Estimation of Heterogeneous Panels with Structural Breaks"

Transcription

Estimation of Heterogeneous Panels with Structural
Breaks
Badi H. Baltagiy
Qu Fengz
Chihwa Kaox
March 27, 2015
Abstract
This paper extends Pesaran’s (2006) work on common correlated e¤ects (CCE)
estimators for large heterogeneous panels with a general multifactor error structure
by allowing for unknown common structural breaks. Structural breaks due to new
policy implementation or major technological shocks, are more likely to occur over
a longer time span. Consequently, ignoring structural breaks may lead to inconsistent estimation and invalid inference. We propose a general framework that includes
heterogeneous panel data models and structural break models as special cases. The
least squares method proposed by Bai (1997a, 2010) is applied to estimate the common change points, and the consistency of the estimated change points is established.
We …nd that the CCE estimator have the same asymptotic distribution as if the true
change points were known. Additionally, Monte Carlo simulations are used to verify
the main results of this paper.
Keywords: Heterogeneous Panels, Cross-sectional Dependence, Structural Breaks,
Common Correlated E¤ects.
JEL Classi…cation: C23, C33
The authors would like to thank the editor Han Hong, the associate editor and three anonymous referees
for their constructive comments and suggestions. We are also grateful to helpful comments from Jushan Bai,
Jiti Gao, Kazuhiko Hayakawa, Dukpa Kim, Tatsushi Oka, Richard Smith, Liangjun Su, and participants of
a seminar at National University of Singapore and from 2013 Asian Meeting of Econometric Society, 2014
Tsinghua Econometrics Conference, 2014 China Meeting of Econometric Society, 2014 International Panel
Data Conference. Financial support from the MOE AcRF Tier 1 Grant M4011314 at Nanyang Technological
University is gratefully acknowledged.
y
Corresponding author. Department of Economics and Center for Policy Research, 426 Eggers Hall,
Syracuse University, Syracuse, NY 13244-1020, USA, and Department of Economics, Leicester University,
University Road, Leicester LE17 6EE, UK. Email Address: [email protected].
z
Division of Economics, School of Humanities and Social Sciences, Nanyang Technological University.
HSS-04-48, 14 Nanyang Drive, Singapore 637332. Email Address: [email protected].
x
Department of Economics and Center for Policy Research, 426 Eggers Hall, Syracuse University, Syracuse, NY 13244-1020, USA. Email Address: [email protected].
1
Introduction
For panel data models, the presence of cross-sectional dependence due to unobservable common factors or spatial spillover e¤ect is a major concern in estimation and inference. It could
lead to invalid inference and inconsistent estimators, see Lee (2002) and Andrews (2005).1
Several tests for cross-sectional dependence in panel data models have been proposed in the
literature. These include Pesaran (2004, 2012), Ng (2006), Pesaran, Ullah and Yamagata
(2008), Sara…dis, Yamagata and Robertson (2009), Chen, Gao and Li (2011), Hsiao, Pesaran and Pick (2012), Baltagi, Feng and Kao (2011, 2012), Halunga, Orme and Yamagata
(2011), Juhl (2011), and Su and Zhang (2011), to mention a few. To deal with cross-sectional
dependence in panels, two general estimation methods have been proposed including spatial estimation methods (Anselin, 1988; Kelejian and Prucha, 1999; Kapoor, Kelejian and
Prucha, 2007; Lee, 2007 and Lee and Yu, 2010, to name a few), and factor models (see
Pesaran, 2006; and Bai, 2009, to name a few).
In particular, Pesaran (2006) develops common correlated e¤ects (CCE) estimators for
large heterogeneous panels with a general multifactor error structure by least squares using
augmented data. The common correlated e¤ects (factors) can be asymptotically partialled
out by means of the cross-sectional average of the dependent variable and the individualspeci…c regressors when the cross-section dimension is large. Kapetanios, Pesaran and Yamagata (2011) show that the CCE estimator can be extended to the case of nonstationary
unobserved common factors. Additionally, the CCE approach is also shown to be applicable
to situations of spatial and other forms of weak cross-sectional dependent errors (Pesaran
and Tosetti, 2011; Chudik, Pesaran and Tosetti, 2011), and heterogenous dynamic panel
data models with weakly exogenous regressors (Chudik and Pesaran, 2013).2
However, this literature assumes that the slope coe¢ cients are constant over time. This
implicit assumption is common in the literature on panel data models with large time dimension, see for example, Kao (1999), Phillips and Moon (1999), Hahn and Kuersteiner
(2002), Alvarez and Arellano (2003), Phillips and Sul (2007), Pesaran and Yamagata (2008),
Hayakawa (2009), to name a few. Due to policy implementation or technological shocks,
1
Baltagi and Pirotte (2010) show that tests of hypotheses based on standard panel data estimators that
ignore spatial dependence lead to misleading inference.
2
In a panel with unobserved common factors, Banerjee and Carrion-i-Silvestre (2011) suggest a test for
panel cointegration based on a pooled CCE estimator of the coe¢ cients.
1
structural breaks are possible especially for panels with a long time span. Consequently,
ignoring structural breaks may lead to inconsistent estimation and invalid inference.
This paper extends Pesaran’s (2006) heterogeneous panels by allowing for unknown common structural breaks in the slopes. This is useful for example when global technological or
…nancial shocks a¤ect all markets or …rms at the same time. Since the framework of heterogeneous panels is fairly general and includes popular panel data models as special cases, it
allows us to examine the issue of structural breaks in panel data models in a less restrictive
way.
By considering both cross-sectional dependence and structural breaks in a general panel
data model, this paper also contributes to the change point literature in several ways. First,
it extends Bai’s (1997a) time series regression model to heterogeneous panels, showing that
the consistency of estimated change points can be achieved with the information along the
cross-sectional dimension. This result con…rms the …ndings of Bai (2010) and Kim (2011).
Second, it also enriches the analysis of common breaks of Bai (2010) and Kim (2011) in a
panel mean-shift model and a panel deterministic time trend model by extending them to
a regression model using panel data. This makes it possible to allow for structural breaks
and cross-sectional dependence in empirical work using panel regressions. In particular,
our methods can be applied to regression models using large stationary panel data, such as
country-level panels and state/provincial-level panels.3
Regarding estimating common breaks in panels, Feng, Kao and Lazarova (2009) and
Baltagi, Kao and Liu (2012) also show the consistency of the estimated change point in a
simple panel regression model. Hsu and Lin (2012) examine the consistency properties of
the change point estimators for nonstationary panels. More recently, Qian and Su (2014)
and Li, Qian and Su (2014) study the estimation and inference of common breaks in panel
data models with and without interactive …xed e¤ects using Lasso-type methods. In terms of
detecting structural breaks in panels, some recent literature includes Horváth and Hušková
(2012) in a panel mean shift model with and without cross-sectional dependence, De Wachter
and Tzavalis (2012) in dynamic panels, and Pauwels, Chan and Mancini-Gri¤oli (2012) in
heterogeneous panels, to name a few.
The paper is organized as follows. Section 2 introduces heterogeneous panels with a
3
Some empirical examples include Fleisher, Li and Zhao (2010) using Chinese provincial-level panel data,
and Huang (2009) using country-level panels, to mention a few.
2
common structural break. Section 3 starts with a simple heterogeneous panel model that
ignores the unobservable common correlated e¤ects. This model can be regarded as a direct
extension of Bai’s (1997a) time series regression model to a panel setup. The least squares
estimation proposed by Bai (1997a) is applied. With the main results established in Section
3, the discussion of the general model with common correlated e¤ects is presented in Section
4. Section 5 brie‡y discusses the case of multiple common breaks. In Section 6, Monte Carlo
simulations are used to verify the consistency of the estimated common change point for
both models considered. Section 7 provides concluding remarks. The mathematical proofs
are relegated to the Appendix.
2
Heterogeneous Panels with a Common Structural
Break
In a heterogeneous panel data model:
yit = x0it
xit is a p
i
+ eit ;
i = 1; :::; N ; t = 1; :::; T;
(1)
1 vector of explanatory variables, and the errors are cross-sectionally correlated,
modelled by a multifactor structure
eit =
where ft is an m
0
i ft
(2)
+ "it ;
1 vector of unobserved factors and
i
is the corresponding loading
vector. "it is the idiosyncratic error independent of xit . However, xit could be a¤ected by
the unobservable common e¤ects ft . Projecting xit on ft , we obtain
xit =
where
i
is an m
0
i ft
+ vit ; i = 1; :::; N ; t = 1; :::; T;
p factor loading matrix. vit is a p
(3)
1 vector of disturbances. Due to
the correlation between xit and eit , OLS for each individual regression could be inconsistent.
