Computation Of Asymptotic Distribution. For Semiparametric GMM Estimators. Hidehiko Ichimura. Graduate School of Public Policy

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1 Computation Of Asymptotic Distribution For Semiparametric GMM Estimators Hidehiko Ichimura Graduate School of Public Policy and Graduate School of Economics University of Tokyo A Conference in honor of Professor Dan McFadden

2 1. Motivation 2. Model 3. Overview and preliminary results 4. Main results 5. Applications 6. Conclusion

3 1 Motivation Two reasons: Many works claried the structure of the asymptotic analysis but given a problem, we still cannot carry out the asymptotic analysis in the same way we can for the standard case. Common structure needs to be claried further to obtain systematic results on smoothing parameter choice problem.

4 2 Model E fg (z; ; (; ))g = 0 if and only if = 0 and (; ) = 0 (; 0 ). (; ) is estimated by ^ (; ). Notations: z 2 R k, 2 R p, for each, is a functions into R d with norm kk. Write instead of (; ) and g i (; ) instead of g (z i ; ; (; )). Often g i (; ) = g (z i ; ; h 1 (h 2 (z i ; ) ; )) for an unknown function h 1 into R` and a known function h 2 so that g can be regarded as a function from Z R` into R m.

5 The added generality is useful to handle applications where is a conditional expectation of an unknown variable which needs to be estimated, for example. The generality is also useful in applications where individuals' decisions depend on the entire distribution of a variable, which in turn is estimated. This is the case for individual decisions in auction models, for example, or more generally any decision under explicitly stated expectation which is to be estimated. Let G n (; ) = 1 n nx i=1 g i (; ) : We study the generalized method of moment estimator which is dened as a solution to the following problem: inf 2 G n (; ^ ) T ^AG n (; ^ )

6 where ^ is an estimator of and ^A is an mm symmetric matrix which converges in probability to a symmetric positive denite matrix A.

7 3 Overview of results Our approach is a direct application of the standard analysis of the two-step GMM estimators. Like the standard analysis, the basic result appeals to the Taylor's series expansion with respect to function (; ) at 0 (; ). Let F be a mapping from B into R K and let F be dened over an open subset O of B. The Taylor series expansion of F is available when the rth Frechet derivative F (r) (x) exists for any x 2 O and is uniformly continuous: F (x + h) = F (x) + F 0 (x) (h) + 1 2! F 00 (x) (h; h) r! F (r) (x) (h; :::; h) +! (x; h) where k! (x; h)k B = o (khk r B ). If the rth derivative satises the Lipschitz condition with exponent > 0, then k! (x; h)k B = O khk r+ B :

8 Let n denote a sample size. We consider an estimator ^ 0 of an element 0 in and assume it is asymptotically linear: there exist a stochastic sequence f ni g n i=1 with ni 2 and E ( ni ) = 0 (when the function is evaluated at each point) and a deterministic sequence fb n g with b n 2 such that ^ 0 0 n 1 nx i=1 ni b n = o p n 1=2 : The norm used for needs to be stronger that the norm used to dene the Frechet dierentiability. Three dierences compared to the standard case: 1. Importance of the norm used and the specication of. 2. Presence of bias. 3. U-statistics CLT rather than the standard CLT.

9 4 Main results Let rg = E frg (z; 0 ; 0 (; 0 ))g where rg (z; ; (; )) (z; ; (; )) =@ +@g (z; ; (; )) (; ) =@; H = (rg) T A (rg), and denote the expectation conditional on z i by E fjz i g. Let n = V ar[g (z 1 ; 0 ; 0 ) +E (z 1 ; 0 ; 0 ) =@ 0 n2jz 1 o +E (z 2 ; 0 ; 0 ) =@ 0 n1jz 1 o ]

10 The main result is the following: Suppose ^ is consistent to 0. Under the conditions below n1=2 ^ 0 converges in distribution to a normal random vector with mean plim n 1=2 n!1 nx i=1 and variance-covariance i ( 0 ; 0 ) =@ 0 b n lim n!1 H 1 (rg) T A n ArGH 1 :

