 # Econ 742: Introductory Econometrics (2)

```Econ 742: Introductory Econometrics (2)
Chapter 3: Large Sample Analysis, the Method of Maximum Likelihood,
Nonlinear Regression, and Asymtotic Tests
3.0 Classical Asymptotic Theory
Some basic asymptotic concepts and useful results:
p
DeÞnition: Convergence in probability Xn −→ X or plimXn = X.
A sequence of random variables {Xn }is said to converge to a random variable X in probability if
lim P (|Xn − X| > ²) = 0,
n→∞
Equivalently, limn→∞ P (|Xn − X| ≤ ²) = 1,
for all ² > 0.
for all ² > 0.
d
DeÞnition: Convergence in distribution Xn −→ X.
A sequence {Xn } is said to converge to X in distribution if the distribution function Fn of Xn converges
to the distribution function F of X at every continuity point of F . (F is called the limiting distribution of
{Xn }).
Chebyshev’s inequality: For any random variable X with mean µ and variance σ 2 ,
P (|X − µ| ≥ λσ) ≤
1
,
λ2
λ > 0.
Continuity mapping theorems:
p
Proposition: Given g : Rk → Rl and any sequence {Xn } such that Xn → α where α is a k × 1 constant
p
vector, if g is continuous at α, then g(Xn ) → g(α).
d
d
Proposition: If g is a continuous function and Xn → X, then g(Xn ) → g(X).
a).
Proposition: (Slutsky) Let {Xn , Yn }, n = 1, 2, · · · be a sequence of pairs of random variables. Then
p
d
p
Xn → X, Yn → 0 ⇒ Xn Yn → 0.
b).
d
p
d
d
d
Xn → X, Yn → c ⇒ Xn + Yn → X + c, Xn Yn → cX, Xn /Yn → X/c,
if c 6= 0.
Law of Large Numbers
Proposition: (Kolmogorov theorem S.L.L.N) Let {Xi }, i = 1, 2, . . . be a sequence of independent random
variable such that E(Xi ) = µi , V (Xi ) = σi2 . Then
∞
X
σ2
i
i=1
i2
a.s
¯n − µ
< ∞ =⇒ X
¯ n → 0.
Proposition: (Chebyshev’s theorem W.L.L.N) Let E(Xi ) = µi , V (Xi ) = σi2 , cov(Xi , Xj ) = 0, i 6= j.
Then
N
1 X 2
p
¯n − µ
σi = 0 =⇒ X
¯n → 0,
lim
n→∞ N 2
i=1
¯n =
where X
1
N
PN
i=1
Xi and µ
¯n =
Central Limit Theorems
1
N
PN
i=1
µi .
1
Theorem: (Lindberg-Levy Theorem C.L.T.) Let X1 , X2 , · · · be a sequence of i.i.d. random variables such
that E(Xn ) = µ and V (Xn ) = σ 2 6= 0 exist. Then the d.f. of Yn ,
Yn =
√
¯ n − µ)/σ → Φ(x),
n(X
d
i.e., Yn → N (0, 1),
where Φ is the standard normal distribution.
Theorem: (Liapounov’s theorem) Let {XnP
} be a sequence of independent
random variables. Let E(Xn ) =
n
1 Pn
2+δ
µn , E(Xn − µn )2 = σn2 6= 0. Denote Cn = ( i=1 σi2 )1/2 . If C 2+δ
E|X
→ 0 for a positive
k − µk |
k=1
n
Pn
(Xi −µi ) d
i=1
→ N (0, 1).
δ > 0, then
Cn
3.1 Large Sample Results for The Linear Regression Model
Features: No normality assumption is needed. Instead appropriate assumptions are assumed such that
certain laws of large numbers and central limit theorems are applicable.
• Consistency:
1. plimn→∞ n1 X 0 ² = 0 under the assumption that limn→∞ n1 X 0 X = Q exists.
Pn
2
Proof: n1 X 0 ² = n1 i=1 x0i ²i . E(x0i ²i ) = x0i E(²i ) = 0 and Var[ n1 X 0 ²] = σn X 0 X/n. The result follows
because the variance of n1 X 0 ² goes to zero.
2. Assuming that Q is invertible, plimn→∞ βˆ = β + Q−1 plimn→∞ n1 X 0 ² = β.
ˆ
• Asymptotic distribution of β:
1. Applying an appropriate CLT (under some regularity conditions on x and ²),
n
1 X 0 d
1
√ X 0² = √
x ²i → N (0, σ 2 Q).
n
n i=1 i
√
2
d
2. n(βˆ − β) → N (0, σ 2 Q−1 ). Hence the asymptotic distribution for βˆ is N (β, σn Q−1 ).