Thus, Pesaran (2006) develops the CCE estimator of
i
by least squares using augmented
data.4
4
For simplicity, observed common e¤ects like seasonal dummies are not included in (1), but they can be
easily handled as in Pesaran (2006).
3
In this paper, we allow for structural breaks to occur in some or all components of the
slopes
5
i.
Following Bai (2010) and Kim (2011), a structural break at a common unknown
date k0 is assumed. This could be due to a macro policy implementation or a technological
shock that a¤ects all markets or …rms at the same time. More formally,
yit = x0it i (k0 ) + eit ; i = 1; :::; N ; t = 1; :::; T;
where some or all components of
i (k0 )
(4)
are di¤erent before and after the date k0 .6 Following
Bai (1997a), this structural break model can be written as
x0it
x0it
yit =
+ eit ;
t = 1; :::; k0 ;
0
+
z
+
e
;
t
= k0 + 1; :::; T;
it
i
it i
i
i = 1; :::; N , where zit = R0 xit denotes a q
the q
q identity matrix with q
(5)
1 subvector of xit with R0 = (0q
(p q) ; Iq ).
Iq is
p. The case where q < p denotes a partial change model,
while the case where q = p is for a pure change model. Pauwels, Chan and Mancini-Gri¤oli
(2012) propose a testing procedure for k0 in this setting.
Substituting zit = R0 xit in (5), we obtain
i (k0 )
so that
2i
1i
=
i
+R
i
= R i , and
1i
1ft > k0 g =
i
2i
=
=
i;
t = 1; :::; k0 ;
i + R i ; t = k0 + 1; :::; T;
denotes the slope jump for i. When
i
= 0 there is no
structural break in the slope.
The case of multiple break points will be discussed in Section 5. In the next two sections,
we consider the simple case of one common break as in model (5). Compared with the
heterogeneous panel data model considered in Pesaran (2006), (5) has the extra component
R
i
1ft > k0 g in the slope, involving the unknown structural change point k0 . Thus,
ignoring the structural break in the slopes may invalidate the CCE estimator proposed by
5
Pesaran (2004) discusses testing for cross-sectional dependence in a heterogeneous panel model with
structural breaks. Kapetanios, Pesaran and Yamagata (2011) examine the performance of the CCE estimator
in case of a structural break in the mean of the unobserved factors using Monte Carlo experiments.
6
As shown in Section 4, to apply the CCE approach, the common break assumption is required. In Kim
(2014), the common break assumption is generalized to handle heterogeneous responses to a common shock
that follow a common distribution. Liao and Wang (2012) also assume a common distribution, instead of a
common break date, and estimate individual-speci…c structural breaks and their cross-sectional distribution
using Bayesian methods. In addition, Li, Chen and Gao (2011) consider a time-varying coe¢ cient panel
data model where the slope coe¢ cient is allowed to be di¤erent for each time period, e.g., (t), but common
for all individuals.
4
Pesaran (2006).7 Compared with the simple mean shift panel data model in Bai (2010), our
model is enriched by adding a regression structure with xit 6= 1 in general, as well as cross-
sectional dependence characterized by a multifactor structure in the errors. When there are
no unobservable common factors ft , our model (4) can also be regarded as an extension of
Bai (1997a) to a panel data setting. In addition, the model (4) above is similar to Kim
(2011), who considered the case of a deterministic time trend with a common break.
Before proceeding to the general model (5), we start with a simple case of heterogeneous
panels in the absence of common correlated e¤ects ft in Section 3, and then extend the main
results to the general case in Section 4.
To estimate the common change point k0 , we need the following additional assumptions:
Assumption 1 k0 = [ 0 T ], where
0
2 (0; 1) and [ ] is the greatest integer function.
Note that unlike the time series model considered by Bai (1997a), the restriction of
0
2 (0; 1) is unnecessary in a panel mean shift setup considered by Bai (2010) as long
as T =N ! 0 . However, this assumption is required in our heterogeneous panels with
general regressors. Enough observations are needed to consistently estimate the slopes in
each regime.
De…ne
thus
N
=
PN
i=1
0
i i.
For series i,
0
i i
measures the magnitude of the structural break,
is an indicator of the break magnitude for all N series sharing a common break.
q
T
T
N
Assumption 2 N ! 1 and (i) N is bounded as N ! 1; (ii) N N ! 1 and N N
!
N
1 as (N; T ) ! 1.
i
could be random with a …nite variance across i, with Assumption 2(i) describing this
case. When
i
is considered as random, Assumption 2 means that
bounded in part (i), and that
natively,
i
N
NT
and
p
N
p
N T
N
N
is stochastically
converge in probability to 0 in part (ii). Alter-
could denote …xed parameters. Since Assumption 2(i) allows for the case where
! 0 as N ! 1, Assumption 2(ii) implies that it cannot converge to 0 too fast. Conse-
N
N
quently, Assumption 2(i) imposes an upper bound on
lower bound on
N
N
N
N
, while Assumption 2(ii) imposes a
.
7
As documented in the time series literature, e.g., Pesaran and Timmermann (2002), structural breaks
could lead to forecast failure. Stock and Watson (2002, 2009) show that forecasts constructed using dynamic
factor models are robust to small structural breaks of factor loadings. However, when the size of the break
is big, ignoring structural breaks may yield biased and inconsistent forecasts, as illustrated by Banerjee,
Marcellino and Masten (2008), Breitung and Eickmeier (2011), and Chen, Dolado and Gonzalo (2014).
5
In case T grows at a comparable rate or faster than N , i.e., T = O(N ) with
Assumption 2(ii) implies that
can diverge at any rate. When
N
N
1,
increases at a rate less
than N , Assumption 2(ii) allows for the possibility of no structural break in some series.8
3
Model 1: No Common Correlated E¤ects
In this section, we assume that there are no unobserved common e¤ects ft in the errors and
regressors. Or the loading vectors
i
x0it
x0it
i
yit =
and
i
are equal to zero. For i = 1; :::; N;
+ "it ;
t = 1; :::; k0 ;
0
+
z
+
"
;
t
= k0 + 1; :::; T:
it
i
it i
(6)
This is the special case of cross-sectionally independent errors, where a common break k0
occurs in the heterogeneous slopes. This model generalizes Bai (1997a), Bai (2010) and
Pesaran and Smith (1995). When N = 1, (6) is the time series model considered in Bai
(1997a). When xit = 1, this model reduces to the one in Bai (2010). In case the lagged
dependent variable is included in xit and
i
= 0, (6) turns out to be the setup in Pesaran
and Smith (1995).9
Assumption 3 (i) The disturbances "it ; i = 1; :::; N; are cross-sectionally independent; (ii)
For each series i ; "it is independent of xit for all i and t; (iii) "it is a stationary process with
absolute summable autocovariances,
"it =
where f
P1
l=0
ail
i;t l
= 1; :::; T g are independent and identically distributed (IID) random variables
P
2
2
with …nite fourth-order cumulants. Assume 0 < V ar("it ) = 1
l=0 ail = i < 1. Also, for
the T
it ; t
1 vector "i = ("i1 ; "i2 ;
; "iT )0 , V ar("i ) =
";i :
When "it is serially uncorrelated, lagged dependent variables are predetermined and can
be included as regressors in (6).
P
P
Assumption 4 For i = 1; :::; N , the matrices (1=j) jt=1 xit x0it , (1=j) Tt=T j+1 xit x0it ,
P0
Pk0 +j
(1=j) kt=k
xit x0it and (1=j) t=k
xit x0it are stochastically bounded and have minimum
0 j+1
0 +1
8
Assumption N ! 1 rules out the case where there is no structural break in the slopes in all series.
Heterogeneous dynamic panel data models with weakly exogenous regressors and unobserved common
factors are studied by Chudik and Pesaran (2013).
9
6
eigenvalues bounded away from zero in probability for all large j. In addition, for each i,
P
(1=T ) Tt=1 xit x0it converges in probability to a nonrandom and positive de…nite matrix as
T ! 1.
This assumption is borrowed from Assumptions A3 and A4 in Bai (1997a).10 Its counterpart across the cross-sectional dimension is also needed.
