Copyright
Lars Peter Hansen.

Econometric evaluation of asset pricing models online

. (page 1 of 4)
Online LibraryLars Peter HansenEconometric evaluation of asset pricing models → online text (page 1 of 4)
Font size
QR-code for this ebook


HD28

.M414



WORKING PAPER
ALFRED P. SLOAN SCHOOL OF MANAGEMENT



ECONOMETRIC EVALUATION OF
ASSET PRICING MODELS

by

Lars Peter Hansen
John Heaton
Erzo Luttmer



WP# 3606-93



August 1993



MASSACHUSETTS

INSTITUTE OF TECHNOLOGY

50 MEMORIAL DRIVE

CAMBRIDGE, MASSACHUSETTS 02139



ECONOMETRIC EVALUATION OF
ASSET PRICING MODELS

by

Lars Peter Hansen
John Heaton
Erzo Luttmer

WP# 3606-93 August 1993



M.I.T. LIBRARIES

OCT 1 1993
RECEIVED



Econometric Evaluation of Asset Pricing Models



August 1993



Lars Peter Hansen
University of Chicago, NBER and NORC



John Heaton
M.I. I. and NBER



Erzo Luttmer
Northwestern University



We thank Craig Burnside, Bo Honore, Andrew Lo, Marc Roston, Whitney Newey,
Jose Scheinkman, Jean-Luc Vila, Jiang Wang, Amir Yaron and especially Ravi
Jagannathan for helpful comments and discussions. We also received valuable
remarks from seminar participants at the 1992 meetings of the Society of
Economic Dynamics and Control, the 1992 NBER Summer Institute and at
Cornell, Duke, L.S.E. and Waterloo Universities. Finally, we gratefully
acknowledge the financial assistance of the International Financial Services
Research Center at M.I.T. (Heaton) and the National Science Foundation
(Hansen) and the Sloan Foundation (Luttmer).



Abstract

In this paper we provide econometric tools for the evaluation of
intertemporal asset pricing models using specification-error and volatility
bounds. We formulate analog estimators of these bounds, give conditions for
consistency and derive the limiting distribution of these estimators. The
analysis incorporates market frictions such as short-sale constraints and
proportional transactions costs. Among several applications we show how to
use the methods to assess specific asset pricing models and to provide
nonparametric characterizations of asset pricing anomalies.



I. Introduction

Frictionless market models of asset pricing imply that asset prices can
be represented by a stochastic discount factor or pricing kernel. For
example, in the Capital Asset Pricing Model (CAPM) the discount factor is
given by a constant plus a scale multiple of the return on the market
portfolio. In the Consumption-Based CAPM (CCAPM) the discount factor is
given by the intertemporal marginal rate of substitution of an investor. If
r is the net return on an asset and m is the marginal rate of substitution,
then the CCAPM implies that:

(l.n 1 = £[m(l+r)|^]

where ? is the information set of the investor today. More generally, if m
is the stochastic discount factor, today's price, nip), of an asset payoff,
p, tomorrow is given by:

(1.2) n(p) = £(mp|^) .

Thus a stochastic discount factor m "discounts" payoffs in each state of the
world and, as a consequence, adjusts the price according to the riskiness of
the payoff. From the vantage point of an empirical analysis, we envision the
stochastic discount factor as the vehicle linking a theoretical model to
observable implications.

Given a particular model for the stochastic discount factor, the
implications of (1.2) can be assessed by first taking unconditional
expectations, yielding



(1.3) Enip) = Eimp).

When m is observable (at least up to a finite-dimensional parameter vector)
by the econometrician, a test of (1.2) can be performed using a time series
of a vector of portfolio payoffs and prices by examining whether the sample
analogs of the left and right sides of (1.2) are significantly different from
each other. Examples of this type of procedure can be found in Hansen and
Singleton (1982), Brown and Gibbons (1985), MacKinlay and Richardson (1991)
and Epstein and Zin (1991).

While tests such as these can be informative, it is often difficult to
interpret the resulting statistical rejections. Further, these tests are not
directly applicable when there are market frictions such as transactions
costs or short-sale constraints. For example, when an asset cannot be sold
short, (1.2) is replaced with the pricing inequality:

(1.4) 7t(p) 2: E{mp\9) .

Finally, these tests can not be used when the candidate discount factor
depends on variables unavailable to the econometrician.