1
2
2
0
3. In practice, n X X estimates Q and σ
ˆ estimates σ .
• Consistency of σ
ˆ2:
plimˆ
σ 2 = plim
²0 ²
²0 X X 0 X −1 X 0 ²
= σ 2 − plim
(
)
= σ 2 − 0 · Q−1 · 0 = σ 2 .
n
n
n
n
ˆ By a mean-value theorem,
• The Delta method: Let α
ˆ = f (β).
√
d
n(α
ˆ − α) → N (0,
∂f (β) 2 −1 ∂f 0 (β)
(σ Q )
).
∂β 0
∂β
• Asymptotic distributions of test statistics:
ˆ −βk
1.) tk = [ˆσ 2 (Xβk0 X)
−1 1/2 is asymptotically normal.
]
kk
σ 2 is asymptotically χ2 (J).
2.) The JFJ,n−K = (Rβˆ − q)0 [R(X 0 X)−1 R0 ]−1 (Rβˆ − q)/ˆ
3.2 The Method of Maximum Likelihood
• Log likelihood function: Suppose that y1 , · · · , yn are independent sample observations,
ln L(θ|y) =
n
X
ln f (yi , θ),
i=1
where f (yi , θ) is the density of yi (if y is continuous r.v. or probability if discrete r.v.)
• Likelihood equation:
∂ ln L(θ|y)
= 0.
∂θ
2
• information³ inequality:
´ E(ln f³(y, θ)) ´< E(ln f (y, θ0 )) for θ 6= θ0 . This follows from the Jensen’s
f (y,θ)
(y,θ)
inequality that E ln f (y,θ0 ) < ln E ff(y,θ
= 0 for θ 6= θ◦ .
0)
(Lemma: (Jensen’s inequality)
Let g : R → R be a convex function on an interval B ⊆ R and let X be a random variable such that
P (X ∈ B) = 1 and E(X) = µ. Then g(E(X)) ≤ E(g(X)). If g is concave on B, then g(E(X)) ≥ E(g(X)).
• A likelihood equality
¸
· 2
¸
∂ ln L(θ|y)
∂ ln L(θ|y) ∂ ln L(θ|y)
+E
= 0.
E
∂θ
∂θ 0
∂θ∂θ 0
·
R
This follows from taking derivatives with θ using the identity f (yi , θ)dyi = 1 for all possible θ.
Under the regularity conditions that the diﬀerentiation operator and the integration operator are interchangeable,
Z
Z
Z
∂L
∂
Ldy =
dy = 0.
Ldy = 1 ⇒
∂θ
∂θ
This implies
Eθ
µ
∂ ln L
∂θ0
¶
=
Z µ
1 ∂L
L ∂θ 0
¶
Z
Ldy =
∂L
dy = 0,
∂θ0
and
¶
µ
¶
µ ·
¸¶
∂ 2 ln L
∂ ∂ ln L
∂ 1 ∂L
=
E
=
E
θ
θ
∂θ∂θ0
∂θ ∂θ0
∂θ L ∂θ0
µ
¶
µ
¶ Z
µ
¶
−1 ∂L ∂L
∂ ln L ∂ ln L
∂2L
∂ ln L ∂ ln L
1 ∂2L
= −Eθ
+
,
= Eθ
+
dy = −Eθ
L2 ∂θ ∂θ0
L ∂θ∂θ0
∂θ ∂θ0
∂θ∂θ0
∂θ ∂θ 0
Eθ
because
R
µ
∂2L
∂θ∂θ0 dy
= 0.
• Information matrix is I(θ) where
I(θ) = E
·
¸
∂ ln L(θ|y) ∂ ln L(θ|y)
.
∂θ
∂θ0
• Large sample properties of the MLE
1.) consistency: plimn→∞ θˆM L = θ.
2.) asymptotic normality: θˆM L is asymptotically N (θ, I −1 (θ)).
Proof: By a Taylor expansion, 0 =
√
By the law of large number,
1
n
∂ ln L(θˆML )
∂θ
n(θˆM L − θ) =
Pn
i=1
µ
=
∂ ln L(θ)
∂θ
1 ∂ 2 ln L(θ∗ )
n ∂θ∂θ0
∂ 2 ln L(θ) p
→
∂θ∂θ 0
E( ∂
2
+
∂ 2 ln L(θ∗ ) ˆ
(θM L
∂θ∂θ 0
¶−1
ln f (y,θ)
)
∂θ∂θ
− θ). Therefore,
1 ∂ ln L(θ)
√
.
n
∂θ
and by a central limit theorem,
f (y,θ) ∂ ln f (y,θ)
)). The Þnal result follows from the likelihood equality.