P Pk0
0
Assumption 5 For any positive …nite integer s, the matrices N1 N
i=1
t=k0 s+1 xit xit and
PN Pk0 +s
1
0
i=1
t=k0 +1 xit xit , i = 1; :::; N , are stochastically bounded, with minimum eigenvalues
N
P
0
bounded away from zero in probability for large N . In addition, for each t, (1=N ) N
i=1 xit xit
is stochastically bounded as N ! 1.
Assumption 6 f i ; i = 1; :::; N g are drawn independently of fxit ; i = 1; :::; N g.
Let bi = ( 0i ; 0i )0 ; i = 1; :::; N; denote the slope parameters. In the random coe¢ cient
model considered by Pesaran and Smith (1995) and Pesaran (2006), we assume:
Assumption 7 For i = 1; :::; N ,
bi = b + vb;i ; vb;i
IID(0;
(7)
b );
v ;i
0
and b =
for i = 1; 2; :::; N , where kbk < 1,
v ;i
0
k b k < 1, and the random deviations vb;i are independent of xit and "jt for all i; j and t.
where b = ( 0 ; 0 )0 , vb;i =
For any matrix or vector A, the norm of A is de…ned as kAk =
p
tr(AA0 ). This as-
sumption is a simpli…ed version of Assumption 4 of Pesaran (2006). Under Assumption
6, f i ; i = 1; :::; N g are not necessarily random. When f i ; i = 1; :::; N g are considered as
random, as part of Assumption 7, Assumption 6 becomes redundant. Under Assumption 7,
6= 0 implies a structural break in the slope.
By (4),
yit = x0it
i
+ x0it R i 1ft > k0 g + "it ;
if the structural break is ignored, the term x0it R i 1ft > k0 g is absorbed in the error term
^"it = x0it R i 1ft > k0 g + "it . This leads to inconsistency of OLS for each series due to
endogeneity. Thus, estimating k0 …rst is essential.
10
As pointed out by Bai (1997a), Bai and Perron (1998, 2003) trending regressors are allowed in the form
of (t=T )l , for (l > 0), or any function of the time trend: g(t=T ).
7
; xiT )0 and "i = ("i1 ; "i2 ;
; yiT )0 , Xi = (xi1 ;
Let Yi = (yi1 ;
; "iT )0 denote the
stacked data and errors for individual i = 1; :::; N over the time periods observed. Similarly,
de…ne Z0i = (0;
; ziT )0 . (6) can be written in matrix form as
; 0; zi;k0 +1 ;
Yi = Xi
The parameters of interest are
i;
i
i
+ Z0i
i
+ "i ;
(8)
i = 1; :::; N:
and the change point k0 . We …rst estimate k0 using least
squares as proposed by Bai (1997a, 2010). For any possible change point k = 1; :::; T
…ne the matrices X2i (k) = (0;
; xiT )0 , and Z2i (k) = (0;
; 0; xi;k+1 ;
; 0; zi;k+1 ;
1, de; ziT )0 .
When k happens to be the true change point k0 , Z2i (k0 ) = Z0i . De…ne X0i = X2i (k0 );
thus Z0i = X0i R. To make the notation more compact, we let Xi (k) = (Xi ; Z2i (k)) and
X0i = (Xi ; Z0i ). Thus, (8) becomes
Yi = Xi
Given any k = 1; :::; T
i
+ Z0i
i
+ "i = X0i bi + "i ;
i = 1; :::; N:
(9)
1, one can estimate bi by least squares,
^bi (k) =
^ (k)
i
î (k)
= [Xi (k)0 Xi (k)] 1 Xi (k)0 Yi , i = 1; :::; N:
(10)
The corresponding sum of squared residuals is given by
SSRi (k) = [Yi
= [Yi
Xi (k)^bi (k)]0 [Yi
Xi ^ i (k)
Xi (k)^bi (k)]
Z2i (k)î (k)]0 [Yi
Xi ^ i (k)
Z2i (k)î (k)],
i = 1; :::; N: Note that both ^bi (k) and SSRi (k) depend on k. For each series i, k0 can be
estimated by arg min1
k T 1
SSRi (k) as in Bai (1997a). Given that the structural break
occurs at a common date for all cross-sectional units in the panel setup, the least squares
estimator of k0 is de…ned as
k^ = arg min
1 k T 1
Weights f
i
2 (0; 1); i = 1; :::; N;
PN
i=1
i
N
X
i SSRi (k):
(11)
i=1
= 1g allow for the possibility of di¤erent magni-
tudes, e.g., di¤erent variances, across series.
When N = 1, k^ de…ned in (11) boils down to the change-point estimator considered by Bai
(1997a) in a time series setting, with k^ k0 = Op (1) for large T . In time series models, only the
8
break fraction
0
= k0 =T , instead of k0 itself, can be consistently estimated. In a multivariate
time series set up, Bai, Lumsdaine and Stock (1998) show that the width of the con…dence
interval of the estimated change point decreases with the number of time series.11 This result
implies that cross-sectional observations with common breaks improve the accuracy of the
estimated change point. In fact, Bai (2010) shows that the least squares estimator of the
change point is consistent in a panel mean-shift model, i.e., k^ k0 = op (1). A similar result is
also obtained by Kim (2011) in a panel deterministic time trend model. In our heterogeneous
panel regression model, (11) combines the information from each series by summing up
SSRi (k). With a large N , k^ uses more information provided by the multiple time series
sharing a common break. Consequently, the panel data estimator k^ is more accurate than
p
the time-series estimator and achieves consistency, i.e., k^ k0 ! 0 as (N; T ) ! 1.
Theorem 1 Under Assumptions 1-6 (or 7), lim(N;T )!1 P (k^ = k0 ) = 1.
The proof of Theorem 1 can be found in the Appendix.
^ the corresponding estimator of the slopes is ^bi =
Given the estimated change point k,
^bi (k);
^ i = 1; :::; N . When bi , i = 1; :::; N , are considered as random variables under Assumption 7, the cross-sectional mean b can be consistently estimated by the mean group estimator
proposed by Pesaran and Smith (1995) and Pesaran (2006):12
N
N
X
X
^bM G = 1
^bi = 1
^ 0 Xi (k)]
^ 1 Xi (k)
^ 0 Yi :
[Xi (k)
i
i
N i=1
N i=1
4
(12)
Model 2: Common Correlated E¤ects
In this section, we extend Model 1 to the general model with common correlated e¤ects (5):
for i = 1; :::; N ,
yit = x0it i (k0 ) + eit =
where eit =
0
i ft
x0it
x0it
+ eit ;
t = 1; :::; k0 ;
0
i + zit i + eit ; t = k0 + 1; :::; T:
i
+ "it . The regressors xit , i = 1; :::; N , are allowed to be correlated with the
unobservable factors ft modelled in (3), xit =
11
0
i ft
+ vit : When
i
= 0, the model reduces to
Change-point estimators in the multivariate time series literature are discussed in Bai (2000) and Qu
and Perron (2006).
12
Note that the pooled estimator of b considered by Pesaran (2006) can be studied similarly. Since the
asymptotic distributions of ^bi and ^bM G are similar to those derived in Bai (1997a) and Pesaran (2006), they
are summarized in the Supplementary Appendix.
9
the one considered by Pesaran (2006). Kim (2011) considers the special case of xit = (1; t)0 .13
In this heterogeneous panel data model with a common break k0 , the parameters of interest
are bi = ( 0i ; 0i )0 , i = 1; :::; N , and the change point k0 . The following assumptions are
needed.
Assumption 8 Common factors ft ; t = 1; :::; T; are covariance sationary with absolute
summable autocovariances, independent of errors "is and vis for all i; s; t:
Assumption 9 Errors "is and vjt are independent for all i; j; s; t: vit , i = 1; :::; N; are linP
ear stationary processes with absolute summable autocovariances, vit = 1
l=0 Sil i;t l ; where
(
it ;
0
it
)0 are (p + 1)
1 vectors of IID random variables with variance-covariance matrix
Ip+1 and …nite fourth-order cumulants, and
V ar(vit ) =
P1
l=0
Assumption 10 Factor loadings
ft for all i; j; t. Assume
i
=
i
+
i = 1; :::N; where the means ,
Sil Sil0 =
and
i;
i
i;v ,
and 0 < k
< 1:
are IID across i, and independent of "jt ; vjt and
IID(0;
i
i;v k
) and
i
=
+
i;
are non-zero and …xed and the variances
IID(0;
i
;
),
are …nite.
