AN IMPROVED NONPARAMETRIC TEST FOR MISSPECIFICATION OF FUNCTIONAL FORM

By Ralph Bradley and Robert McClelland1
1. Introduction

In two recent papers, Bierens (1990,1982) derives a consistent conditional moment test of functional form that can

be a powerful tool to verify the specification of a nonlinear model. Unlike other conditional moment tests that test

for the nonzero product of the residual from an estimated regression and a function of the explanatory variables of

that regression, the Bierens test does not specify an alternative and encompassing form. It is also robust in that the

null is not restricted to any particular form, such as the linearity restriction of the notable tests of Ramsey (1969)

and Lee, White and Granger (1993), and it does not require a pre-specified parametric alternative that can be

consistently estimated, as is true of the test of Hausman (1978).

However, Bierens' test has several shortcomings. While the test is consistent against all deviations from the null,

i e. in the limit the test statistic over the sample size is nonzero for all alternative hypotheses, this limit is not

shown to be the largest in the class of conditional moment tests. Further, because a direct estimate of the

misspecification is not used in the test there is a suspicion that it may be smaller than necessary. Instead, the test is

a reduced form test that is based on the first moment of the residual from a regression multiplied by the anti-log of

a bounded function of the explanatory variables. This also means that there is little information about the nature of

any misspecification detected. Finally, there are important parameters that adversely affect the distribution of the

test if chosen to maximize the power of the test. Instead the power is increased through selecting several penalty

parameters and then solving a somewhat unintuitive constrained optimization problem.

In this paper, we attempt to develop a test that improves on Bierens' test while keeping its robustness. Our test is

based on the intuition that if the model is not properly specified then the vector of explanatory variables, X, can

better predict the residuals than the residual's mean. Thus, while many X-measurable functions, such as the

antilog of a bounded transform of X, can be used to detect misspecification an appealing function to use is the

conditional expectation of the residuals given X. We implement this intuition in the first stage of our test by

estimating the residuals as a function of the explanatory variables. Because we obtain our estimate with a kernel

regression, our test is nonparametric. The estimate is then a component of the second stage, which is similar to

2
the Bierens test with the misspecification estimate in the first stage replacing the antilog function.2 By directly

estimating the misspecification with a conditional expectation function, the limit of our test over the sample size is

the largest of all conditional moment tests, has fewer arbitrary parameters than the Bierens test and can help

researchers better understand the nature of any misspecification.

Another advantage of our test is in the potential for researchers to increase its finite sample power by increasing

the predictive accuracy of the explanatory variables. To do this, we cross validate the bandwidth of a kernel

regression, e.g., minimizing a quadratic loss function whose argument is the out-of-sample error of the kernel

regression. The quadratic loss function is particularly appealing in our case because the expected value is

minimized by predicting the residuals with their conditional expectation. This connects with our intuition that the

regressors can better predict the residuals than the residual's unconditional mean in a misspecified model, so that

procedures that increase this predictability increase the power of our test.

The minimization procedure we use does not impose as many arbitrary constraints as those imposed by the

optimization procedure in Bieren's test. Instead we use fewer a priori constraints, attempting to allow the data

itself to determine parameters such as the bandwidth. Therefore, our implementation of the test is robust to all

misspecifications and is optimized solely on its ability to use the explanatory variables to predict the residuals. Yet

under the null, the best predictor of the residuals is its unconditional mean when the loss function is quadratic,

and we show that the kernel regression converges to the sample mean of the residuals. Therefore, we can show

that our statistic has a asymptotic χ2(1) distribution under the null.

It should be noted that even if the population has independently and identically distributed errors, there is no

guarantee that the real size of our test will converge to its nominal size because of dependence inherent in the

residuals produced by a regression. To solve this, we use standard bootstrapping techniques to obtain a subsample

from the entire sample of residuals and regressors that ensures that under the null hypothesis, the real size of our

test converges to its nominal size regardless of the choice of the window width.

Finally, we compare Monte Carlo simulations of our test and those in Bierens (1990). While both tests have

approximately the correct distribution under the null hypothesis, our test rejects a greater proportion of the

3
simulations at all sample sizes when the hypothesis is false. We then choose one simulation at random under this

alternative hypothesis and show how the kernel estimation in the first stage our test can be used to describe the

nature of the misspecification.

The remainder of our paper is organized as follows. Section 2 briefly discusses Bierens test. Section 3 derives our

test. Section 4 discusses results from Monte Carlo experiments that compares our test to the Bierens test and

conclusions are in Section 5. The assumptions on the random vectors (y,x) are listed in Appendix A, and the

Proofs and Lemmas are in Appendix B.

2. The Bierens Test

We first describe the calculation of the Bierens Test, and then briefly discuss the theory behind his test. Suppose

the random vector (y,x), y ∈ R and x ∈ Rk, has a joint probability distribution F(y,x) where the following holds:

E(y|x) f(x) =

y f(x) u = + ,

where u ∈ R is the error of the model. It is clear that E(u|x) = 0 for all x and E(u)=0. In parametric regression

estimation, it is assumed that f(x) falls in a family of known real parametric functions f(x,θ) on Rk×Θ where Θ is

the parameter space and is a compact subset of Rm . Using the notation of Bierens (1990), we denote D(f) as the

set of all probability distribution functions F(y,x) on R × Rk such that P(f(x)) = E(y|x) = 1. Therefore, we wish to

test that a specific parameterization of f(x) which is denoted f(x,θ) satisfies the null hypothesis:

(1.) Ho: The distribution F(y,x) belongs to D D(f(., ) 0 =

∈

θ

θ Θ

U )

In other words, the null states that there is at least one θ in Θ such for all x values in the support of x except for

subsets with zero measure, E(y|x) = f(x,θ). The alternative hypothesis is that

(2.) H1 : The distribution belongs to D D(f) D 1 0 =

f

U \

where the f is the union over all Borel measurable real functions, f, on Rk. Given a consistent estimator for θ

which we denote as θ

ˆ, Bierens (1990) proposes to test the null hypothesis based on the following statistic from the

random sample {yi

, xi}, i = 1,...,n and t ∈ Rk:

4
(3.a) $ ( , $) [ $ ( , $ Wt n t s t θ θ = M )] / $ ( ) 2 2 ,

where

(3.b) $ ( , $) (/ ) ( ( , $ Mt n y )) exp( ' ( )) j f xj j

n

t x j θ θ = −

=

1 ∑

1

Φ .