N (0, E( ∂ ln ∂θ
∂θ0
∂ ln L(θ) d
√1
→
∂θ
n
3.) asymptotic eﬃcient — it achieves the Cramer-Rao lower bound for consistent estimators (uniform convergence in distribution over any compact set of the parameter).
• Estimates of the Variance of MLE: The asymptotic variance I −1 (θ) can be estimated by
Ã
∂ 2 ln L(θˆM L )
−
∂θ∂θ 0
3
!−1
.
For independent sample, it can also be estimated by
Ã
n
X
∂ ln f (yi , θˆM L ) ∂ ln f (yi , θˆM L )
∂θ 0
∂θ
i=1
!−1
.
• (Cramer-Rao Lower Bound)
Under some regularity conditions, the variance of an unbiased estimator of θ will be at least as large as
I −1 (θ).
ln L
), and unbiasedness of
This follows from the postive deÞniteness of the variance matrix of (θ˜ − θ, ∂ ∂θ
∂
ln
L
˜ The latter implies that the covariance E((θ˜ − θ) 0 ) = I.
θ.
∂θ
i
h
¡ ∂ ln L ∂ ln L ¢
ln L
˜
. The covariance matrix of θˆ and
Proof: Let P = V (θ), R = E ∂θ ∂θ0 , and Q = E (θ˜ − θ) ∂∂θ
0
¶
µ
P Q
∂ ln L
, which is nonnegative deÞnite. By premultiplying and postmultiplying this covariance
is
∂θ
Q0 R
matrix by [I, −QR−1 ] and its transpose, it follows that P − QR−1 Q0 ≥ 0. This is so, by pre- and postmulplying this matrix with (I, −QR−1 ) and its transpose,
µ
µ
¶
¶
P Q
P − QR−1 Q0
−1 0
−1
(I, −QR−1 )
)
=
(I,
−QR
)
(I,
−QR
= P − QR−1 Q0 .
0
Q0 R
˜ = θ implies that
The matrix Q equals an identity matrix. This is so, since Eθ (θ)
I=
∂
(
∂θ
Z
˜
θL(y,
θ)dy) =
Z
∂L(y, θ)
dy,
θ˜
∂θ0
¶ Z
µ
∂L
∂ ln L
˜
= θ˜ 0 dy = I.
Q=E θ
0
∂θ
∂θ
Hence P − R−1 ≥ 0.
Q.E.D.
3.3 Maximum Likelihood Estimation of The Linear Regression Model under Normality
Assumption: ² is N (0, σ 2 In ).
• Log likelihood function of y given X is
ln L = −
n
1
n
ln 2π − ln σ 2 − 2 (y − Xβ)0 (y − Xβ).
2
2
2σ
• Likelihood equations:
1
∂ ln L
= 2 X 0 (y − Xβ) = 0,
∂β
σ
∂ ln L
−n
1
=
+ 4 (y − Xβ)0 (y − Xβ) = 0.
∂σ 2
2σ 2
2σ
2
0
• MLE: βˆM L = (X 0 X)−1 X 0 y; σ
ˆM
L = e e/n.
• The information matrix of the linear regression model:
µ 2 0 −1
σ (X X)
2
I(β, σ ) = −
00
This follows from
Ã
∂ 2 ln L
∂β∂β 0
∂ 2 ln L
∂σ 2 ∂β 0
∂ 2 ln L
∂β∂σ2
∂ 2 ln L
∂σ 2 ∂σ 2
!
=
µ
4
− σ12 X 0 X
− 2σ1 4 ²0 X
0
2σ 4 /n
¶
.
− 2σ1 4 X 0 ²
n
²0 ²
2σ4 − σ 6
¶
,
by taking expectations.
• The MLE βˆM L and the least squares estimate βˆ are identical. It is the eﬃcient unbiased estimator as
it attains the Cramer-Rao variance bound.
2
• The MLE σ
ˆM
L is biased downward.
2
2
2
E(ˆ
σM
L ) = (n − K)σ /n < σ
for any Þnite sample size n.