Together with Assumptions 3 and 7, Assumptions 8, 9 and 10 given above are the same
as Assumptions 1-3 of Pesaran (2006), with the additional restrictions
6= 0 and
6= 0.
The correlation between xit and eit due to unobserved common factors ft renders OLS
inconsistent. If ft were observable, it could be treated as a regressor, and this correlation can
be removed using a partitioned regression. Let F = (f1 ; f2 ;
orthogonal projection matrix is given by Mf = IT
; fT )0 ; then the corresponding
F (F 0 F ) 1 F 0 : In this case, (5) can be
written in matrix form as
Yi = Xi
i
+ Z0i
i
+F
i
+ "i ;
i = 1; :::; N
(13)
Premultiplying (13) by Mf , we get
Yi = Xi
i
+ Z0i
i
+ "i ; i = 1; :::; N;
(14)
which is of the same form as equation (8) considered in Section 3, with transformed data
Yi = Mf Yi , Xi = Mf Xi = Mf Vi , Z0i (k0 ) = Mf Z0i and "i = Mf "i . For each i = 1; :::; N , the
13
In a similar panel set up without exogenous regressors, Bai and Carrion-i-Silvestre (2009) develop unit
root tests applicable to situations of multiple structural breaks and unobserved common dynamic factors.
10
T
p vector Vi denotes (vi1 ; :::; viT )0 . Conditional on F , (Xi ; Z0i ) and "i are uncorrelated
under Assumption 9.
However, ft ; t = 1; :::; T; are unobservable. To proceed, we follow Pesaran’s (2006) idea
of using the cross-sectional averages of yit and xit as proxies for ft . Combining (5) and (3)
yields
wit
yit
xit
=
(p+1) 1
= Ci (k0 )0 ft + uit (k0 );
(p+1) mm 1
where
Ci (k0 ) = ( i ;
1
0
i (k0 ) Ip
i)
m (p+1)
Note that like
i (k0 ),
(15)
(p+1) 1
and uit (k0 ) =
"it + vit0 i (k0 )
:
vit
the slope Ci (k0 ) in (15) also shifts at k0 .
Ci (k0 ) =
C1i = (
C2i = (
+
i+
i
i 1i ;
i );
t = 1; :::; k0 ;
i ); t = k0 + 1; :::; T:
i 2i ;
(16)
Common break k0 splits the data generating process for all individuals into two regimes,
and each regime has the same structure as that considered in Pesaran (2006). Consequently,
unobserved common factors ft can be partialled out by using cross-sectional averages in the
same spirit.
Let wt =
PN
i=1
i wit
be the cross-sectional averages of wit using weights
i,
i = 1; :::; N .
In particular,
wt = C(k0 )0 ft + ut (k0 )
(17)
PN
P
C1 = i=1 i C1i ; t = 1; :::; k0 ;
P
where C(k0 ) = N
i=1 i Ci (k0 ) =
C2 = N
i=1 i C2i ; t = k0 + 1; :::; T:
The common break assumption is needed, otherwise C(k0 ) is not well de…ned. ut (k0 ) is
de…ned as
ut (k0 ) =
N
P
i uit (k0 ) =
i=1
As in Pesaran (2006), the weights
P
and N
i=1 j i j < 1.
8
>
>
<
>
>
:
i,
"t +
"t +
PN
i=1
v
PN t
i=1
0
i vit 1i
0
i vit 2i
vt
; t = 1; :::; k0 ;
(18)
; t = k0 + 1; :::; T:
i = 1; :::; N , satisfy conditions:
Assumption 11 Rank(C1 ) = Rank(C2 ) = m
11
p + 1:
i
= O( N1 ),
PN
i=1
i
=1
We assume that rank condition is satis…ed. Pesaran (2006) shows that in the case of
de…cient rank, it is impossible to obtain consistent estimators of the individual slope coe¢ cients, but their cross-sectional mean can be consistently estimated. When C(k0 ) is of full
rank, ft can be written as
1
ft = C(k0 )C(k0 )0
C(k0 )(wt
ut (k0 )):
From (16), the matrix C(k0 )C(k0 )0 has two regimes, shifting at k0 ,
C10 C1 ; t = 1; :::; k0 ;
C20 C2 ; t = k0 + 1; :::; T:
C(k0 )C(k0 )0 =
Assumption 11 implies that C(k0 )C(k0 )0 is invertible. As shown in Lemma 1 of Pesaran
(2006), the cross-sectional average of the errors vanish in both regimes as N ! 1, where
PN
P
"t = N
i=1 i vit , yielding
i=1 i "it , vt =
C(k0 )C(k0 )0
ft
1
p
C(k0 )wt ! 0:
(19)
This suggests that it is asymptotically valid to use wt as observable proxies for ft . Let
; wT )0 denote the T
W = (w1 ; w2 ;
T
(p + 1) matrix of cross-sectional averages. Denote the
W (W 0 W ) 1 W 0 : Thus, similar to the result Mf F = 0, by
T matrix Mw by Mw = IT
(19) it is expected that the terms involving Mw F are ignorable asymptotically as N ! 1.
Premultiplying (13) by Mw instead of Mf , we obtain
Mw Yi = Mw Xi
i
+ Mw Z0i
i
+ Mw F
i
+ Mw "i ; i = 1; :::; N .
(20)
~ i = Mw Xi = (~
p matrix X
xi1 ;
; x~iT )0 denote the transformed regressors.
Similarly, de…ne Y~i = Mw Yi ; Z~0i = Mw Z0i and ~"i = Mw "i . Thus, (20) becomes
Let the T
~i
Y~i = X
where ~"0i = Mw F
i
i
+ Z~0i
i
+ Mw F
i
~i
+ ~"i = X
i
+ Z~0i
i
+ ~"0i ; i = 1; :::; N;
(21)
+ ~"i .
Lemma 6 in the Appendix shows that each element of Mw F
i
is of order Op ( p1N ) and
vanishes as (N; T ) ! 1, implying that ~"0i can be treated as ~"i asymptotically. Based on this
intuition, we can follow the procedure proposed in Section 3 to estimate k0 and bi = ( 0i ; 0i )0 ,
~ i ; i = 1; :::; N g.
using transformed data fY~i ; X
12
For any possible change point k = 1; :::; T
1, de…ne matrices Z~2i (k) = Mw Z2i (k),
~ i (k) = (X
~ 0i = (X
~ i ; Z~2i (k)) and X
~ i ; Z~0i ). With new notation, (21) becomes
X
~ 0i bi + ~"0 ; i = 1; :::; N:
Y~i = X
i
(22)
Given k, slope bi can be estimated by least squares,
~bi (k) =
~ (k)
i
~i (k)
~ i (k)0 Y~i ; i = 1; :::; N:
~ i (k)] 1 X
~ i (k)0 X
= [X
(23)
The resulting sum of squared residuals is
] i (k) = [Y~i
SSR
= [Y~i
~ i (k)~bi (k)]0 [Y~i
X
~ i ~ i (k)
X
~ i (k)~bi (k)]
X
Z~2i (k)~i (k)]0 [Y~i
~ i ~ i (k)
X
Z~2i (k)~i (k)]; i = 1; :::; N;
and the estimator of k0 is de…ned similarly as
P
k~ = arg min
i
1 k T 1
where
i
]
(24)
i SSRi (k);
are weights, as in (11).
Assumption 12 For i = 1; :::; N , the matrices
1
Xi0 Mw Xi
T
and
1
Xi0 Mf Xi
T
are nonsingular,
and their inverses have …nite second-order moments.
This assumption of identifying bi and b is adopted from Pesaran (2006).
~ i , i = 1; :::; N . To identify k0 , we need a modi…ed
Let x~0it be the tth element of matrix X
version of Assumptions 4, 5, 6:
P
P
Assumption 13 For i = 1; :::; N , the matrices (1=j) jt=1 x~it x~0it , (1=j) Tt=T j+1 x~it x~0it ,
P0
P 0 +j
(1=j) kt=k
x~it x~0it and (1=j) kt=k
x~it x~0it are stochastically bounded and have minimum
0 j+1
0 +1
eigenvalues bounded away from zero in probability for all large j. In addition, for each i,
P
(1=T ) Tt=1 x~it x~0it converges in probability to a nonrandom and positive de…nite matrix as
T ! 1.