In (3.a), sˆ2(t), is a consistent estimator of the asymptotic variance of n Mˆ (t,θ

ˆ), and in (3.b), Φ(x) is a bounded x

measurable function from Rk to Rk and there is finite vector C ∈ Rk such that supx Φ(x)<C. Bierens uses the

inverse tangent function where the ith element in the vector is tan-1(xi

).

The test in (2.a) is a conditional moment test, which means that it is a member of the set of tests {Wh}, defined by

(4.) {Wh}=

1

1

2

n

y fx hx

s

hx HX

j jj j n ( (, $)) ( )

$ () ( )

− L

N

M

M

M

M

O

Q

P

P

P

P

∈

R

S

|

|

T

|

|

U

V

|

|

W

|

|

∑ = θ

,

where H(X) is the set of X-measurable functions and X is the support of x. One can use the probability limit of

W/n, which is equal to E(u(h(x))/E(u2h(x)2), to characterize the asymptotic power of the test.3 Newey (1985)

describes this test in which h(x) is equal to a finite sum of weights, each of which imposes a restriction upon the

conditional moment. As Bierens points out, this finite number of moment conditions implies that the Newey test

cannot be consistent against all possible alternative hypotheses.

In contrast, the use of h(x) =exp(t' Φ(x)) makes the test in (2.a) consistent against all deviations from the null

hypothesis. The theoretical underpinning for this comes from Theorem 1 in Bierens (1982), where he shows that if

f(x,θ) is misspecified, then E{u*exp(t' Φ(x))}is nonzero for some t ∈Rk. Bierens further shows that E{[yf(x,θ)]exp(to' Φ(x))}is nonzero for some to in a neighborhood of zero. The expectation of the right-hand side of

(3.b), which is used as an estimate of E{[y-f(x,θ)]exp(to' Φ(x))}, will then be nonzero if the model is misspecified.

These results are used in Bierens (1990) Theorem 2 to show that Wˆ (t,θ

ˆ) has asymptotic power against the general

alternative (2.). Unlike the Ramsey RESET test (1969), a wide range of nonlinear nulls can be tested for correct

specification. Likewise, unlike the Hausman (1978) test, this test is robust against any departure from the null that

is Borel-measurable.

5
Although Bierens shows that Wˆ (t,θ

ˆ) has asymptotic power against all alternatives, he says nothing about its

asymptotic power relative to other conditional moment tests in the set {Wh}. Further, because the function

exp(t'Φ(xj

)) does not directly estimate the misspecification, it is reasonable to suspect that a more powerful test

exists. This suspicion is reinforced by the dependence of power of the test upon t, and as Bierens points out, one

cannot choose t to maximize Wˆ (t,θ

ˆ) because the asymptotic distribution of maxt Wˆ (t,θ

ˆ) converges to the supremum

of a χ2 process and not a χ2 itself. In a sense, maxt

 Wˆ (t,θ

ˆ) overfits the sample moments under the null.

In order for Wˆ (t,θ

ˆ) to achieve an asymptotic χ2(1) distribution, Bierens implements the test by using random

uniform draws of t from elements in a compact hypercube whose boundaries must be picked in an arbitrary

manner. In other words, t is a vector of random variables whose parameters cannot be chosen to optimize the

power of the test. Given these random draws, the statistic does not have a χ2 distribution under the null unless the

researcher further solves a constrained maximization problem, using constraints on Wˆ (t,θ

ˆ) formed by a priori

penalty parameters. Because of the innovative nature of the problem there is little intuition to guide researchers as

to the choice of these parameters. The dependence of the test upon the hypercube boundaries and these penalty

parameters is therefore unfortunate because Bierens claims that the finite sample power of Wˆ (t,θ

ˆ) is sensitive to

their selection.

Finally, the lack of a direct estimate of the misspecification means that there is little to direct the researcher toward

a functional form that better specifies E(y|x). In order to gain insight into the nature of the misspecification when

using (3.a), Bierens uses an additional model with the residuals as a dependent variable and exp(t'Φ(xj

)) as a righthand side variable needs to be estimated. However, if the residual is a function of the explanatory variables, why

not improve the efficiency of the test by incorporating an estimate of this function into the test itself?

3. The Kernel Test

In this section we propose a test in the set of conditional moment tests {Wh} that uses a direct estimate of the

misspecification g(x)=E(y-f(x,θ

ˆ)|x) as the function h(x). Like the Bierens test, our test is consistent against all

deviations from the null hypothesis. However, our test has the greatest asymptotic power of all tests in {Wh}in the

6
sense that the probability limit of the kernel estimate is g(x), which we show is the function that maximizes plim

(W/n), where W is defined in (4.). We also increase its finite sample size power by using a cross validated window

width. This cross validation increases the power of the test over a fixed window width even though the kernel

regression estimator converges slowly to its asymptotic limit. In addition, the test presented here does not rely on

random draws of t from a hypercube with arbitrary boundaries or the initial choice of nonintuitive penalty

parameters. Instead, we use standard kernel regression and bootstrapping techniques that have been extensively

studied and set two parameters: an upper bound on the window width and the size of the bootstrap sample.

Finally, we can use the estimate of g(x) to gather insight into the nature of any problem detected by the test.

To start, we restrict ourselves to an independently and identically distributed (IID) random sample {yi

, xi}, i=1,...,n

from a distribution F(y,x) on R×Rk for which E(y2) < ∞. As in Bierens, the extension to serially correlated series

should be possible but is outside the bounds of this paper. Suppose we use the functional form f(x,θ) to estimate

E(y|x). Under Ho, we assume that the true value of θ which we denote as θο satisfies:

(5.) θο = argminθ∈θ E([yi - f(xi

,θ )]2 ) and E([yi-f(xi,θο )]|xi) = 0.

Let θ

ˆ be a consistent estimate of θo. This study is based on assumptions, outlined in Appendix A, that ensure that

under Ho plimθ

ˆ-θo= 0 and under H1 plim θˆ exists. If Ho in (1.) is correct, then the asymptotic expectation of the

product of y-f(x,θ

ˆ) and any x-measurable function equals zero and

(6.) plim E(y-f(x,θ

ˆ)|x) = 0 for all x.