As an alternative to testing directly pricing errors using (1.3), we
consider a different set of tests and diagnostics using the
specification-error bounds of Hansen and Jagannathan (1993), and the
volatility bounds of Hansen and Jagannathan (1991). We also consider
extensions of these tests and diagnostics, developed by He and Modest (1992)
and Luttmer (1993), that handle transactions costs, short-sale restrictions
and other market frictions. We develop an econometric methodology to provide



consistent estimators of the specification-error and volatility bounds.
Further, we develop asymptotic distribution theory that is easy to implement
and that can be used to make statistical inferences about asset pricing
models and asset market data using the bounds. The specification-error and
volatility bounds, along with the econometric methodology that we develop,
can be applied to address several related issues.

The specification-error bounds of Hansen and Jagannathan (1993) can be
used to examine a discount factor proxy that does not necessarily correctly
price the assets under consideration (see also Bansal, Hsieh and Viswanathan
1992 for an application). This is important since formal statistical tests
of many particular models of asset pricing imply that the hypothesis that
their pricing errors are zero is a very low probability event. Since these
models are typically very simple, it is perhaps not surprising that they do
not completely capture the complexity of pricing in financial markets. The
specification-error bounds give measures of the maximum pricing error made by
the discount factor proxy. This provides a way to assess the usefulness of a
model even when it is technically misspecif led. Further, this tool can
easily accommodate market frictions such as transactions costs and short-sale
constraints.

Given a vector of asset payoffs and prices, (1.3) typically does not
uniquely determine m. Instead there is a whole family of m' s that will
work. Any parametric model for m imposes additional restrictions on that
family, often sufficient to identify a unique stochastic discount factor.
Rather than imposing these extra restrictions, Hansen and Jagannathan (1991)
showed how asset market data on payoffs and prices can be used to construct
feasible sets for means and standard deviations of stochastic discount
factors. The boundary points of these regions provide lower bounds on the



volatility (standard deviation) indexed by the mean. He and Modest (1992)
and Luttmer (IS 3) showed how to extend this analysis to the case where some
of the assets are subject to transactions costs or short-sales constraints.

These fea- Ue sets of means and standard deviations of the stochastic
discount factor can be used to isolate those aspects of the asset market data
that are most informative about the stochastic discount factor. One way to
do this is to ask whether the volatility bound becomes significantly sharper
as more asset market data is added to the analysis. This would help one
assess the incremental importance of additional security market data in an
econometric analysis without having to limit a priori the family of
stochastic discount factors. More generally, it is valuable to have a
characterization of the sense in which an asset market data set is puzzling
without having to take a precise stand on the underlying valuation model.

When testing a particular model of asset pricing in which the candidate
m is specified, it is often useful to examine whether the candidate is in the
feasible region. Moreover, when diagnosing the failures of a specific model,
it is valuable to determine whether the candidate discount factor is not
sufficiently volatile or wh \er it is other aspects of the Joint
distribution of asset payoffs and the candidate discount factor that are
problematic.

As we remarked previously, sometimes it is not possible to construct
direct observations of m, making pricing-error tests infeasible. However,
it may still be possible to calculate the moments of a stochastic discount
factor implied by a model which can then be compared to the volatility
bounds. For example in Heaton (1993) a consumption-based CAPM model is
examined in which the cous-umption lata is time averaged and preferences are
such that a simple linearization of the utility function can not be done to



consistency result for estimators of the arbitrage bounds. Those
uninterested in the consistency results, but who are interested in the
calculations necessary for conducting statistical inference, need only read
Sections III. A., III.C and III.D before moving to Section IV.

In Section IV we present several applications and extensions of our
results each of which can be read independently after reading Sections II,
III. A, III.C and III.D. In Section IV. A we discuss the sense in which the
entire feasible set of means and standard deviations for the stochastic
discount factor can be estimated. Section IV. B provides a discussion of
tests of whether the volatility bound becomes sharper with additional asset
market data. Section IV. C shows how to use the volatility bounds to test
models of the discount factor. Finally in Section IV. D we extend the
specification-error bound to a case where there are parameters of the
discount factor proxy that are unknown and must be estimated. Section V
contains some concluding remarks.