P Pk0
Assumption 14 For any positive …nite integer s, the matrices N1 N
~it x~0it and
i=1
t=k0 s+1 x
PN Pk0 +s
1
~it x~0it , i = 1; :::; N , are stochastically bounded, with minimum eigenvalues
i=1
t=k0 +1 x
N
P
bounded away from zero in probability for large N . In addition, for each t, (1=N ) N
~it x~0it
i=1 x
is stochastically bounded as N ! 1.
13
Assumption 15 f i ; i = 1; :::; N g are drawn independently of the process of f~
xit ; i = 1; :::; N g.
Alternatively, under a random coe¢ cient model, we have a slightly di¤erent version of
Assumption 7.
Assumption 16 For i = 1; :::; N ,
bi = b + vb;i ; vb;i
IID(0;
b );
v ;i
0
and b =
for i = 1; 2; :::; N , where kbk < 1,
v ;i
0
k b k < 1, and the random deviations vb;i are independent of j ; j ; "jt ; and vjt for all i; j
where b = ( 0 ; 0 )0 , vb;i =
and t.
Under Assumption 16, bi is independent of
E(C1i ) = ( +
j,
p
implying that as N ! 1, C1 =
PN
i=1
i C1i
; ) and C2 ! E(C2i ) = ( + ( + R ); ). In this case, rank condition
Assumption 11 requires non-zero means for
and
in Assumption 10 when N is large. Sim-
ilarly in Model 1, When f i ; i = 1; :::; N g are considered as random, as part of Assumption
16, Assumption 15 becomes redundant.
After the transformation (20), it can be shown that the change point estimator k~ is still
consistent in a linear model with a multifactor error structure (5), i.e., k~ k0 = op (1).
Theorem 2 Under Assumptions 1-3, 8-15 (or 16), lim(N;T )!1 P (k~ = k0 ) = 1.
Theorem 2 can be proved similarly to Theorem 1, see the Appendix.
~ the CCE estimator of the slope coe¢ cients can be
Given the change point estimator k,
written as
~bi = ~bi (k)
~ = [X
~ 0X
~ 1X
~ 0 Y~i ; i = 1; :::; N:
~ i (k)
~ i (k)]
~ i (k)
~ the asymptotics of ~bi can be established.
With the consistency of k,
Proposition 1 Under Assumptions 1-3, 8-15, and
p
T (~bi
d
bi ) ! N 0;
p
1
~
X;i
T =N ! 0 as (N; T ) ! 1; for each i,
~ ";i
X~
1
~
X;i
where
~
X;i
1 ~0 ~
X0i and
= plimT !1 X
T 0i
~ ";i
X~
1 ~0
= plimT !1 X
T 0i
14
~
";i X0i ,
i = 1; :::; N:
p
!
An additional condition
that Mw F
p
T =N ! 0 as (N; T ) ! 1 is required here, due to the fact
is included in ~"0i = Mw F
+ ~"i , the error term of transformed model (21)
p
using cross-sectional averages. This yields an extra term in T (~bi bi ) whose order is
p
p
p
Op ( T =N )+Op (1= N ) which is asymptotically ignorable when T =N ! 0 as (N; T ) ! 1.
i
i
See the Supplementary Appendix.
As discussed above, Assumption 2 allows that T can grow faster than N , i.e., T = O(N )
p
with
1. Here, the relative speed of N and T , T =N ! 0 as (N; T ) ! 1 imposes an
upper bound on
, i.e., < 2. Therefore, in the case of T = O(N ) with 1
< 2, both
p
Assumption 2 and T =N ! 0 as (N; T ) ! 1 required by Proposition 1 are satis…ed.
As discussed by Pesaran (2006), a consistent Newey-West type estimator of
~ ";i
X~
can be
obtained using the transformed data,
~~
X~
";i
!
P
= ~ i0 + (1
!
1 P
j
^ it (k)
^ 0;
)( ~ ij + ~ 0ij ); ~ ij =
e~it e~i;t j Xit (k)X
!+1
T t=j+1
j=1
~ ~bi and X
~ is the tth
~ i (k)
~ it (k)
where ! is the window size. e~it is the tth element of e~i = Y~i X
~ 0X
~ Thus, a consistent
~ Since ~ can be consistently estimated by 1 X
~ (k)
~ i (k).
~ i (k).
row of X
X;i
T i
estimator of
1
~
X;i
~ ";i
X~
1
~
X;i
is given by
1
1 ~ ~ 0~ ~
Xi (k) Xi (k)
T
~~
X~
";i
1 ~ ~ 0~ ~
Xi (k) Xi (k)
T
1
:
(25)
~ has the same limiting distribution as ~bi (k0 ), parameters bi , i = 1; :::; N , in model
Since ~bi (k)
(5) can be inferred as if k0 were known.
The mean group estimator with a common break can be de…ned similarly:
N
N
X
X
~bi = 1
~ 0X
~ 1X
~ 0 Y~i :
~bM G = 1
~ i (k)
~ i (k)]
~ i (k)
[X
N i=1
N i=1
Proposition 2 Under the assumptions 1-3, 8-14, 16,
p
As in Pesaran (2006),
b
N (~bM G
d
b) ! N (0;
b) :
can be consistently estimated by
1
N
1
N
X
(~bi
~bM G )(~bi
i=1
15
~bM G )0 :
(26)
For detailed proofs of Propositions 1 and 2, see the Supplementary Appendix. Unlike
Pesaran (2006), an additional step is needed, that of estimating k0 . As shown in the proposi~ the convergence rate of k~ is not required for deriving
tions above, with the consistency of k,
the asymptotic distributions of ~bi , for i = 1; :::; N; and ~bM G .
5
Multiple Common Break Points
(1)
(Bk )
When multiple common break points, k0 ; :::; k0
regimes for each individual:
8 0
xit
>
>
>
< x0
it
yit =
>
>
>
: 0
xit
+ eit ;
0
i + zit 1i + eit ;
..
.
i
i
+ zit0
Bk ;i
, occur in the slopes, there are Bk + 1
(1)
t = 1; :::; k0 ;
(1)
(2)
t = k0 + 1; :::; k0 ;
..
.
(Bk )
+ eit ; t = k0
(27)
+ 1; :::; T;
for i = 1; :::; N .
Estimation of multiple break points has been discussed by Bai (1997b) and Chong (1995)
in a mean-shift model, Bai and Perron (1998) in linear regression models and Bai (2010) in a
panel mean-shift model. To deal with this issue in model (27), we can follow the sequential
or one at-a-time approach discussed by Bai (1997b, 2010). The number of common breaks,
Bk , is assumed known.14 The idea of the sequential approach is to estimate break points one
(1)
(2)
(3)
by one. For example, if Bk = 3, the estimation of k0 ; k0 and k0 can be completed in 3
steps. In the …rst step, one break point is assumed as in Model 1 (or Model 2) above, and can
be estimated by (11) (or (24)), denoted by k^(1) (or k~(1) ). In the second step, in each of the
two sub-panels split by k^(1) (or k~(1) ), the same procedure (11) (or (24)) is applied. Thus, two
single break estimators are obtained in these two sub-panels. k^(2) (or k~(2) ) is de…ned as the
one associated with a larger reduction in the sum of squared residuals. Similarly, k^(1) and k^(2)
(or k~(1) and k~(2) ) yield 3 sub-panels. In the third step, in each of these 3 sub-panels, one break
point can be estimated as in Section 3 (or 4). Among these 3 break estimators, we choose
the one associated with the largest reduction of sum of squared residuals, denoted as k^(3) (or
k~(3) ). As suggested by Bai (2010), it can be shown that after rearranging (k^(1) ; k^(2) ; k^(3) ) (or
14
In a time series regression model, a sup FT (l+1jl) test is proposed by Bai and Perron (1998) to determine
the number of structure breaks. Bai and Perron (2003) report the simulation results of this test and compare
it with other tests based on information criteria. A panel version of the sup FT (l + 1jl) test can be applied
to determine the number of common breaks in our setup.
16
(k~(1) ; k~(2) ; k~(3) )) in temporal order, (k^(1) ; k^(2) ; k^(3) ) (or (k~(1) ; k~(2) ; k~(3) ) in Model 2) is consistent
(1)
(2)
(3)
for (k0 ; k0 ; k0 ) as long as the assumptions listed in Section 3 (or 4) hold in each of the
sub-panels.