Under H1 ,there is a set X of non zero measure such that:

7.) plim E(y-f(x,θ

ˆ)|x) ≠ 0 for all x ∈ X.

since Pr[lim f(x,θ

ˆ) - E(y|x)] < 1.

Define ûi

 = yi - f(xi,θ); then the Rao-Blackwell theorem tells us that under H1 :

(8.) lim E(ûi

-E(ûi|xi))2 < lim E(ûi - E(ûi))2

which implies lim E(ûiE(ûi|xi))=lim E(E(ûi|xi)2) > 0. Under Ho, since plimθ

ˆ

=θo , the inequality in (8.) becomes an

7
equality. This implies that under Ho, lim E(ûi

E(ûi|x) = 0. Therefore, our test is based on a nonparametric

estimator for E(ûi

E(ûi|x). The features in (5.) through (8.) suggest the following four steps to test the null in (1):

(i) Estimate θ and denote this estimate as θ

ˆ.

(ii) Generate ûi

 = yi - f(xi,θ

ˆ) for i = 1,..,n.

(iii) For each observation i, estimate E(ûi

|xi) with a kernel regression by selecting a random subsample

with replacement of size n' < n from the existing sample. Specifically, for the original sample of size n,

Let N={1,...,n} index the sample. We then choose for each sample point i a random subsample with

replacement from N of size n' which we denote as Ni

'

 and calculate for each i a kernel estimator denoted

as gˆ

i

(xi, γi

) where:

(9.) $

$ ( )

( )

g (x , )

u(x )K

x x

x x i i

j

i j

j Ni'

i j

j Ni'

γ γ

γ

i

i

i

=

−

−

∈

∈

∑

∑ Κ

and K(.) is a kernel function which has the following properties:

(P.1) ∫ K(u) du = 1

(P.2) ∫ uK(u)du = 0

(P.3) K(u) > 0 for all u.

(P.4) 0 = argmax ( ) K u , K'(u) > 0 for u<0 and K'(u) < 0 for u>0.

(iv) Calculate the statistic:

(10.) W( , ) $ $ γ θ ∗ = n Tˆ 2(γ∗ ,θ

ˆ)/sˆ2(γi

),

where

(11.) $ ( *, $ T n ux g x ) (/ ) $( )$ (,) i

i

n

iii γθ γ = L

N

M O

Q

P =

1 ∑

1

,

γ* = {γ1,γ2,...,γn}

(12.) $ ( ) $( )($(,) $( ) $ ( / )( , $)) * * s

n

ux gx b A fx i ii i

i

n 2 2 12

1

1 γ γ γ ∂ ∂θ θ = − −

=

∑

(13.) $( *) ( / )( , $ b f γ ∂ ∂θ θ γ = )$(,)

=

∑ xi g xi i

i

n

1

8
(14.) $ {( / ' ) ( , $)}{( / ) ( , $ A = )}

=

∑ ∂ ∂θ θ ∂ ∂θ θ f xi f xi

i

n

1

.

Given the properties (P.1) through (P.4) our test, like Bierens', is consistent against all alternative hypotheses.

This follows from the fact that if E{û ⋅ exp(t' Φ(x))} is nonzero, then E{û⋅gˆ(x,γ)}is also nonzero (see Bierens

(1987)). But because the first stage of our test estimates E(û|x) with gˆ(x,γ), if gˆ(x,γ) is consistent then the

asymptotic power of our test should be at least as great as the Bierens test. The reason for this difference is that

E(û|x) is the function h(x) in the set of bounded functions which maximizes E{ûh(x)}. In essence, Tˆ in equations

(10.) and (11.) is an estimator of E{û E(û|x)} and therefore the numerator in equation (10.) uses an estimator that

maximizes the value of E{ûh(x)}.

To increase the predictability of gˆ(x,γ) in finite samples, we choose γi

 as the cross validated window width by

solving:

γ γ

γ

i

Bn

= − ux g

L

N

M O

Q

P < < <∞

−

∈

argmin x ∑ k kk

k Ni 0

2 ( $ ( ) $

$ ( , ))

'

where

$

$ ( ,)

$( )( )

( )

',

',

g

ux K

x x

K

x x

j

k j

j Ni j k

k j

j Ni j k

−

∈ ≠

∈ ≠

=

−

−

∑

∑

k k x γ

γ

γ

and the bound Bn to have two properties:

(P.5) Bn ≤ Bn+1

(P.6) lim sup Bn = B < ∞ .

The limit, B, on the window width γi

 is the only parameter we must choose a priori and has another important

feature under Ho.

4 The bound B along with the property (P.4) for the kernel K(⋅) guarantees that sˆ(γ*)2 is always

positive as long as the variance of x is greater than zero. In addition, searching for the optimal γ over a bounded

set guarantees a solution in a finite time.

9
Given that this test uses standard a standard kernel regression to estimate the misspecification, it is perhaps

surprising that this test has not been used earlier. Two possible complications may explain this. First, gˆi

(xi,γi) is

not an IID process even though the errors in the true model are IID under Ho, because there is finite sample

dependence among observations of ûi

 arising from the first moment restriction, u x $( ) i i

n = ∑ = 0 1 when there is a

constant term. We solve this problem in step three by using a bootstrap from the original sample. This

bootstrapping allows us to use a projection theorem that shows that the moments of {ûi

gˆi(xi,γi)}, i=1,2,..n converge

in probability to the moments an IID random process. The only parameter we must choose is the size of the

bootstrap, which can be chosen according to standard bootstrap criteria, as long as the bootstrapped sample is of

size O(n).

A second problem may be the bias inherent in a kernel estimate. However, the power of our test does not depend

on the lack of bias in gˆi

(xi,γi). Rather it depends upon the ability of gˆi(xi,γi) to predict ûi. If any X measurable

nonconstant function can better predict the residuals using a quadratic loss function, then by the Rao-Blackwell

theorem E(ûi

,|xi) is non-constant and therefore we can detect the misspecification.

We now state the basic results of this paper:

THEOREM 1: Let the assumptions in Appendix A hold. The statistic Wˆ (γ∗ ,θ

ˆ) generated in equation (10.) is

asymptotically distributed as χ2(1) under Ho.

PROOF: See Appendix B.