3.4 Numerical Methods
Newton-Raphson Method
This method is applicable for either maximization or minimization problems. Let Qn (θ) be the object
function for optimization. Let θˆ1 be an initial estimate of θ. By a quadratic approximation, Qn (θ) ≡
2
ˆ
Qn (θˆ1 )
n (θ 1 )
Qn (θˆ1 ) + ∂Q∂θ
(θ − θˆ1 ) + 12 (θ − θˆ1 )0 ∂ ∂θ∂θ
(θ − θˆ1 ). Maximizing (or minimizing) the right-hand side
0
0
approximation provides a second-round estimator θˆ2 ,
"
∂ 2 Qn (θˆ1 )
θˆ2 = θˆ1 −
∂θ∂θ0
#−1
∂Qn (θˆ1 )
.
∂θ
The iteration is to be repeated until the sequence {θˆj } converges. Other modiÞcation of this algorithm is
ˆ
n (θ 1 )
to replace ∂Q
∂θ∂θ0 by a negative deÞnite matrix in each iteration (e.g., Quadratic Hill Climbing algorithm).
The step sizes in the iteration can also be modiÞed as
"
∂ 2 Qn (θˆ1 )
θˆ2 = θˆ1 − λ
∂θ∂θ 0
#−1
∂Qn (θˆ1 )
,
∂θ
which λ is a scalar.
Asymptotic Properties of the Second-round Estimator
√
Suppose that θˆ1 is a consistent estimator of θ◦ such that n(θˆ1 − θ◦ ) has a proper distribution, then
∗
n (θ )
the second-round estimator θˆ2 has the same asymptotic distribution as a consistent root θ∗ of ∂Q∂θ
. By
the mean-value theorem,
¯
∂Qn (θˆ1 )
∂Qn (θ◦ ) ∂ 2 Qn (θ)
(θˆ1 − θ◦ ),
=
+
∂θ
∂θ
∂θ∂θ0
where θ¯ lies between θˆ1 and θ◦ . It follows that


#−1 ·
"
¸
2
2
ˆ1 )
¯
√
√
Q
(
θ
(θ
)
Q
(
θ)
∂Q
∂
∂
n
n
n
◦
n(θˆ2 − θ◦ ) = n ˆ
(θˆ1 − θ◦ ) 
+
θ1 − θ◦ −
∂θ∂θ 0
∂θ
∂θ∂θ 0


#−1
#−1
"
"

2
¯ √
ˆ1 )
1
1 ∂ 2 Qn (θ)
Q
(
θ
1 ∂Qn (θ◦ )
∂
1 ∂ 2 Qn (θˆ1 )
n
√
n(θˆ1 − θ◦ ) −
.
= I−

n ∂θ∂θ0
n ∂θ∂θ0 
n ∂θ∂θ0
n
∂θ
Since the Þrst term on the right-hand side converges to zero in probability, we conclude that
¶−1
µ
√
1 ∂Qn (θ◦ ) √ ∗
1 ∂ 2 Qn (θ◦ )
√
n(θˆ2 − θ◦ ) ≡ − plim
≡ n(θ − θ◦ ),
0
n ∂θ∂θ
n
∂θ
i.e., the second round consistent estimator has the same asymptotic distribution as the extremum estimator.
5
3.5: Asymptotic Tests
The null hypothesis is h(θ) = 0, where h : Rk → Rq with q < k. Let θˆ be the unconstrained MLE and
θ¯ the constrained MLE.
• Likelihood ratio test statistic:
¸
·
maxh(θ)=0 L(θ|y)
ˆ − ln L(θ|y)].
¯
= 2[ln L(θ|y)
−2 ln
maxθ L(θ|y)
• Wald test statistics:

ˆ
ˆ  ∂h(θ)
h0 (θ)
∂θ0
Ã
• Eﬃcient Score test (Rao) statistics:
¯
∂ ln L(θ)
0
∂θ
ˆ
∂ 2 ln L(θ)
−
∂θ∂θ 0
!−1
−1
ˆ
∂h(θ)
ˆ
 h(θ).
∂θ
µ 2
¯ ¶−1 ∂ ln L(θ)
¯
∂ ln L(θ)
.
−
0
∂θ∂θ
∂θ
• All these three test statistics are asymptotically chi-square distributed with q-degrees of freedom.
The eﬃcient score test statistics has the same limiting distribution as the likelihood ratio test. This
follows from the arguments.