(1)
(Bk )
Once the consistent estimators of (k0 ; :::; k0
) are obtained, the parameters
i ; 1i ; :::; Bk ;i ,
i = 1; :::; N , can be estimated by least squares as in (10) (or (23)). Thus, their mean-group
estimators can be obtained similarly.
6
Monte Carlo Simulations
This section employs Monte Carlo simulations to examine the consistency of the estimated
break points k^ and k~ summarized in Theorems 1 and 2. Since the CCE estimators in Model
2 have the same asymptotic distributions as if the true common breaks were known, their
asymptotic properties are not examined here. Two di¤erent designs are used for Models 1
and 2, respectively. In Model 1, there are no common correlated e¤ects in the errors and
regressors, so least squares can be run for each individual series. While in Model 2, the
regressors and errors are correlated due to common correlated e¤ects ft . A transformation,
using cross-sectional averages of the dependent variable and regressors proposed by Pesaran
(2006), is needed to remove such e¤ects asymptotically.
In the following experiments, the focus is on the histograms of k^ and k~ in setups with
di¤erent combinations of (N; T ).
6.1
Model 1: No common correlated e¤ects
The data generating process of Model 1 is modi…ed from that in Pesaran (2004, p.24):
Here we set
i
yit =
i
+
eit =
i ft
i (k0 )yi;t 1
+ eit ; i = 1; :::; N ; t = 1; :::; T ;
+ "it :
= 0, so there is no cross-sectional dependence in the errors. Instead, in this
dynamic heterogeneous panel model, there is a common break k0 = 0:5T in the slopes
for i = 1; :::; N , i.e.,
i (k0 )
=
1i ;
2i
t = 1; :::; k0 ;
= 1i + i ; t = k0 + 1; :::; T;
17
i,
where
i
is the jump in the slope for each series. We assume
iidU (0; 0:2). We set
i
=
i (1
In addition, we assume yi;0
1i );
i
= "0i +
iidN (0; 1) and "it
i
where "0i
iidN (0;
In (11), for any possible change point k = 1; :::; T
1i
iidU (0; 0:8) and
iidN (0; 1) and
2
i );
with
i
i
iidN (1; 2).
2
2 =2.
2
i
1, the estimated change point k^ is
the one that minimizes the sum of N individual sum of squared residuals. 1000 replications
are performed to obtain the histogram of k^ for each setup.
Figure 1 reports the histograms of k^ for T = 20 and N = 1; 10; 50; 200. It shows that the
distribution of k^ shrinks with N . The frequency of choosing the true value k0 increases from
8% to 58% when N increases from 1 to 200. In case T = 50, as Figure 2 shows, the frequency
of choosing the true value k0 improves to almost 90% for N = 200. This …nding supports
Theorem 1, con…rming that multiple individual series provide additional information on k0 ,
and that k^ converges to k0 as the number of series goes to in…nity.
To consider the case where there is no structural break in slopes in some series, we set
= 0 in [N=4] series, implying that N increases with N at a rate of O(N 3=4 ). Figure 3
reports the histograms of k^ for this case with T = 50. Similar to Figure 2, the pattern that
k^ converges to k0 as N increases remains. However, the frequency of choosing the true value
i
k0 is signi…cantly smaller than that in Figure 2. For example, for N = 50, the frequency of
choosing the true value k0 drops from 44% in Figure 2 to 34% in Figure 3. This suggests
that for the accuracy of the estimated change point, allowing for no break in some series is
equivalent to reducing the number of series or the magnitude of break
6.2
N.
Model 2: Common correlated e¤ects
The data generating process for Model 2 is as follows:
where
"it
i
yit =
i
eit =
1i ft
iidN (1; 1) and
iidN (0;
2
i)
and
2
i
i (k0 )
where
1i
=1+
i;
i
+
i (k0 )xi;t
+ eit ; i = 1; :::; N ; t = 1; :::; T
+ "it ;
iidN (1; 0:2). The idiosyncratic errors are generated as
1i
iidU (0:5; 1:5). There is a common break in the individual slopes:
=
1i ;
2i
=
t = 1; :::; k0 ;
k = 0:5T;
+
;
t
= k0 + 1; :::; T; 0
i
1i
iidN (0; 0:04) and
iidN (0; 0:04).
i
18
Unlike Model 1, the error eit and the regressor xit contain the common correlated e¤ect
ft :
xit = ai +
where ai
iidN (0:5; 0:5),
2i
2i ft
+ vit ;
iidN (0:5; 0:5) and vit
iidN (0; 1
2
vi );
with
vi
= 0:5.
The factor ft is generated by the stationary process:
ft =
f
f ft 1
= 0:5; vf t
+ vf t ; t =
49; :::; 0; 1; :::; T ;
iidN (0; 1
2
f ); f 50
= 0:
The correlation between xit and eit renders OLS inconsistent in the individual regressions.
Thus, transformation (20) using cross-sectional averages of yit and xit is needed to remove
ft before conducting least squares estimation of k0 .
The setup above is a simpli…ed version of the design in Pesaran (2006). First, as in model
(4), the observed factors are omitted for simplicity. Second, the number of regressors and
unobservable factors are reduced to 1, respectively. Third, the correlation structures in vit
and "it are removed. The only new feature of this model is that there is a common break at
k0 , speci…ed as 0:5T .
The …rst row of Figure 4 presents the histograms of the estimated change point k~ for
T = 20. It replicates the pattern in Figure 1, showing that after the transformation, the
frequency of choosing the true value k0 increases signi…cantly with N . Figure 4 also reports,
in the second row, the histograms of the estimated change point k^ without conducting
transformation (20). It indicates that in the presence of common correlated e¤ects, crosssectional information using multiple series fails to improve the accuracy of the estimated
change point.
Figures 5 and 6 report the histograms of k~ and k^ for T = 50 and 200, respectively.
The same pattern emerges, suggesting that the distribution of k~ shrinks to k0 as N ! 1.
^ the estimator without conducting transformation
Di¤erent from Figure 4, the frequency of k,
(20), choosing the true break date increases with N in Figure 6 when T is large, although
not at a rate as high as that of k~ using the transformed data. Whether jk^ k0 j shrinks to
0 or not as (N; T ) ! 1 depends upon the correlation between xit and eit . In Figure 7, we
increase this correlation by changing the distribution of
1i
from N (1; 0:2) to N (2; 0:2). In
this case, the cross-sectional information using multiple series fails to improve the accuracy
19
Figure 1: Histograms of 𝑘� in Model 1: T =20
The Frequency of Estimated Change Point
0.12
0.1
0.1
0.08
0.08
Frequency
Frequency
0.12
0.06
0.06
0.04
0.04
0.02
0.02
0
0
2
4
6
8
10
12
14
16
18
0
20
0
2
4
6
8
10
k0=10
k0=10
N =1
N =10
12
14
16
18
20
18
20
0.25
0.7
0.6
0.2
0.15
Frequency
Frequency
0.5
0.1
0.4
0.3
0.2
0.05
0.1
0
0
2
4
6
8
10
12
14
16
18
0
20
0
2
k0=10
6
8
10
k0=10
N =50
N =200
Note: The DGP is similar to Pesaran (2004, p.24):
𝑦𝑖𝑖 = 𝛼𝑖 + 𝛽𝑖 (𝑘0 )𝑦𝑖,𝑡−1 + 𝑒𝑖𝑖 , 𝑖 = 1, … , 𝑁; 𝑡 = 1, … , 𝑇,
where 𝛽𝑖 (𝑘0 ) = �
4
𝛽1𝑖 ,
𝑡 = 1, … , 𝑘0 ,
𝛽2𝑖 = 𝛽1𝑖 + 𝛿𝑖 , 𝑡 = 𝑘0 + 1, … , 𝑇.
𝑘0 = 0.5𝑇 = 10, 𝑒𝑖𝑖 = 𝛾𝑖 𝑓𝑡 + 𝜀𝑖𝑖 , 𝛾𝑖 = 0 and 𝛼𝑖 = 𝜇𝑖 (1 − 𝛽1𝑖 ), 𝜇𝑖 = 𝜀0𝑖 + 𝜂𝑖 .