11 Here are the conditions: 1. fz i g n i=1 are independent and identically distributed. 2. ( 0 ; 0 (; 0 )) is an interior point of f(; (; ))g plim n!1 ^A = A where A is symmetric and positive denite. 4. g (z; ; ) is Frechet dierentiable with respect to (; ) in and the Frechet derivatives satises the Lipschitz continuity conditions: for C j (z) > 0 E n C j (z) o < 1 (j = 1; 2; 3; (z; ; ) =@ C 1 (z; ; ) =@ C 3 z; 0 ; 0 =@ 0 R p + C 2 (z) 0 R p + C 4 (z) z; 0 ; 0 =@ 0 R mp L

12 5. (z; ; (; )) =@k L + k@g (z; ; (; )) =@k R mp C 0 (z) and E fc 0 (z)g < E frg (z; 0 ; 0 (; 0 ))g rg is nite and has full rank. 7. 7! 0 (; ) as a mapping from into is continuous at sup 2N (0 ;") k^ (; ) 0 (; )k = o p n 1=4 and that ^ (; 0 ) is asymptotically linear for 0 (; 0 ) in with rate n 1=2. 9. plim n!1 n 3=2 P n i ( 0 ; 0 ) =@ 0 ni = 0.

13 10. plim n!1 n 1=2 P n i ( 0 ; 0 ) =@ 0 b n exists. Typically we will nd conditions under which the limit is 0. Under the condition i ( 0 ; 0 ) =@ 0, the term can be bounded: n 1=2 nx i=1 nx 1 n i ( 0 ; 0 ) =@ 0 b n C 0 (z i ) n 1=2 kb n k : R m Thus if n 1=2 kb n k = o (1), then the limit is ^ (; ) is continuously dierentiable and sup k@^ (; ) 0 (; = o p (1) : 2N ( 0 ;")

14 5 Applications To carry out these computations, we need to nd out the relevant Frechet derivatives and know what the asymptotic linear expressions are for the nonparametric estimators used in the estimation. For the kernel density estimators the following are the terms in the asymptotic linear expressions: ni = 1 zi z h dk h 1 E h dk 1 b n = E h dk zi zi h h z z and f (z) : For kernel regression estimators of g (x) = E (Y jx = x) ;

15 denoting " = Y g (X), the asymptotic linear approximation of (^g g) I ^f > b takes the following form: ni = " ih d K ((x i x) =h) I (f (x) > b) and f (x) b n = E[I (f (x) > b) (g (x i) g (x)) xi x h d K =f (x)]: h To control the bias a certain type of kernel function needs to be used. The following \higher order kernel" by Bartlett (1963) is a standard device in the literature. Let j0 = 1 if j = 0 and 0 for any other integer value j. K`, ` 1 is the class of symmetric functions k : R! R around zero such that Z 1 1 tj k (t) dt = j0 for j = 0; 1; :::; ` 1

16 and for some " > 0 lim k (t) = 1 + jtj`+1+" < 1: jtj!1 Dimension d kernel function K of order ` is constructed by K (t 1 ; :::; t d ) = k (t 1 ) k (t d ) for k 2 K`. In order to improve the order of bias by the higher order kernel, the underlying density is required to be smooth accordingly. The following notion of smoothness is used by Robinson (1988). Let [] denote the largest integer not equal or larger than. G, > 0, > 0, is the class of functions g : R d! R satisfying: g is []-times partially dierentiable for all z 2 R d ;

17 for some > 0, n y2 ky sup o zk R d< jg (y) g (z) Q (y; z)j ky zk R d h (z) for all z; Q = 0 when [] = 0; Q is a []-th degree homogeneous polynomial in (y z) with coecients the partial derivatives of g at z of orders 1 through [] when [] 1; and g (z), its partial derivatives of order [] and less, and h (z) have nite th moments. Bounded functions are denoted by G 1. Let K be a higher order kernel constructed as above. Robinson (1988) has shown the following results: he E h d K ((z 2 z 1 ) =h) jz 1 f (z 1 ) i 2 = O h 2 when f 2 G 1 for some > 0 and k 2 K []+1.

18 and E n (g (z2 ) g (z 1 )) h d K ((z 2 z 1 ) =h) = O h min(;+1;+) o when f 2 G 1, g 2 G, and k 2 K []+[]+1. The following estimator ^ of E ffg is examined by Schweder (1975), Ahmad (1978) and Hasminskii and Ibragimov (1979): 0 = n 1 n X i=1 h ^f (zi ) i : In this application g (z; ; f) = f (z). Note (z; ; f) =@ = 1 and (z; ; f) =@f = I z where I z evaluates a given function at point z. Since neither derivative depends on or f, condition 4 holds trivially.