¯
ˆ
∂ ln L(θ)
∂ ln L(θ)
∂ 2 ln L(θ∗ ) ¯ ˆ
(θ − θ),
=
+
∂θ
∂θ
∂θ∂θ0
and
¯ − ln L(θ)
ˆ =
ln L(θ)
2
2
∗
∗
ˆ
∂ ln L(θ)
ˆ = 1 (θ¯ − θ)
ˆ
ˆ + 1 (θ¯ − θ)
ˆ 0 ∂ ln L(θ ) (θ¯ − θ)
ˆ 0 ∂ ln L(θ ) (θ¯ − θ),
(θ¯ − θ)
0
0
∂θ
2
∂θ∂θ
2
∂θ∂θ
therefore,
¯
∂ ln L(θ)
0
∂θ
µ 2
2
¯ ¶−1 ∂ ln L(θ)
¯ D
∂ ln L(θ)
D
ˆ 0 ∂ ln L(θ0 ) (θ¯ − θ)
ˆ =
ˆ − ln L(θ)).
¯
−
= −(θ¯ − θ)
2(ln L(θ)
0
∂θ∂θ
∂θ
∂θ∂θ 0
3.6: Nonlinear Regression Models
A nonlinear regression model is
yi = h(xi , β) + ²i ,
where E(²|x) = 0. A standard case assumes the ²s have the similar properties in the standard linear regression
model.
• The method of nonlinear least squares:
min
β
• Normal equations:
n
X
i=1
2
• Consistent estimator of σ :
n
X
i=1
(yi − h(xi , β))2 .
[yi − h(xi , β)]
n
∂h(xi , β)
= 0.
∂β
1X
ˆ 2.
[yi − h(xi , β)]
σ
ˆ =
n i=1
2
6
Pn
√
d
• asymptotic distribution: n(βˆNL − β) −→ N (0, σ 2 C −1 ), where C = plim n1 i=1
Pn
Proof: Let S(β) = i=1 (yi − h(xi , β))2 . By a Taylor series expansion,
∂h(xi ,β) ∂h(xi ,β)
.
∂β
∂β 0
1 ∂S(βˆN L )
1 ∂S(β)
1 ∂S(β ∗ ) √ ˆ
0= √
· n(βN L − β),
=√
+
n
∂β
n ∂β
n ∂β∂β 0
which implies that
√
n(βˆNL − β) = −
Furthermore,
·
1 ∂ 2 S(β ∗ )
n ∂β∂β 0
¸−1
1 ∂S(β)
√
.
n ∂β
n
2 X ∂h(xi , β) d
1 ∂S(β)
√
²i
→ N (0, 4σ 2 C)
= −√
∂β
n ∂β
n i=1
and
n
n
1 ∂ 2 S(β)
1 X ∂h(xi , β) ∂h(xi , β)
1 X ∂ 2 h(xi , β) p
=2
²i
−2
→ 2C.
0
0
n ∂β∂β
n i=1
∂β
∂β
n i=1
∂β∂β 0
Gauss-Newton Method
This method is applicable to the estimation of a nonlinear regression equation: yi = fi (β) + ui .
ˆ
i (β1 )
Let βˆ1 be an initial estimate of β◦ . By a Taylor series expansion, fi (β) ≡ fi (βˆ1 ) + ∂f∂β
(β − βˆ1 ) and
0
Sn (β) =
n
X
i=1
2
(yi − fi (β)) ≡
n
X
i=1
Ã
!2
ˆ1 )
(
β
∂f
i
yi − fi (βˆ1 ) −
(β − βˆ1 ) .
∂β 0
By minimizing the right-hand quadratic approximation with respect to β,
βˆ2 = βˆ1 +
!−1 n
Ã n
X ∂fi (βˆ1 ) ∂fi (βˆ1 )
X ∂fi (βˆ1 )
(yi − fi (βˆ1 ))
0
∂β
∂β
i=1
i=1
!−1 n
Ã n
X ∂fi (βˆ1 ) ∂fi (βˆ1 )
X ∂fi (βˆ1 )
∂fi (βˆ1 ) ˆ
(yi − fi (βˆ1 ) +
=
β1 ).
0
∂β
∂β
∂β
∂β 0
i=1
i=1
Alternatively, yi ≡ fi (βˆ1 ) +
∂β
∂fi (βˆ1 )
∂β 0 (β
− βˆ1 ) + ui , or equivalently,
∂fi (βˆ1 ) ˆ
∂fi (βˆ1 )
β + ui .
yi − fi (βˆ1 ) +
β1 =
0
∂β
∂β 0
The second round estimator can be interpreted as the least squares estimation applied to the above equation. The second-round estimator of the Gauss-Newton
iteration is asymptotically as eﬃcient as the NLLS
√
estimator if the iteration is started from an n-consistent estimator.
7
``` # ECG751 - Econometric Methods - Fall 2014 Instructor: Denis Pelletier # Math 3080 § 1. First Midterm Sample Rroblems Name: Sample # Econometrics Lecture 4: Maximum Likelihood Estimation and Large Sample Tests 1 