Values used: 𝛽1𝑖 ~𝑖𝑖𝑖𝑖(0, 0.8), 𝛿𝑖 ~𝑖𝑖𝑖𝑖(0, 0.2), 𝜀𝑖𝑖 ~𝑖𝑖𝑖𝑖(0, 𝜎𝑖2 ), 𝜎𝑖2 ~𝜒22 /2,
𝜀0𝑖 ~𝑖𝑖𝑖𝑖(0, 1) and 𝜂𝑖 ~𝑖𝑖𝑖𝑖(1, 2), 𝑦𝑖,0 ~𝑖𝑖𝑖𝑖(0, 1), 𝑓𝑡 ~𝑖𝑖𝑖𝑖(0, 1).
These variables are mutually independent.
The replication number is 1000.
20
12
14
16
Figure 2: Histograms of 𝑘� in Model 1: T =50
0.07
0.18
0.16
0.06
0.14
0.05
Frequency
Frequency
0.12
0.04
0.03
0.1
0.08
0.06
0.02
0.04
0.01
0
0.02
0
5
10
15
20
25
30
35
40
45
0
50
0
5
0.35
0.7
0.3
0.6
Frequency
Frequency
0.8
0.25
0.2
0.1
0.2
0.05
0.1
25
30
35
40
40
45
50
45
50
0.4
0.3
20
35
0.5
0.15
15
30
0.4
10
25
N =10
0.9
5
20
N =1
0
15
k0=25
0.45
0
10
k0=25
45
0
50
k0=25
0
5
10
15
20
25
k0=25
N =50
N =200
Note: The DGP is the same as in Figure 1. 𝑘0 = 0.5𝑇 = 25.
21
30
35
40
Figure 3: Histograms of 𝑘� in Model 1 with No Break in Some Series: T =50
0.07
0.12
0.06
0.1
0.05
Frequency
Frequency
0.08
0.04
0.03
0.06
0.04
0.02
0.02
0.01
0
0
5
10
15
20
25
30
35
40
45
0
50
0
N =1
N =10
40
50
0.8
0.3
0.7
0.6
Frequency
0.25
Frequency
30
20
k0=25
0.35
0.2
0.15
0.1
0.5
0.4
0.3
0.2
0.05
0
10
k0=25
0.1
0
10
30
20
40
0
50
k0=25
0
10
30
20
k0=25
N =50
N =200
Note: The DGP is the same as in Figure 1 except that there is no break in [N/4] series. 𝑘0 = 0.5𝑇 = 25.
22
40
50
Figure 4: Histograms of 𝑘� and 𝑘� in Model 2: T =20
N =10
N =50
The Frequency of Estimated Change Point: transformation
0.5
0.09
0.18
0.45
0.08
0.16
0.4
0.07
0.14
0.35
0.06
0.12
0.05
0.04
Frequency
0.2
0.1
0.08
0.2
0.15
0.06
0.02
0.04
0.1
0.01
0.02
0.05
0
2
4
6
8
10
12
14
16
18
0
20
0
2
4
6
8
k0
10
12
14
16
18
0
20
0.09
0.08
0.09
0.08
0.08
0.07
0.05
0.04
0.03
0.05
0.04
0.02
10
12
14
16
18
20
0
14
16
18
20
0.05
0.04
0.01
0.01
8
12
0.02
0.02
0.01
10
0.03
0.03
6
8
0.06
0.06
Frequency
Frequency
0.06
4
6
0.07
0.07
2
4
The Frequency of Estimated Change Point: no transformation
0.1
0
2
k0
0.09
0
0
k0
Frequency
0.3
0.25
0.03
0
𝑘�
0.1
Frequency
𝑘�
Frequency
N =200
0
2
4
6
8
k0
10
k0
12
14
16
18
20
0
0
2
4
6
8
10
12
14
16
18
20
k0
Note: The DGP is constructed by simplifying the design of Pesaran (2006):
𝛽1𝑖 ,
𝑡 = 1, … , 𝑘0 ,
𝑦𝑖𝑖 = 𝛼𝑖 + 𝛽𝑖 (𝑘0 )𝑥𝑖,𝑡 + 𝑒𝑖𝑖 , 𝑖 = 1, … , 𝑁; 𝑡 = 1, … , 𝑇. 𝛼𝑖 ~𝑖𝑖𝑖𝑖(1, 1), 𝛽𝑖 (𝑘0 ) = �
𝑘 = 0.5𝑇 = 10, 𝛽1𝑖 ~𝑖𝑖𝑖𝑖(1, 0.04), 𝛿𝑖 ~𝑖𝑖𝑖𝑖(0, 0.04).
𝛽2𝑖 = 𝛽1𝑖 + 𝛿𝑖 , 𝑡 = 𝑘0 + 1, … , 𝑇. 0
𝑒𝑖𝑖 = 𝛾𝑖1 𝑓𝑡 + 𝜀𝑖𝑖 , 𝑥𝑖,𝑡 = 𝑎𝑖 + 𝛾𝑖2 𝑓𝑡 + 𝑣𝑖𝑖 ; 𝑓𝑡 = 𝜌𝑓 𝑓𝑡−1 + 𝑣𝑓𝑓 , 𝑡 = −49, … ,0,1, … 𝑇, 𝑣𝑓𝑓 ~𝑖𝑖𝑖𝑖(0, 1 − 𝜌𝑓2 ), 𝜌𝑓 = 0.5, 𝑓−50 = 0. 𝜀𝑖𝑖 ~𝑖𝑖𝑖𝑖(0, 𝜎𝑖2 ), 𝜎𝑖2 ~𝑖𝑖𝑖𝑖(0.5, 1.5),
2
𝛾𝑖1 ~𝑖𝑖𝑖𝑖(1, 0.2), 𝛾𝑖2 ~𝑖𝑖𝑖𝑖(0.5, 0.5), 𝑎𝑖 ~𝑖𝑖𝑖𝑖(0.5, 0.5), 𝑣𝑖𝑖 ~𝑖𝑖𝑖𝑖(0, 1 − 𝜌𝑣𝑣
), 𝜌𝑣𝑣 = 0.5. These variables are mutually independent. The replication number is 1000.
23
N =10
N =50
N =200
0.09
0.08
0.35
0.8
0.3
0.7
0.07
0.6
0.25
0.05
0.04
Frequency
Frequency
𝑘�
Frequency
0.06
0.2
0.15
0.5
0.4
0.3
0.03
0.1
0.2
0.02
0.05
0.01
0
0
5
10
15
20
25
30
35
40
45
0
50
0.1
0
5
10
15
20
k0
25
30
35
40
45
0
50
0
5
10
15
20
k0
0.06
25
30
35
40
45
50
k0
0.08
0.09
0.07
0.08
0.05
0.07
0.06
0.06
0.03
0.05
Frequency
Frequency
𝑘�
Frequency
0.04
0.04
0.03
0.02
0.05
0.04
0.03
0.02
0.02
0.01
0.01
0
0
5
10
15
20
25
k
30
35
40
45
50
0
0.01
0
5
10
15
20
25
k
0
0
24
30
35
40
45
50
0
0
5
10
15
20
25
k0
30
35
40
45
50
N =10
N =50
N =200
0.18
0.7
1
0.9
0.16
0.6
0.8
0.14
0.5
0.7
0.1
0.08
Frequency
Frequency
𝑘�
Frequency
0.12
0.4
0.3
0.06
0.2
0.1
0
50
100
150
0
200
0
50
100
k0
150
0
200
50
100
150
200
k0
0.45
0.25
0.08
0.4
0.2
0.07
0.35
0.05
0.04
Frequency
0.3
Frequency
0.06
Frequency
0
k0
0.09
0.15
0.1
0.03
0.25
0.2
0.15
0.02
0.1
0.05
0.01
0
0.4
0.1
0.02
𝑘�
0.5
0.3
0.2
0.04
0
0.6
0.05
0
50
100
150
200
0
0
50
100
k0
k0
25
150
200
0
0
50
100
k0
150
200
Figure 7: Histograms of 𝑘� and 𝑘� in Model 2 (with increased the correlation between 𝑥𝑖𝑖 and 𝑒𝑖𝑖 ): T =200
N =10
N =50
N =200
0.18
0.7
1
0.9
0.16
0.6
0.8
0.14
0.5
0.7
0.1
0.08
Frequency
Frequency
𝑘�
Frequency
0.12
0.4
0.3
0.06
0.5
0.4
0.3
0.2
0.04
0.2
0.1
0.02
0
0.6
0.1
0
50
100
150
0
200
0
50
100
k0
150
0
200
0
50
k0
200
0.035
0.03
0.03
0.025
150
k0
0.03
100
0.025
0.025
0.015
Frequency
0.02
Frequency
𝑘�
Frequency
0.02
0.02
0.015
0.01
0.015
0.01
0.01
0.005
0
0.005
0.005
0
50
100
150
200
0
0
50
100
k0
k0
150
200
0
0
50
100
150
k0
Note: The DGP is the same as in Figure 4, except that the correlation between 𝑥𝑖𝑖 and 𝑒𝑖𝑖 increases by changing the distribution of 𝛾𝑖1 from 𝑖𝑖𝑖𝑖(1, 0.2) to 𝑖𝑖𝑖𝑖(2, 0.2).
