama and French (1993). That is,

 = the one-month Treasury bill rate, observed at the beginning of the month, and = the value-weighted

monthly percent return on the stocks in their 25 size-BE/ME portfolios, plus the negative-BE stocks excluded

from the portfolios. 

The sources of the data employed here on the and are explained in Fama and French

(1992). The variables and were obtained from the FAME data base maintained by

the Board of Governors of the Federal Reserve System. The index t denotes the months that occurred in the

period from July 1963 through December 1993. The subscript i of variables in equations (7), (10), and (11)

should not be confused with the i of nsibje. 

5. Empirical Results and Applications 
ˆ(0it (0it

ˆ(1it (1it

(1it

$

ˆ(1it

ˆ(0it

ˆ(1it ˆ(0it ˆ(1it

$

(rMt & rft)

B00

Bˆ00

B01

B03

B06

16

Tables 1a-c show the estimates for Conjunction I. The maxima, minima, and ranges of the estimates

( ) of in Table 1a show considerable variation over time for all 10 portfolios. By contrast, for the

same portfolios, the volatilities of the estimates ( ) of are quite low. When estimated without the

nonnegativity constraint, the estimates of were negative only for 5 of 341 months and for 1 (ns1b5e) of 10

portfolios. 

Black (1995) provides three theoretical reasons why the line relating expected return and is flatter

than suggested by the CAPM, as stated in the Introduction. The question arises then, do the values in Tables

1a-c support such a flat line? The answer is “no,” as is shown below. Under the conditions stated below

equation (13), the arithmetic means of in Table 1a give measures of portfolio risks. These arithmetic

means are positive for all 10 portfolios and are significantly different from zero for 7 of these portfolios. The

arithmetic means of are not significantly different from zero for these 7 portfolios and are significantly

different from zero for the 3 highest-BE/ME portfolios (ns2b5e, ns3b5e, and ns4b5e) for which the arithmetic

means of are insignificant. The significant means of and the insignificant means of for the 3

portfolios cannot be interpreted as evidence of a flat expected return- line because they arise as a direct

consequence of the significant estimates of some of the coefficients on z’s and on the interactions between

 and each of the z’s that are discussed, in detail, below. 

Table 1b shows that the estimates of the intercept ( ) of equation (15) are insignificant for all 10

portfolios. What does this finding say about Conjunction I? The answer follows from Merton’s (1973) work,

revealing that a well-specified asset-pricing model produces intercepts that are indistinguishable from 0. Fama

and French (1993, p. 5) also state that “judging asset-pricing models on the basis of the intercepts in excessreturn regressions imposes a stringent standard.” The insignificance of the in Table 1b shows that, at least

in the cases considered here, Conjunction I shares a property with a well-specified asset-pricing model and

meets Fama and French’s stringent standard. 

Table 1b also shows that the estimates of the coefficient ( ) on log(size) are significant for 2

portfolios (ns2b1e and ns2b5e), the estimates of the coefficient ( ) on the dividend-price ratio of the S & P

500 are significant for 4 portfolios (ns2b5e, ns3b5e, ns5b1e, and ns5b5e), and the estimates of the coefficient

( ) on the January dummy variable are significant for 7 portfolios (ns1b1e, ns1b5e, ns2b5e, ns3b5e, ns4b5e,
zi6,t&1

zi2,t&1, zi4,t&1, zi5,t&1

B10 (rMt & rft)

(rMt & rft)

zi4,t&1

B10 E($it|zit)

ˆ(1it ˆB10

(rMt & rft)(log(size)

(rMt & rft)((BE/ME)

(rMt & rft)((default premium)

(rMt & rft)(zi5,t&1B13 B16

$

B10 % μ1i

zit

(rMt & rft) zit

a1it

a1it a0it

17

ns5b1e, and ns5b5e). This result shows, among other things, that except for 3 portfolios (ns2b1e, ns3b1e, and

ns4b1e), the strong January seasonals in the returns on 7 stock portfolios are not absorbed by strong seasonals

in the explanatory variables of equation (15) other than the January dummy variable ( ). The estimates

of the coefficients on and are insignificant for all 10 portfolios. 