Wˆ (γ∗ ,θ

ˆ) is essentially a Wald Test for the restriction that E(û[Eû|x]) = 0. Because we are imposing one restriction

the limiting distribution must have one degree of freedom.

While both Wˆ(t) and W( , ) $ $ γ θ ∗ are conditional moment tests consistent against all alternatives we now show that

W( , ) $ $ γ θ ∗ is most asymptotically powerful in the set of statistics W. We do this by first showing that a function

h(x) that is proportional to g(x) is the solution to the problem:

10
(15.)

max

( )

lim ( ( ))

h HX ( ()) p

W

n

E uh x

∈ Euhx = L

N

M O

Q 2 2 P

2

and then observing that plim gˆi(xi,γi) = g(x) implies that plimWˆ (γ∗ ,θ

ˆ)/n is the largest among from the set W.

PROPOSITION 1: Let {x,u} be a random vectors for some probability space where E(u)=0, and E(u2) = σ2 < ∞.

Denote g(x) = E(u|x) and g(x) is an X measurable continuous function. Then the solution to (1.) is proportional to

g(x).

PROOF: See Appendix B.

COROLLARY 1: Given the assumptions Appendix A, plim Wˆ (γ∗ ,θ

ˆ)/n equals the maximand corresponding to the

solution of (15.).

4. Monte Carlo Experiments

In this section we briefly explore the finite sample size and convergence properties of our test by conducting an

experiment identical to that conducted in Bierens (1990). We then give an example of how the information from

g

ˆ

i

(xi,γi) can be used to understand the nature of the misspecification.

In Bierens (1990) the following data generating mechanism is examined:

x1i = zi

 + v1i , v1i ~ N(0,1)

x2i = zi

 + v2i , v2i ~ N(0,1)

yi

 = 1 + x1i + x2i + ui .

The residual ui

 is generated by either one of two data generating processes

(DGP1) ui

 = v1i v2i + ei

or

(DGP2) ui

 = 2 ei.

where

ei

 ~ N(0,1).

Both processes are fitted with the following model:

11
(16.) yi

 = θ0 + θ1x1i + θ2x2i + ui

The null is false for DGP1 and true for DGP2.

In Table I we compare the results of our test (labeled Kernel) with Ni

=.95N and a Gaussian kernel on these to data

generating processes with the results in the rows in Table 1 of Bierens (1990) where the penalty parameters γ and ρ

yielded the most rejections of DGP1 while still maintaining reasonable size properties for DGP2. The hypercube

chosen was [1,5]×[1,5] from which [N/10] - 1 draws were taken and the weighting function Φ(x) is set to tan1(x/2).

Our test rejects DGP1 more often than the Bierens test at all data sizes while offering similar size characteristics.

In particular it does well at small sample sizes, rejecting the null hypothesis at the 5 per cent critical level in more

than two thirds of the simulations using 50 observations. With one hundred observations, it rejects more than 90

per cent of the simulations and with two hundred observations it rejects all but three of the simulations at the 5 per

cent critical value.

It should be pointed out that Bierens states that his choice of hypercube accidentally falls in one of the worst areas

in terms of power. A better choice may yield more rejections for DGP1. Unfortunately, choosing the hypercube to

maximize the statistic is not allowed. This itself points out a problem with the test: it crucially depends upon a

parameter which the researcher cannot choose to increase the power of the statistic.

As an example of how to use the information from our test to better understand the nature of the misspecification,

we more closely examine the final simulation of DGP 1 in which Wˆ (γ∗ ,θ

ˆ) is equal to 20.40. In figure one we

show the true conditional mean of the residual as a function of the random variables x1

 and x2. It is clear from the

parabolic shape that a quadratic term is missing from the original specification. In Figure 2, we plot the true

conditional mean of the residual against the realizations of x1

. The parabolic shape in figure one is clearly reflected

in this figure as well. The plot of the kernel estimate also reflects the parabolic shape, although the finite sample

bias of the kernel estimate causes an upward shift. Figure 3 plots the same two variables against the realizations of

x2

. Again the parabolic shape is evident in both the expectation of the residual and the kernel estimate.

12
While it would seem useful to plot the kernel estimate and its confidence intervals with the expected residuals in a

diagram similar to figure one, the complexity of such a figure renders it practically useless. Instead Figure 4 plots

the kernel estimate and its 95 per cent confidence intervals against the expected residual, holding x 2

 fixed at the

arbitrary level of 0.5. In this figure the kernel estimate hovers around expected residual, and the residual is within

the confidence intervals in about 95 per cent of the cases. Combining the information provided by the kernel

estimates in the figures, a researcher might suspect that a quadratic term is missing from the regression.

As a final step we run two additional regressions. In the first, we include the missing quadratic components as

explanatory variables. In the second, we include the kernel regression as an explanatory variable. Running our

test on these new regressions yields the insignificant values of 0.12 and 3.30, respectively.

5. Conclusions

The test of Bierens (1990) is an important advance in attempt to detect misspecification of the functional form

through a conditional moment test. In this paper we describe an improvement upon the Bierens test that directly

estimating E(û|x) with a standard kernel regression on a bootstrapped sample of the data. It retains all of the

robustness of Bierens' test, but does not requires the arbitrary selection of penalty parameters and boundaries and

needs fewer parameters to be determined by the researcher. Through the use of E(û|x) our test is the most powerful

in the class of conditional moment tests. Because E(û|x) is estimated, researchers obtain an estimate of any

misspecification. In addition, cross-validation of the bandwidth in the kernel regression can be used to increase the

power of the test, which we conjecture is at least as great as the Bierens test. We show the results of a Monte Carlo

simulation study that suggests that our test has good power against misspecification. Finally, we use the results

from an arbitrary simulation to demonstrate how the estimate of E(û|x) can be used to understand the nature of the

misspecification.

US Bureau of Labor Statistics

13
APPENDIX A

Assumptions for the random vector (y,x)

(A.1) {yi,xi}, i=1,...,n are a simple random sample from a continuous probability distribution on R×Rk with

E(yi

2) < ∞ .