II. General Model and Bounds

Our starting point is a model in which asset prices are represented by
a stochastic discount factor or pricing kernel. To accommodate security
market pricing subject to transactions costs, we permit there to be
short-sale constraints for a subset of the securities. Although a
short-sale constraint is an extreme version of a transactions cost, other
proportional transactions costs such as bid-ask spreads can also be handled
with this formalism. This is done as in Foley (1970), Jouni and Kallal
(1992) and Luttmer (1993) by constructing two payoffs according to whether a
security is purchased or sold. A short-sale constraint is imposed on both



artificial securities to enforce the distinction between a buy and a sell,
and a bid-ask spread is modeled by making the purchase price higher than the
sale price.

Suppose the vector of security market payoffs used in an econometric
analysis is denoted x with a corresponding price vector q. The vector of
X is used to generate a collection of payoffs formed using portfolio weights
in a closed convex cone C of IR :

(2.1) P = {p : p = a' X for some a e C>.

The cone C is constructed to incorporate all of the short-sale constraints
imposed in the econometric investigation. If there are no price distortions
induced by market frictions, then C is r". More generally, partition x into
two components: x' = [x"',x°'] where x" contains the k components whose
prices are not distorted by market frictions and x contains the I components
subject to short-sale constraints. Then the cone C is formed by taking the
Cartesian product of R and the nonnegative orthant of R .

Let q denote the random vector of prices corresponding to the vector x
of securities payoffs. These prices are observed by investors at the time
assets are traded and are permitted to be random because the prices may
reflect conditioning information available to the investors. Since it is
difficult to model empirically this conditioning information, we instead work
with the average or expected price vector Eq.

While information may be lost in our failure to model explicitly the
conditioning information of investors, some conditioning information can be
incorporated in the following familiar ad hoc manner. Suppose some of the
security payoffs used in an econometric analysis are one-period stock or



bond-market returns with prices equal to one by construction. Additional
synthet ic payoffs can be formed by an econometrician by taking one of the
original returns, say x , and multiplying it by a random variable, say z, in
the conditioning information set of economic agents. The corresponding
constructed payoff is then x z with a price of z. Hence the price of the
synthetic payoff is random even though the price of original security is
constant. If x is subject to a short-sale constraint, then z should be
nonnegative.

The vehicle linking payoffs to average prices is a stochastic discount
factor. To represent formally this link and provide a characterization of a
stochastic discount factor, we introduce the dual of C, which we denote C .

This dual consists of all vectors in IR whose dot product with every element

n •
of C is nonnegative. For instance, when C is all of R , C consists only of

the zero vector. More generally, if x can be partitioned in the manner

described previously, the elements of C are of the form {0,^')' where g is

nonnegative.

A stochastic discount factor m is a random variable that satisfies the

pricing relation:



(2.2) Eq - Emx e C .

To interpret this relation, first consider the case in which C is r". Then
there are no market frictions and we have linear pricing. In this case
relation (2.2) is the familiar pricing equality because C has only one
element, namely the zero vector. Consider next the case in which x can be
partitioned into the two components described previously. Partition q
comparably, and relation (2.2) becomes:



(2.3) £q" - Eflix" =
Eq^ - Emx^ a 0.

The inequality restriction emerges because pricing the vector of payoffs x^
subject to short-sale constraints must allow for the possibility that these
constraints bind and hence contribute positively to the market price vector.

I I. A: Maintained Assumptions

There are three restrictions on the vector of payoffs and prices that
are central to our analysis. The first is a moment restriction, the second
is equivalent to the absence of arbitrage on the space of portfolio payoffs,
and the third eliminates redundancy in the securities.

For pricing relation (2.2) to have content, we maintain:



2
Assumption 2.1: E\x\ < a, E\q\ < oo.



2

Assumption 2.2: There exists an m > satisfying (2.2) such that Em < oo.



The positivity component of Assumption 2.2 can often be derived from
the Principle of No-Arbitrage (e.g., see Kreps 1981, Prisman 1986, Jouni and
Kallal 1992 and Luttmer 1993). The Principle of No-Arbitrage specifies that
the smallest cost associated with any payoff that is nonnegative and not
identically equal to zero must be strictly positive. Notice that from
(2.2), a stochastic discount factor m satisfies:

(2.4) a' Eimx) s a' Eq for any a e C,



which shows that Assumption 2.2 implies the Principle of No-Arbitrage
(applied to expected prices).

Next we limit the construction of x by ruling out redundancies in the
securities:

• • •

Assumption 2.3: If a'x = a 'x and oc' Eq = a ' Eq for some a and a in C, then

a = a .