19 Condition 5 would require to be restricted to a continuous and bounded class of functions. rg (z; ; f) = 1 so that rg = (z 1 ; 0 ; f 0 0 n2 = 1 h dk 1 + E h (z 2 ; 0 ; f 0 0 n1 = 1 h dk 1 + E h dk z2 z 1 h z2 z 1 h z1 z 2 Thus (z 1 ; 0 ; f 0 ) =@f 0 n2 jz 1 (z2 ; E 0 ; f 0 0 n1jz 1 1 z1 z = E 2 h dk jz 1 +E h ) h z1 z 2 h = 0 and 1 h dk jz 1 jz 2 : z1 z 2 h

20 Therefore n = V ar[ 0 f 0 (z 1 ) 1 z1 z E 2 1 z1 z h dk jz 1 +E 2 h h dk h! 4E n [ 0 0 (z 1 )] 2o : ] Also, Robinson's result allows us to nd conditions under which the asymptotic bias is 0. Another example is the partial linear regression model of Cosslett (1984), Schiller (1984) and Wahba (1984). For x 2 R K, y 2 R, w 2 R d the model is where E ("jw; x) = 0. y = x T 0 + (w) + "

21 Consider an estimator which solves the following equations: 0 = n 1 n X i=1 h yi x 0 i^ ^E (yjw i ) + ^E x 0 jw i ^i ^I i x i where ^I i = I ^f (wi ) > b and I is the indicator function. The following lemma allows us to consider 0 = n 1 X n [y i x 0 i^ ^E (yjw i )+ ^E x 0 jw i ^]Ii x i i=1 instead of the feasible GMM. Let I i = I (f (w i ) > b). Pr at least one of ^I i I i 6= 0! 0 when f 2 G 1, for some > 0, k 2 K []+1, jk (0)j < 1, b is positive and bounded, nh d b 2 = log n! 1, b=h! 1, and when there is no positive probability that f (w i ) = b.

22 Let z = (w; x; y). In this example, so that g (z; ; ) = h y x T 1 (w) 2 (w) T i I x rg (z; ; ) = I x [x 2 (w)] T : Since rg (z; ; ) is linear in and does not depend on, the verication of the conditions are easy. One can verify when b! 0 and E n x [x E (xjw)] T o R K 2 < 1 the following holds: rg = E n I x [x E (xjw)] T o! E n x [x E (xjw)] T o To examine the asymptotic distribution note that the Frechet derivative of g with respect to is

23 (h) = h 1 (w) h 2 (w) T 0 x so that writing u = y E (yjw) and v = x E (xjw) and " = y x T 0 (w) = (z 1 ; 0 ; 0 0 n2 u2 v2 T 0 h d K ((w 2 f (w 1 ) I (f (w 1 ) > b) x (z 2 ; 0 ; 0 0 n1 u1 v1 T 0 h d K ((w 1 f (w 2 ) I (f (w 2 ) > b) x 2 : Thus noting that u = y E (xjw) T 0 (w) = " + v T 0 w 1 ) =h) w 2 ) =h) E ( (z1 ; 0 ; 0 0 n2jz 1 = 0

24 E ( (z2 ; 0 ; 0 0 n1jz 1 = " 1 E[ h d K ((w 1 w 2 ) =h) I (f (w 2 ) > b) x 2 jw 1 ] f (w 2 ) so that n = V ar[" 1 [x 1 I 1 E[ h d K ((w 1 w 2 ) =h) I (f (w 2 ) > b) x 2 jw 1 ]]] f (w 2 )! V ar [" 1 [x 1 E (x 1 jw 1 )]] as b! 0: 6 Conclusion More examples need to be worked out. Analogous results for series estimators should be established.

25 First stage expansion result can be obtained including parameters so that conditions can be simplied. general M-estimator is considered in the joint work with Simon Lee (UCL) without assuming smoothness.

Garrett: `Bernstein's analytic continuation of complex powers' 2 Let f be a polynomial in x 1 ; : : : ; x n with real coecients. For complex s, let f

Garrett: `Bernstein's analytic continuation of complex powers' 2 Let f be a polynomial in x 1 ; : : : ; x n with real coecients. For complex s, let f 1 Bernstein's analytic continuation of complex powers c1995, Paul Garrett, garrettmath.umn.edu version January 27, 1998 Analytic continuation of distributions Statement of the theorems on analytic continuation