(A.2) Under Ho, the parameter space Θ is a compact and convex subset of Rm and f(x,θ ) is a Borel measurable real

function on real k and for each k vector x is a twice continuously differentiable real function on θ. E[supθ∈Θ f(xi

,θ )2] <

∞ and for i1,i2 = 1,..,m,

E fx fx i i

sup {( / ) ( , )}{( / ) ( , )} θ

∂ ∂θ θ ∂ ∂θ θ

∈

L

N

M O

Q

P < ∞

Θ 1 2 1 1

E y fx fx fx i i

sup { ( , )} {( / ) ( , )}{( / ) ( , )} θ

θ ∂ ∂θ θ ∂ ∂θ θ

∈ − L

N

M O

Q

P < ∞

Θ 1 1

2

1 1 1 2

E y fx fx i i

sup 

θ

θ ∂ ∂θ ∂ ∂θ θ

∈ − L

N

M O

Q

P < ∞

Θ { ( , )}( / )( / ) ( , ) 11 1 1 2

(A.3) E(y1 - f(x1,θ)) takes on a unique minimum on Θ at θo. Under Ho, the parameter vector θo is an interior point of 

Θ.

(A.4) The matrix A defined in (13.) is nonsingular.

14
APPENDIX B

Proofs

In the standard literature on kernel regression estimation the optimal window with γi is O(n-1/(k+4)); however under

Ho, we show:

LEMMA 1: Let v be a random variable satisfying E|v| < ∞ and E(v) = 0. Let x be a bounded random vector in

Rk. Suppose that h-i

ˆ (x,γ ) is the kernel regression of v on x with the i-th observation excluded from the calculation

of the kernel regression:

$ ( )

)

)

h x

v

x x

x x i i

j

i j

j

j i

n

i j

j

j i

− n

=

≠

=

≠

=

−

−

∑

∑

K(

K(

γ

γ

1

1

.

Denote gn* as the γ that minimizes sample variance,

Sn(γ )= ( $ v h , )) / i(xi n i i

n − − =

∑ γ 2

1 .

If P[E(v|x)=0]=1 , then plim gn* = ∞.

PROOF:

gn

* = argminγ Sn(γ). If Pr(E(v|x) = E(v)= 0)=1, we show that plim g*

n=∞.

Our proof uses the following facts:

i) Let μ be a constant and S_ = E(v - μ )2. Then S_ is minimized when μ = 0 , and ∂S / ∂μ is continuous in

μ.

ii) $

h (x v , ) -i i γ = at γ = ∞ where v_

 is the sample average of v.

iii) Sn(γ) converges in probability uniformly in γ to a nonstochastic function, S(γ), and min

γ

S(γ)= S(∞)=S

_

.

Let ψnbe the half closed interval [n,∞ ) and let δn= min

γ ψ∉ n S(∞ ) - S(γ ). Define An to be the event

 |Sn(γ)-S(γ )|<δn/2 for all γ. Then

15
(B.1) An ⇒ S(g*

n) <δn/2+Sn(g*n)

and by definition of Sn(g*

n), we know:

(B.2) Sn(g*

n) ≤ Sn(∞ )

(B.3) An ⇒ Sn(∞) <δn/2+ S(∞).

Combining (B.1), (B.2), and (B.3) we obtain:

(B.4) An ⇒ S(g*

n ) < S(∞)+δn

Therefore An ⇒ g*

n ∈ψn which implies,

(B.5) P(An) ≤ P( g*

n ∈ψn ) .

By (iv) above P(An)→1

Then since inf ψn → ∞ , plim g*

n = ∞ .

QED

LEMMA 2

Given H0, the assumptions in Appendix A, and the properties (P.1) through (P.6),

(i) lim $ , n

E

n

ugx i ii

i

n

→∞

F

H

G I

K

J =

=

∑ 1 0

1

b g γ

where g $ , , $ xi i e j γ θ is the regression defined by (8.) based on resampling x u x i i m r , $b g and by errors validation is with

replacement, where the size of the replacement is O(n). Furthermore,

(ii) 1

n

ugx i ii $ $ , , $ ∑ e j γ θ is Op(1)

 which implies

(iii) plim 1 0

n

uΣ i i $ $ , , $ g x , i e j γ θ =

16
PROOF: For simplicity we drop the argument $

θ and write $ , , $ g xi i e j γ θ as g x $ , i i b g γ . To prove (i) first note

that

(B.6) 1

n

ugx i i ∑ $ $b g , γ = 1

n 1

u uw i j ij

i j N

n

i

$ $

= ∈

′

∑ ∑ ,

where

(B.7) w

K

x x

x x

ij

i j

i

i k

k i

n =

F −

H

G I

K

J

F −

H

G I

K

J =

′

∑

 

 

γ

γ 1

,

and K(.), Ni and n' are as defined in the text. Then to prove (i), it is sufficient to show that

(B.8) lim $ $ n

Euuw i j ij →∞= 0 ∀ i ∈ {1,...,n}, j ∈ Nj .

Based on sampling with replacement

Euuw EuEu u w i j ij i j i ij d ie j $$ $ = ($ $ ) .

Because

u$ , i

i

n

=

=

∑ 0

1

we know

- u u $ $ j i

j i

=

≠

∑ ,

and

Eu u

u n

u j i

i

i

$ | $ $ ( )

$ d i = R−− ≠ S

T

1 if j i

if j = i

 

 .

Therefore, based on resampling, the assumptions in appendix A and (P.1) to (P. 6),

(B.8) E u u w i j E u w i j ij i ii d i b gd i $ $ = = Pr $ 2 - Pr $ i jE u

n

w i ≠ ij −

F

H

G I

K

J b g

2

1

or

(B.9) E u u w i j ij d i $ $ = − 1 2

n

Eu w w i ii ij { } $ d i .

It can be shown that

lim $ nEu w w i ii ij →∞− = 2 d i 0 ,

17
and therefore

lim $ $ nEuuw i j ij →∞= 0 .

We can prove (ii) by showing that

n Var 1 1

n

ugx i ii ∑ $ $b g af , γ F

H IK = O

or equivalently that

nE

n

ugx i ii

i

n 1

1

2

$ $b g , γ

=

∑ F

H

G I

K

J = O(1).

To show this first note that because ûi

gˆ(xi,γi) is identically distributed,

nE

n

ugx i ii

i

n 1

1

2

$ $b g , γ

=

∑ F

H

G I

K

J = − −

−

F

H

G I

K

J

L

N

M

M

O

Q

P

P Eug x

n n

n

E u

n

i i gx gx iii jj $ $ , $ $ , $ , 2 2

2 1

1 d i b g γ γγ a f b gd i .