Furthermore, the estimates of the coefficient ( ) on are significant for 7 portfolios

(ns1b1e, ns1b5e, ns2b1e, ns3b1e, ns4b1e, ns5b1e, and ns5b5e)(see Table 1b). For the remaining 3 portfolios

(ns2b5e, ns3b5e, and ns4b5e), the estimated coefficients on the interaction between and the

default premium ( ) are significant (see Table 1b). Under the conditions stated below equation (13), the

estimates of can be viewed as the estimates of and the difference between the arithmetic

mean of in Table 1a and in Table 1b gives an estimate of the arithmetic mean of the sum of

mismeasurement effects and omitted-variables biases. These differences do not appear to be insignificant for

most of the 10 portfolios. 

The interactions, the estimates of whose coefficients are significant, are 

for 3 portfolios (ns1b1e, ns3b1e, and ns5b5e), for the portfolio ns5b5e,

 for 4 portfolios (ns2b5e, ns3b5e, ns4b5e, and ns5b5e), and

 for the portfolio ns5b1e. The estimates of the interaction coefficients and are

insignificant for all 10 portfolios. 

All the estimates in Tables 1a and 1b unambiguously support only one conclusion: the relation

between the observable counterparts of expected return and is not a flat line but is nonlinear. This finding

provides possible answers to Black’s (1995) questions stated in the Introduction. The cross-section of average

returns on U.S. common stocks probably shows little relation to the defined in footnote 3 and

shows significant relation to the , as implied by Fama and French’s (1992) results, if the interactions

between and each of the , and the heteroscedasticity and serial correlation of the error term

in equation (15) are neglected. 

Table 1c indicates the extent to which Conjunction I satisfies set (12) of restrictions. In 7 of 10 cases

shown in this table, the whole set is nearly satisfied. In addition, in all 10 cases, the estimated variance of 

and the estimated covariance between and are very small in magnitude relative to the estimated
a0it

(1it

(1it

ˆ(0it ˆ(1it

ˆB00 Bˆ10

ˆ(0it (ˆ1it

B

B00

B00 B10

$

(1it

(1it

ˆB00 ˆB10 ˆB00 ˆB10

18

variance of . It should be noted that exclusion of a concomitant variable from equation (10) because its

estimated coefficient is insignificant is improper if the variable is needed to explain the variation in . It is

better to include a concomitant that substantially explains the variation in than to exclude it even if its

inclusion means reducing the t ratios of the estimates of the coefficients of equation (10). 

Tables 2ab and 2c display the results for Conjunction II. These results provide information on the

effects of set (14) of restrictions on the estimates in Tables 1a-c. The values in every column of Table 2ab can

be compared with those in the corresponding column of Table 1a. Plots (not included here) show that while the

time profiles of in Tables 2ab and 1a are the same in all 10 cases, those of in these tables are the

same only in 4 of 10 cases. The t ratios of and in Table 2ab are generally higher in magnitude than

those of the arithmetic means of and in Table 1a. It is possible that the extra precision obtained by

imposing set (14) of restrictions is spurious because several of the estimates of in Table 1b are significant. In

4 of 10 cases, the estimates of in Table 2ab are significant. This shows that Conjunction II does not always

satisfy the property of a well-specified asset-pricing model noted by Merton (1973). In 3 of 10 cases shown in

Table 2ab, the estimates of and are significant and insignificant, respectively, supporting the

conclusion that a relation between expected return and is flat. However, such a conclusion is not credible

because it is based on Conjunction II, which is inconsistent. 

Tables 1c and 2c might be compared to determine whether the regressors of equation (10) are

appropriate and sufficient. It can be seen from these tables that for 8 of the 10 portfolios, set (12) of restrictions

is better satisfied when set (14) of restrictions is not imposed than when it is imposed. This result supports the

conclusion that the regressors of equation (10) are the appropriate explanatory variables for but some

additional concomitants are needed to completely explain all the variation in for all portfolios. Further

work is needed to find such additional concomitants. 

Tables 3 and 4 show parameter estimates for Conjunctions III and IV, respectively. The t ratios of

 and in these tables are generally higher than those of and in Table 2ab in absolute value. The

spuriousness of the extra precision obtained by imposing inconsistent sets (16) and (17) of restrictions is more

pronounced than than that of the extra precision obtained by imposing inconsistent set (14) of restrictions.