In the absence of transaction costs, Assumption 2.3 precludes the
possibility that the second moment matrix of x is singular. Otherwise,
there would exist a nontrivial linear combination of the payoff vector x
that is zero with probability one. In light of (2.2), the (expected) price
of this nontrivial linear combination would have to be zero, violating
Assumption 2.3. To accommodate securities whose purchase price differs from
the sale price, we permit the second moment matrix of the composite vector x
to be singular. Assumption 2.3 then requires that distinct portfolio

weights used to construct the same payoff must have distinct expected

1
prices.

II. B: Minimum Distance Problems

There are two problems that underlie most of our analysis. Let M
denote the set of all random variables with finite second moments that
satisfy (2.2), and let M* be the set of all nonnegative random variables in
M. In light of Assumption 2.2, both sets are nonempty. Let y denote some
"proxy" variable for a stochastic discount factor that, strictly speaking,
does not satisfy relations (2.2). Following Hansen and Jagannathan (1993),



10



we consider the following two ad hoc least squares measures of
misspecif ication:



(2.5) 8^ = min £[(y - m)^] ,
meM



and



(2.6) 5^ = minjliy - m)^]
meM



When the proxy y is set to zero, the minimization problems collapse to
finding bounds on the second moment of stochastic discount factors as
constructed by Hansen and Jagannathan (1991), He and Modest (1992) and
Luttmer (1993). In particular, the bounds derived in Hansen and Jagannathan
(1991) are obtained by setting y to zero and solving (2.5) and (2.6) when
there are no short-sale constraints imposed (when C is set to IR ) ; the bound
derived in He and Modest is obtained by solving (2.5) for y set to zero; and
the bound derived by Luttmer (1993) is obtained by solving (2.6) for y set to
zero. These second moment bounds will subsequently be used in deriving
feasible regions for means and standard deviations. Clearly, the second
moment bound implied by (2.6) is no smaller than that implied by (2.5) since
it is obtained using a smaller constraint set.

Next consider the case in which the proxy y is not degenerate. Hansen
and Jagannathan (1993) showed that the least squares distance between a
proxy and the set M of (possibly negative) stochastic discount factors has
an alternative interpretation of being the maximum pricing error per unit
norm of payoffs '" P, where the norm of a payoff is the square root of its



11



second moment. When the constraint set is shrunk to M as in problem (2.6),
the dual interpretation takes account of potential pricing errors for
hypothetical derivative claims. While Hansen and Jagannathan (1993)

abstract from short-sale constraints in their analysis, pricing-error

2

interpretations are applicable more generally.

II. C: Conjugate Maximization Problems

In solving the least squares problems (2.5) and (2.6) and in our
development of econometric methods associated with those problems, it is
most convenient to study the conjugate maximization problems. They are
given by



(2.7) 5^ = maxiEy^ - Eliy - x'a.)^] - Zol' Eq) ,
aeC



and



(2.8) S^ = max {Ey^ - £[(y - x'a)*^] - 2a' Eg}
aeC



where the notation h denotes max{h,0}. The conjugate problems are obtained
by introducing Lagrange multipliers on the pricing constraints (2.2) and
exploiting the familiar saddle point property of the Lagrangian. The a' s
then have interpretations as the multipliers on the pricing constraints.

The conjugate problems in (2.7) and (2.8) are convenient because the
choice variables are finite-dimensional vectors whereas the choice variables
in the original least squares problems are random variables that reside in
possibly infinite-dimensional constraint sets. The specifications of the



12



conjugate problems are justified formally in Hansen and Jagannathan (1993)
and Luttmer (1993). Of particular interest to us is that the criteria for
the maximization problems are concave in a and that the first-order
conditions for the solutions are given by:

(2.9) Eq - £[(y - x'a)x] e C*
in the case of problem (2.7) and

(2.10) Eq - £[(y - x'a)*x] e C*

along with the respective complementary slackness conditions



(2.11) a'Eq - a'£[(y - x'a) x] = 0,



and



(2.12) a'Eq - a'£[(y - x'a)*xl = .

In fact, optimization problem (2.7) is a standard quadratic programming
problem. Interpreting the first-order conditions for these problems, observe
that associated with a solution to problem (2.7) is a random variable m = {y
- x'a) in M and associated with a solution to problem (2.8) is a nonnegative
random variable m = (y - x'a) in M . These random variables are the unique
(up to the usual equivalence class of random variables that are equal with
probability one) solutions to the original least squares problems.