Since

u O $ i p

2 = a f1 ,

it suffices to show that under Ho,

gx O $ , ii p b g af γ

2 = 1

and

gx gx O $ , $ , ii jj p b gd i af γ γ = 1 .

Note that

(B.10) gx uw $ , $ i i j ij

j

b g γ

2

2

1

=

F

H

G I

K

J ∈

∑

Ν

 = uu w w $ $j k ij ik

j k ∈ i ∈ i

∑ ∑

Ν Ν

.

Given the properties (P.1) through (P.6), it follows:

 (B.11) wij

j∈ i

∑

F

H

G I

K

J = Ν

2

w ww ij

j

ik ij

i k j

k j i

2 1 ∈

∑ ∑ + =

≠

∈ Ν , Ν

.

 From (B.11) we can show that wijwik is O(n'-2) for all but finitely many i, j and k. With the additional condition

that x is continuously distributed if there is at least one wijwik is greater than O(n'-2), then there exists some

neighborhood around xj

 with an infinite number of elements such that wijwik is greater than O(n'-2), contradicting

the limit of (B.11) as n goes to infinity. This implies that (B.10) is

18
nO n O p p

2 2 1 − d i = ( ).

To show that gx gx $ , $ , ii jj b gd i γ γ is no greater than Op(1), we rewrite

gx gx u w uw $ , $ , $ $ i i j j k ik j jh

k h i j

b gd i γ γ = ∈ ∈

∑ ∑

Ν Ν

= uuw w $ $ k

h

h ik jh

k∈ i ∈ j

∑ ∑

Ν Ν

.

If k≠ h then

Eu uw w k h ik jh d i $ $ = −

F

H

G I

K E u J u

n 1 k w w k

ik jh $ $ .

Because wikwjh is O(n-2), we know that ûk

ûhwikwjh is O(n-3) for all but finitely k and h. If k=h then

Eu uw w uw w k h ik jh j ik jh d i $$ $ = 2

is O(n-2) for all but finitely many k. Therefore, defining nj as card (Ni ∩ Nj)

E uuw w k h

h

ik jh

k i j

$ $

∈ ∈

∑ ∑ F

H

G I

K

J

Ν Ν

 = En u w w j

k ik jh $ 2 d i - ′ − −

F

H

G I

K n n J u

n

w w j k

ik jk e j a f2

1

$

Since nj

 is O(n') the order of

E uuw w k h

h

ik jh

k i j

$ $

∈ ∈

∑ ∑ F

H

G I

K

J

Ν Ν

is of order

On On On On O ′ + ≤ − − a f d i d i d i af 2 23 1

(i) and (ii) imply (iii) by Chebychev's inequality. This completes the proof.

LEMMA 3: (Projection Theorem for Sampling with Replacement).

Let zi

 be a random sequence of IID variables. And define

(B.11) T

n

E pz z z i j

j N

i

i

n

i

= L

N

M

M

O

Q

P = ∈ P ∑ ∑ 1

1

d i ,

where Ni is a sample with replacement from the set of integers {1, ... , n} when each point has a probability of

(1/n), n' = card (Ni

) < 0(n) and p(.,.) is a k-dimensional symmetric kernel. Define

(B.12) $

T , npz z i j

i j

n

i

= = ∈

∑ ∑ 1

1

d i

Ν

 .

If

19
E pz z on i j d i di , 2 L 1

N

M O

Q

P = −

and

Epz z pz z On ij kh d i , , b g = d i −2 for i,j ≠ k,h

 then

nT T op $ − = a f 1 .

PROOF:

Define

qz z pz z

n

E pz z z ij ij ij

j N

i

i

, , ' d id i d i = − , L

N

M

M

O

Q

P ∈ P ∑ 1

so that

$

T T , nqz z i j

i j

n

i

− = = ∈

∑ ∑ 1

1

d i

Ν

,

Eqz z i j e j d i , = 0

and

ET T

n

Eq z z q z z i j

k j h

n

i

n

k h

i k

$ − = , , F

H IK = = ∈ ∈

e j d ib g ∑ ∑ ∑ ∑

2 2

1 1

1

Ν Ν

= F

H IK = ∈

∑ ∑ 1 2 2

n 1

E qz z i j

i j

n

i

d i , Ν

+F

H IK

F

H

G I

K

J = = + ∈ ∈

∑ ∑ ∑ ∑ 1 2

2

n 1 1

E qz z qz z ij kh

k i j h

n

i

n

i k

 ,, d ib g

Ν Ν

The expectations are such that

Oqz z O pz z ij ij di di , , 2 2 F

H IK = FH IK

and

Oqz z qz z Opz z pz z ij kh ij ih e j d ib g d ib g , , , ,. = e j

Since

Eqz z qz z ij kh e j d i , , b g = 0

when i, j, k and h are not equal to each other, and based on sampling with replacement card (Ni ∩ Nj

) is O(n'),

implying that

 Ij h On

j h i k

= = ′ ∈ ∈

∑ ∑ b g af

Ν Ν

.

20
Then nE(Tˆ - T)2 is of order

nO n o n − − 2 1 d id i + ′ − − nO n O n O n 2 2 d iafd i= o(1),

which implies that n|Tˆ - T| is op

(1).

QED

THEOREM 1: Let the assumptions in Appendix A hold. The statistic Wˆ (γ∗ ,θ

ˆ) generated in equation (9.) is

asymptotically distributed as χ2

(1) under Ho.

PROOF:

Under H0:

y = f(x,θ) + u

is the true data generating process. Define the function Tˆ(γ∗ ,θ

ˆ) as in the text. Note that Tˆ(γ∗ ,θˆ) uses the

estimated function $ , , $ g xi i e j γ θ and residuals û. Now define a function that uses g x $ , , i i b g γ θ , the true residuals u,

b(θ,γ), A and f(x,θ):

(B.14) nT g

n

u gx b A f x

i ii

i

n

γ θ γ θ θγ ∂ θ

∂θ

* * , , $ $ ,, , , d i b gd i a f = − R

S

T

U

V

W −

=

∑ 1 1

1

where

b E f x θ γ g x

∂ θ

∂θ , γ θ , b g $ , , a f = b g L

N

M O

Q

P ,

$ , , $ g xi i e j γ θ is defined in (8.) with û , g x $ , , i i b g γ θ is analogously define with u (the true residuals) and

A E fx fx

d = ′

F

H

G I

K

J ∂ θ

∂θ

∂ θ

θ

a fa f , , .