Since the explanatory variables of equation (15) are not orthogonal to one another, and its error covariance
ˆB00

ˆB00 N00

,0it ˆF

2

a

ˆ*00a

a0it ˆF

2

a

ˆ*00a

ˆF

2

a

a0it

(0it (rMt & rft)

19

matrix is not equal to a scalar times an identity matrix under all Conjunctions I-IV, the estimates in Tables

2ab, 3, and 4 are not comparable to the estimates in Table 1b. The estimates of in Tables 1c, 2c, and 3

offer little support for the presence of serial correlation among the . The estimates ( ) of the

variance of in Table 1c are generally smaller than the estimates in Table 2c and the estimates 

in Tables 3 and 4, indicating that in most cases the variance of is reduced as the z’s are added to equation

(10). This reduction in variance helps to weaken the correlations between and . 

The values of log likelihood in Tables 1a, 2ab, 3, and 4 might be compared to determine whether one

of the Conjunctions I-IV has greater support of the time-series data used for estimation than other

Conjunctions. With one exception, these values in Table 1a are higher than those in Tables 2ab, 3, and 4. The

exception corresponds to the portfolio ns2b1e in Table 2ab. Even in this case, the value of the log likelihood in

Table 1a is only slightly smaller than that in Table 2ab. This shows that the support of the data to Conjunction I

is either greater or only slightly less than the support to Conjunctions II-IV. 

Since the t ratios in Tables 1a and 1b are based on a consistent set of general assumptions that are not

used by earlier tests of the CAPM in the literature, they do not fall into Black’s (1995, p. 2) category of “the

simplest kind of data mining.” Still, it is appropriate to seek RMSE and MAE measures for each portfolio of a

conjunction’s success in predicting the out-of-sample values of the dependent variable of equation (7). Tables

1c, 2c, 3, and 4 report for each portfolio the values of such measures in the RMSE and MAE columns. The

RMSEs for Conjunction I are smaller for 7 portfolios and slightly higher for 3 portfolios than those for

Conjunctions II-IV. For a conjunction which has at least 10 more unknown parameters and hence uses up at

least 10 more degrees of freedom than any of Conjunctions II-IV, this is not a bad performance. Perhaps

Conjunction I would have produced lower RMSEs than Conjunctions II-IV for all 10 portfolios if all the

parameters of equations (11) and (15) were known. Based on the RMSEs, Conjunctions II-IV cannot be

preferred to Conjunction I. Even though the MAEs for Conjunction I are smaller than those for Conjunctions IIIV in 5 of 10 cases, they are much bigger than those for Conjunctions II-IV in 2 of the remaining 5 cases. Two

reasons for this result are: (i) the predictor used to generate forecasts of the dependent variable of equation (7)

is optimal relative to a quadratic loss function but is not optimal relative to an absolute error loss function, and

(ii) sometimes inconsistent models appear to predict better than consistent models if inappropriate formulas are
(1it B10

E((1it|zit) E((1it)

(1it

(ˆ1it

$

20

used to measure the accuracy of forecasts. In the long run, only consistent models are able to tell the truth. 

Any empirical result that holds for all 4 Conjunctions can be considered robust. The arithmetic means

of the estimates of under Conjunction I and the estimates of under Conjunctions II-IV are the

estimates of the conditional and unconditional expectations, and , respectively. It can

be seen from Tables 1a, 2ab, 3, and 4 that these estimates are close to one another and hence are robust. 

Fama and French (1993, p. 53) list four applications--(a) selecting portfolios, (b) evaluating portfolio

performance, (c) measuring abnormal returns in event studies, and (d) estimating the cost of capital--that

require estimates of risk-adjusted stock returns. The estimates in Table 1b can be substituted into the first term

on the right-hand side of equation (15) to obtain the estimates of risk-adjusted stock returns because is a

comprehensive descriptor of stock risk. The preceding discussion shows that these estimates do a better job in

all four applications than those of previous studies. The methodology used to obtain the estimates in Tables 1ac can also be used to obtain accurate predictions about as yet unobserved values of the dependent variable of

equation (15). The discussion in this section and in Sections 2 and 3 shows that the measures of market or

“systematic” risks of portfolios given by the arithmetic means of in Table 1a are theoretically and

empirically superior to estimates of presented so far in the literature. 