Since Assumption 2.3 eliminates redundant securities and the random



13



variable (y - x'a) is uniquely determined, the solution a to conjugate
problem (2.7) is also unique. This follows because the value of the
criterion must be the same for all solutions, implying that they all must
have the same expected price a'Eq. The solution to conjugate problem (2.8)
may not be unique, however. In this case the truncated random variable (y -
x'a) is uniquely determined, as is the expected price a'Eq. On the other
hand, the random variable (y - x'a) is not necessarily unique, so we can not
exploit Assumption 2.3 to verify that the solution a is unique. As we will
now demonstrate, the set of solutions is convex and compact.

The convexity follows immediately from the concavity of the criterion
function and the convexity of the constraint set. Similarly, the set of
solutions must be closed because the constraint set is closed and the
criterion function is continuous.

Boundedness of the set of solutions can be demonstrated by investigating
the tail properties of the criterion functions. We consider two cases:
directions 9 for which 9'x is negative with positive probability and

directions 9 for which 9'x is nonnegative. To study the former case we take

2

the criterion in (2.8) and divide it by 1 + |a| . For large values of |a|

the scaled criterion is approximately:

(2.13) - £[(-x'9)*^] where 9 = a/[l + |a|^]^'^^ .

Hence |9| is approximately one for large values of Ia|. Moreover, 9'x is a
payoff in P. Consequently, the unsealed criterion will decrease (to -co)
quadratically for large values |a|.

Consider next directions 9 for which 9'x is nonnegative. From
Assumption 2.2 and relation (2.4) we have that



14



(2.14) £fli(G'x) s e'Eq

for some m that is strictly positive with probability one. Hence e'Eq must
be strictly positive unless G'x is identically zero. However, when G'x is
identically zero, it follows from Assumption 2.3 and inequality (2.14) that
e'Eq is still strictly positive.

For directions 9 for which the payoff e'x is nonnegative, we study the

2 1/2
tail behavior of the criterion after dividing by (1 + |al ) , which yields

approximately - e'Eq for large values of |a|. Hence in these directions the

the unsealed criterion must diminish (to -m) at least linearly in |a|. Thus

in either case, we find that the set of solutions to conjugate problem (2.8)

is bounded.

For some but not all of the results in the subsequent sections, we will

need for there to exist a unique solution to conjugate problem (2.8). Since

the set of solutions is convex, local uniqueness implies global uniqueness.

To display a sufficient condition for local uniqueness, let x denote the

component of the composite payoff vector x for which the pricing relation is

satisfied with equality:



(2.15) £mx* = Eq'



where q is the corresponding price vector. Also, let lf~>Q» be the the
indicator function for the event {fli>0>. A sufficient condition for local
uniqueness is that



Assumption 2.4: Ex x '^i~yn\ is nonsingular.



15



To see why this is a valid sufficient condition, observe that from the
complementary slackness condition (2.12), m is given by (y - x '3) for some
vector p. Consequently,



(2.16) Eq' = £yl^~^Q^ - £(xV' 1^-^^^ )3



When the matrix £(x x '1,~^„.) is nonsingular, we can solve (2.16) for S.

\in>vr

II. D: Volatility Bounds and Restrictions on Meains

The second moment bounds described in the previous subsection can be
converted into standard deviation bounds via the formulas:



(2.17) 0- = [5^ - (£m)^]^^^
? = [6^ - (£m)^]^^^



"2 ~2

where 5 and 5 are constructed by setting the proxy to zero. When P

contains a unit payoff, Em is also equal to the average price of that payoff

and hence is restricted to be between the sale and purchase prices of the

unit payoff. However, data on the price of a riskless payoff is often not

available, so that it is difficult to determine Em. In these circumstances,

bounds can be obtained for each choice of Em by adding a unit payoff to P

(augmenting x with a 1) and assigning a price of v to that payoff (augmenting

£q with v). In forming the augmented cone, there should be no short sale

constraints imposed on the additional security and hence no new price

distortions should be introduced. The price assignment v is equivalent to a

mean assignment for m. Mean-specific volatility bounds can then be obtained



16



using (2.7). (2.8) and (2.17).

The Principle of No-Arbitrage puts a limit on the admissible values of
v. V e \ ,v ] where X is the lower arbitrage bound and i; is the upper

'^

arbitrage bound. These bounds are computed using formulas familiar from


1 3 4

Online LibraryLars Peter HansenEconometric evaluation of asset pricing models → online text (page 1 of 4)