We show

(i) nT T g Op $ , $ , , $ * * e j d i af γθ γθ − = 0 1

(ii) nT g T g op γθ γθ * * , , $ , , 0 0 d i d i af − = 1

where

g Egx x = ii i c h $b g , , γ θ .

21
Note that T(γ*, θ0

, g) is the sum of IID random variables with finite variance. This, the Levy Lindeberg Central

Limit Theorem and the fact that γ* converges to B implies that

nT SB ,g) d ( *, , γ θ0

2  → Νd i 0 a f

where

S B E u gx bBA

f x

i i

22 1 i af b g af b g = − R

S

|

T

|

U

V

|

W

|

L

N

M

M

O

Q

P

P − , , , γ θ

∂ θ

∂θ .

To show (i), we use the Mean Value Theorem

(B.16) n nT g T γθ γθ * * , $ ej d i − , , 0 = − n

T

 ∂

γ θ

∂θ

θ θ

* ,% $ e j e j 0

where

% : $ $ θ θθ θ θ θ θ θ θ θ ∈ − ′ − ≤− ′ − R

S

T

U

V

W 00 00 b gb g e je j .

Let $ , * bd i θ γ be defined as in the text. Then

$ %, $ , , % %

% , , % * *

*

b

n

g x

f x

u

g x

i

i

i

i

i

n

θγ γ θ

∂ θ

∂θ

∂ γθ

∂θ ej e j ej e j

=− +

R

S

|

T

|

U

V

|

W = | ∑1

1

and

 % ,% u y fx ii i == θ e j.

By the triangle inequality

(B.17) n b E gx f x

→∞ − −R

S

|

T

|

U

V

|

W

| plim

<B γ θγ γ θ

∂ θ

∂θ

sup $ %, $ , , , * * ej d i b g

0

0

≤ →∞

F

H

G

G

I

K

J

J

+

F

H

G I

K

J

R

S

|

T

| =

 ∑ plim

<B

sup n

n

g x

f x

E gx

f x

i

n

i

i

i γ γ θ

∂ θ

∂θ γ θ

∂ θ

∂θ

1

1

0

0 $ , , % , %

$ , , , * * e j e j d i b g

+

U

V

|

W

| ∑1

n

u

g x

i

i %

$ , , * % ∂ γθ

∂θ

e j = 0

By the uniform law of large numbers and assumption (A.1) and Lemma 2 the first term goes to zero, and the last

term is bounded by:

n

p

B n

u

g x

u

g x

n

u

g x

i

i

i

i

i

n

i

n

i

i

i

n

→∞ < − +

R

S

|

T

|

U

V

|

W = = = | ∑ ∑ ∑ lim %

$ , , %

$ $ , , $$ $ , , * * * $ sup

γ

∂ γθ

∂θ

∂ γθ

∂θ

∂ γθ

∂θ

11 1

1 1 1

ej ej ej

Ν

The first term goes to zero from assumption (A.1) and

22
 plim $ $ θθ θθ − ′ e je j 0 0 − = 0 ,

and the second goes to zero by Lemma 2. Since

b E gx

f x

j

o θγ γθ

∂ θ

∂θ 0 0 , $ , , , * bg d i b g = F

H

G I

K

J

it follows from B.16 and B.17 that

(B.18) n

p

B →∞ nT nT g < − lim sup , $ , , * * γ ej d i γθ γθ0 + −= b n θγ θθ 0 0 , 0 $ b ge j .

From standard nonlinear least squares theory, it follows that

(B.19) n

p A

n

u

f x

i

i

i

n

→∞ − − −

=

∑ lim $ , n = 0 θ θ

∂ θ

∂θ 0

1 0

1

1 e j b g

where

u y fx ii i = − ,θ0 b g.

Substituting the second argument on the right-hand side of (B.19) into (θ

ˆ - θ) in (B.18), we get

n

p

B →∞ n T nT g < − = lim

*

sup *, $ * , , γ ej e j γθ γθ0 0

or

nT T g Op $ , $ , , $ * * e j d i af γθ γθ − = 0 1 .

This completes part (i) of the proof. To prove part (ii), we apply the results of Lemma 3. Using the definitions of

Lemma 4, we set

zi

 = {ui, xi}

and

(B.20) p z z u u w i j i j ij d i , = .

It follows from the definitions above that Lemma 3 can be applied to ( T(γ*, θo, gˆ) - T(γ*, θo, g)). Therefore, to

prove (ii) it suffices to show that

(a) E u u w o n i j ij d i2 1 = − ( )

(b) E u u w u u w o n i j ij k h kh ( )( ) ( ) = −2

To show (a) note that

Eu u w w i j ij ii

22 2 4 2 d i < σ .

It is therefore sufficient to note that E(wii

2) can be shown to be O(n-2) by the method in Lemma 2.

23
To show (b) note that there are two cases. Either the four indexes i, j k, h have two sets of equalities, e.g. i=k and

j=h, or they do not, e.g. i ≠ j ≠ k ≠ h. In the first case, we know from (a) that E(ui

ujwij)(ukuhwkh) is at most O(n-2).

In the second case

Euuw u u w i j ij k h kh ( )( ) = 0 .

Together these imply that (b) holds. This completes part (ii).

Since nT g ( *, , ) γ θ is the sum of an IID. sequence of random variables, we get our final results by applying the

Lindberg Levy Central Limit Theorem to nT g ( *, , ) γ θ .

PROPOSITION 1: Let {x,u} be a random vectors for some probability space where E(u)=0, and E(u2) = σ2 < ∞.

Denote g(x) = E(u|x) and g(x) is an X measurable continuous function. Then the solution to (1.) is proportional to

g(x)

Proof: We first show that g(x) is the solution to :

(17.) max

( )

( ( ))

( ()) h HX

E uh x

Euhx ∈ 2 2

By the conditions of the Proposition, u can be decomposed as

(18.) u=g(x)+ε

Since E(g(x)ε)=0,

E(u2)= E(g(x)2) + Eε2

 = σux2 + σεε 2

Substituting (3.) into the denominator of (2.) and taking expectations over ε , we rewrite (2.) as:

 (19.) max

( )

( ( ))

(() () ()) h HX

E uh x

Egx hx hx ∈ + 2 22 2 σεε

without loss of generality, we normalize (4.) by letting

D2 = Ehx

Egx

(())

(())

2

2

and define

h

~

(x) = h(x)/D .