6. Conclusions

This paper has extended the CAPM to account for the effects of differences between unobservable

and observable stock and market portfolio returns, of excluded variables, and of departures from a linear

relationship between the observable returns on individual stock and market portfolios. The extended CAPM is

tested using a stochastic-coefficients methodology. For purposes of comparison, both consistent and

inconsistent sets of assumptions are made in these tests. Tests based on a consistent set of assumptions show

that the relation between the observable returns on stock and market portfolios is nonlinear. 
21

REFERENCES

Amihud, Yakov and Haim Mendelson, 1986, Asset pricing and the bid-ask spread, Journal of Financial

Economics 17, 223-249.

Black, Fischer, 1972, Capital market equilibrium with restricted borrowing, Journal of Business 45, 444-455.

 , 1995, Estimating expected return, Journal of Financial Education 21, 1-4. 

Bodie, Zvi, Alex Kane and Alan J. Marcus, 1993, Investments 2nd ed. (Irwin, Boston, Massachusetts). 

Chang, I-Lok, Charles Hallahan and P.A.V.B. Swamy, 1992, Efficient computation of stochastic coefficient

models, in Hans M. Amman, David A. Belsley and Louis F. Pau, eds.: Computational Economics and

Econometrics (Kluwer Academic Publishers, Boston). 

Constantinides, George M., 1989, Theory of valuation: Overview and recent developments, in Sudipto

Bhattacharya & George M. Constantinides, eds.: Frontiers of Modern Financial Theory, Vol. 1 

(Rowman & Littlefield, Savage, Maryland). 

Dawid, A.P., 1979, Conditional independence in statistical theory, Journal of the Royal Statistical Society

Series B, 41, 1-31.

Fama, Eugene F. and Kenneth R. French, 1992, The cross-section of expected stock returns, The Journal of

Finance 47, 427-465. 

 , and Kenneth R. French, 1993, Common risk factors in the returns on stocks and bonds, Journal of

Financial Economics 33, 3-56. 

Friedman, Milton and Anna J. Schwartz, 1991, Alternative approaches to analyzing economic data, The

American Economic Review 81, 39-49. 

Greene, William H., 1993, Econometrics Analysis (Macmillan Publishing Company, New York). 

Ingersoll, Jr., Jonathan E., 1987, Theory of Financial Decision Making (Rowman & Littlefield, Savage,

Maryland). 

Jagannathan, Ravi and Zhenyu Wang, 1996, The conditional CAPM and the cross-section of expected returns,

The Journal of Finance 51,3-53.

Lehmann, Erich L., 1983, Theory of Point Estimation (Wiley, New York). 
22

Lintner, John, 1965, The valuation of risk assets and the selection of risky investments in stock portfolios and

capital budgets, Review of Economics and Statistics 47, 13-37.

Merton, Robert C., 1973, An intertemporal capital asset pricing model, Econometrica 41, 867-887. 

Mosteller, Frederick and John W. Tukey, 1977, Data Analysis and Regression Addison-Wesley, Reading, ( 

Massachusetts). 

Pratt, John W. and Robert Schlaifer, 1984, On the nature and discovery of structure, Journal of the American

Statistical Association 79, 9-21, 29-33. 

 , and Robert Schlaifer, 1988, On the interpretation and observation of laws, Journal of Econometrics,

Annals 39, 23-52. 

Roll, Richard, 1977, A critique of the asset pricing theory’s tests: Part I: On past and potential testability of the

theory, Journal of Financial Economics 4, 129-176.

Sharpe, William F., 1964, Capital asset prices: a theory of market equi- librium under conditions of risk,

Journal of Finance 19, 425-442.

Swamy, P.A.V.B. and Peter A. Tinsley, 1980, Linear prediction and estimation methods for regression models

with stationary stochastic coefficients, Journal of Econometrics 12, 103-142. 

 , and George S. Tavlas, 1995, Random coefficient models: Theory and applications, Journal of Economic

Surveys 9, 165-196. 

 , Jatinder S. Mehta and Rao N. Singamsetti, 1996, Circumstances in which different criteria of estimation

can be applied to estimate policy effects, Journal of Statistical Planning and Inference 50, 121-153. 

 , Thomas J. Lutton and George S. Tavlas, 1995, How should diversifiable and nondiversifiable portfolio

risks be defined?, Journal of Economics and Finance, 1997, forthcoming. 

 
23