Since Eh~

(x)2) = E(g(x)2)=σux, the problem in (1.) is equivalent to the following:

24
(20.) 

max ( % ( ))

(() % () ) % { ( ): (% h HX Eh x Eg x ( ) ( ( ))} ux

E uh x

∈ = Egx hx + 2 2 2 2 22 σ σεε

The fraction in (5.) is bounded above by 1, and with the constant σux2σεε 2 in the denominator, the solution to (5.)

is the same solution to:

(21.) 

max

( %( )) % { ( ): ( % h HX Eh x Eg x ( ) ( ( ))}

E uh x ∈ = 2 2

We can now show that (6.) is maximized by h~(x)=g(x). By the Rao Blackwell Theorem,

E(u-h~(x))2 ≥ E(u-g(x))2

which implies

-E(2uh~(x)) + E(h~(x)2) ≥ -E(g(x)2) since E(g(x)2) = E(ug(x)).

Since E(h~(x)2 ) = E(g(x)2) .

E(ug(x)) ≥ E(uh~(x).

Therefore, g(x) is the solution to (2.). To be the solution to (1.), we need to show that there exists no function h(x)

∈ H(X) such that

E uh x

Euhx

E ug x

Eugx

( ( ))

( ())

( ( ))

( ()) 22 22 >

and

E uh x ( ( )) < 0.

To show this note that if E(uh(x)) is negative, then

E(u(-h(x)))= -E(uh(x)) >0

Eu hx

Euhx

Eu hx

Euhx

E ug x

Eugx

( ( ( )))

( ())

( ( ( )))

( ())

( ( ))

( ())

− = − > 22 22 2 2 ,

contradicting the fact that g(x) maximizes (2.).

QED

25
REFERENCES

Bierens, H.J.:, "Consistent Model Specification Tests," Journal of Econometrics, 20 (1982), 105-134.

-----: "Kernel Estimators of Regression Functions," in Advances in Econometrics, Fifth World Congress, vol. 1, ed. T. R.

Bewley, 1987.

-----.: "A Consistent Conditional Moment Test of Functional Form," Econometrica, 58 (1990), 1443-1458.

Lee, T.H., H. White, and C.W.J. Granger: "Testing for Neglected Nonlinearity in Time Series Models", Journal of

Econometrics, 56 (1993), 269-290.

Manski, C: "Nonparametric Estimation of Expectations in the Analysis of Discrete Choice Under Uncertainty" in

Nonparametric and Semiparametric Methods in Econometrics and Statistics. Cambridge, Mass.: Cambridge

University Press, 1991.

Ramsey, J.B.: "Tests for Specification Errors in Classical Least Squares Regression Analysis," Journal of the Royal

Statistical Society Series B, 31 (1969), 350-371.

26
TABLE I

Simulations of (DGP1) and (DGP2)

Percent of 1,000 Trials Rejected at the Asymptotic 10 per cent and 5 per cent Significance Level

Sample Size

Significance Level

50

10% 5%

100

10% 5%

200

10% 5%

DGP1

 Bierens 24.4 12.8 35.2 25.2 57.6 46.2

 Kernel 74.1 66.8 92.4 90.1 100.0 99.7

DGP2

 Bierens 12.4 4.0 10.0 6.0 12.0 5.6

 Kernel 11.3 6.7 10.8 5.3 11.1 6.0

ENDNOTES

1The authors would like to thank Tim Erikson and participants of seminars at the American Statistical

Association, the Bureau of Labor Statistics and George Mason University. The authors are especially grateful to

Anna Sanders for her extraordinarily patient editing. The views expressed in this paper are solely those of the

author, and do not necessarily reflect the policy of the Bureau of Labor Statistics(BLS) or the views of other staff

members of BLS.

2For an example of a two-stage procedure for semi-parametrically estimating a binary choice model under

uncertainty using estimated conditional expectations, see Manksi (1991).

3For example, see Bierens (1990) Theorem 2.

4 Simulations suggest that our test is insensitive to the limit B.

27
AN IMPROVED NONPARAMETRIC TEST FOR MISSPECIFICATION OF FUNCTIONAL FORM

Ralph Bradley and Robert McClelland

 US Bureau of Labor Statistics

2 Massachusetts Ave NE

Room 3105

Washington, DC 20212

In this paper, we develop a two-stage test for misspecification of a conditional mean that is similar to one recently

developed by Bierens(1990). Our test uses the idea that in misspecified models the explanatory variables, X, better

predict the residuals, u^, than their mean. Thus, an appealing function that detects misspecification is E(u^|X). We

estimate this with a kernel regression which is then a component of the second stage which tests for the nonzero

correlation of u^ and the kernel estimate. By directly estimating the misspecification our approach should have

good power while making fewer a priori restrictions than Bierens, while also helping researchers better understand

the nature of any misspecification.

KEYWORDS: Consistent Tests, misspecification of functional form, conditional moment tests, kernel regression.
-2

-1.8

-1.6

-1.4

-1.2

-1

-0.8

-0.6

-0.4

-0.20

0.2

0.4

0.6

0.81

1.2

1.4

1.6

1.82

-2

-1.2

-0.4

0.4

1.2

2

-4

-3.5

-3

-2.5

-2

-1.5

-1

-0.5

0

0.5

1

FIGURE 1 - Conditional Expectation of the Residual as a Func
-4

-3-2-1

0123

-5 -4 -3 -2 -1 0 1 2 

Conditional Expectation of theResidualKernel Estimate

FIGURE 2 - A Comparison of the Conditional Expection of the Residual and the Kernel Est
-4

-3-2-1

0123

-4 -3 -2 -1 0 1 2 

Conditional Expectation of theResidualKernel Estimate

FIGURE 3 - A Comparison of the Conditional Expectation of the Residual and the Kernel E
-3

-2-1

01234

FIGURE 4 - A Comparison of the Conditional Expection of the Residual and the Kerne

Intervals, holding X2 Constant

Expected Residual

Kernel Estimate