𝑘0 = 0.5𝑇 = 100.
26
200
^ This is consistent with the …ndings of Kim (2011).15
of the estimated change point k.
7
Conclusion
The recent literature on panel data models with large time dimension assumes that the
slopes are constant over time. However, due to global policy or technological shocks, slope
parameters are likely to fall into di¤erent regimes over a longer time span. This paper
tackles both structural breaks and cross-sectional dependence by extending Pesaran’s (2006)
framework of heterogeneous panels to the situation of unknown common breaks in the slopes.
The least squares method proposed by Bai (1997a, 2010) is applied to estimate the common
change points. This method is married to the CCE estimators proposed by Pesaran (2006)
in this setup. Di¤erent from the time series change point models, in which the estimated
change point is inconsistent, our paper establishes its consistency in a general panel data
setup. The properties of the CCE estimators of the slopes at estimated points are also
examined. We …nd that the CCE estimators have the same asymptotic distributions as if
the true change points were known.16
In this paper, we assume that the rank condition is satis…ed. In case of de…cient rank,
Westerlund and Urbain (2013) show that the CCE estimators could be inconsistent when
factor loadings in the error term and in the explanatory variables are correlated. In this case,
the iterative principal component approach, proposed by Bai (2009) in homogeneous panels
and extended by Song (2012) to dynamic heterogeneous panels, can be an alternative, which
allows for correlated and zero-mean factor loadings. However, the issue of structural breaks
in this setup is beyond the scope of this paper.
15
Experiments with di¤erent values of parameters and distributions are conducted and similar histograms
are obtained. Results are not included here to save space.
16
Since the convergence rate of the change point estimator is not required to derive the asymptotic distributions of the CCE estimators of slopes and their cross-sectional mean, we leave the derivations of convergence
rate and asymptotic distribution of the change point estimator in heterogeneous panels for future research.
27
Appendix: Mathematical Proofs
Since the panel data model (6) considered here includes the time series model in Bai
(1997a) as a special case of N = 1, it can be shown similarly that k^
the proofs that follow, we assume k^
k0 = Op (1). In
k0 is stochastically bounded. With more information
along the cross-sectional dimension under the common break assumption, we further show
that k^
p
k0 ! 0 as (N; T ) ! 1.
For i = 1; :::; N , let SSRi be the sum of squared residuals of regressing Yi on Xi in case
there is no break, i.e., Z2i (k) = 0T
SSRi
SSRi (k) = [Yi
q.
Using the identity
Xi ^ i (k)
Z2i (k)î (k)]0 [Yi
Xi ^ i (k)]0 [Yi
[Yi
Xi ^ i (k)
Z2i (k)î (k)]
Xi ^ i (k)]
= î (k)0 [Z2i (k)0 Mi Z2i (k)]î (k)
Xi (Xi0 Xi ) 1 Xi0 ,
with Mi = I
k^ = arg min
1 k T 1
N
X
i SSRi (k)
= arg max
1 k T 1
i=1
N
X
i SVi (k)
i=1
= arg max
1 k T 1
N
X
i [SVi (k)
SVi (k0 )];
i=1
0
Mi Z0i ] î (k0 ) is
where SVi (k) = î (k)0 [Z2i (k)0 Mi Z2i (k)]î (k). Note that SVi (k0 ) = î (k0 )0 [Z0i
not a function of k. For simplicity, we assume i = 1=N , i = 1; :::; N .17
P
SVi (k0 )] can be decomposed into a deterministic
To prove Theorem 1, N
i=1 [SVi (k)
part and a stochastic one. Partitioned regression gives
î (k) = [Z2i (k)0 Mi Z2i (k)]
Substituting Yi = Xi
i
+ Z0i
i
i
0
+ (Z0i
Mi Z0i )
1
Z2i (k)0 Mi Yi , i = 1; :::; N:
+ "i into the equation above, we obtain
î (k) = [Z2i (k)0 Mi Z2i (k)]
and î (k0 ) =
1
1
Z2i (k)0 Mi Z0i
i
+ [Z2i (k)0 Mi Z2i (k)]
1
Z2i (k)0 Mi "i
Z0i 0 Mi "i :
17
Since the weights i , i = 1; :::; N , are used to balance di¤erent variances across series, we can always
employ i = 1=N , i = 1; :::; N , by using di¤erent notations for Xi and 2i .
28
To simplify notation, k is suppressed in î (k) and Z2i (k) when no confusion arises. Since
0
0
SVi (k) = î (Z2i
Mi Z2i )î =
0
0
0
1
0
i (Z0i Mi Z2i )(Z2i Mi Z2i ) (Z2i Mi Z0i ) i
0
0
0
0
0
Mi Z2i )(Z2i
Mi Z2i ) 1 Z2i
Mi "i + "0i Mi Z2i (Z2i
Mi Z2i ) 1 Z2i
M i "i ;
+ 2 0i (Z0i
it follows that
SVi (k)
0
0
i [(Z0i Mi Z0i )
SVi (k0 ) =
0
0
0
Mi Z0i )]
(Z0i
Mi Z2i )(Z2i
Mi Z2i ) 1 (Z2i
0
0
0
M i "i
Mi Z2i ) 1 Z2i
Mi Z2i )(Z2i
+2 0i (Z0i
0
0
+"0i Mi Z2i (Z2i
Mi Z2i ) 1 Z2i
Mi "i
(28)
i
0
M i "i
2 0i Z0i
(29)
0
0
"0i Mi Z0i (Z0i
Mi Z0i ) 1 Z0i
Mi "i :(30)
The deterministic part is denoted by
J1i (k) =
0
0
i [(Z0i Mi Z0i )
0
0
0
(Z0i
Mi Z2i )(Z2i
Mi Z2i ) 1 (Z2i
Mi Z0i )] i ;
(31)
and the stochastic part is denoted by
0
0
0
J2i (k) = 2 0i (Z0i
Mi Z2i )(Z2i
Mi Z2i ) 1 Z2i
M i "i
0
0
+"0i Mi Z2i (Z2i
Mi Z2i ) 1 Z2i
M i "i
Thus SVi (k)
SVi (k0 ) =
k^ = arg max
1 k T 1
De…ne
X
X
Z
i
i
i
N
X
0
2 0i Z0i
M i "i
0
0
"0i Mi Z0i (Z0i
Mi Z0i ) 1 Z0i
M i "i :
J1i (k) + J2i (k) and
[SVi (k)
SVi (k0 )] = arg max
1 k T 1
i=1
"
N
X
J1i (k) +
i=1
N
X
#
J2i (k) :
i=1
= X2i X0i = (0;
; 0; xi;k+1 ;
; xi;k0 ; 0;
; 0)0 ;
for k < k0 ;
0
= (X2i X0i ) = (0;
; 0; xi;k0 +1 ;
; xi;k ; 0;
; 0) ; for k k0 :
can be de…ned similarly.
For a …nite large number Ck and arbitrarily small positive number a <
K(Ck ) = fk : 1
jk
k0 j < Ck ; aT < k < (1
1 (k)
> 0 and
1
> 0.
29
de…ne the set
a)T g. Since k^ k0 is stochastically bounded,
we only consider the values of k that belong to set K(Ck ).
P
0
0
Let 1 (k) be the minimum eigenvalue of N1 N
i=1 R (X i X i )R. De…ne
Under Assumption 5,
0,
1
= mink2K(Ck )
1 (k).
Lemma 1 Under Assumptions 1-7, for all large N and T , with probability tending to 1,
inf
K(Ck )
N
X
J1i (k)
1 N:
i=1
This lemma is similar to Lemma A.2 in Bai (1997a). The proof can be found in the
Supplementary Appendix.
Lemma 2 Under Assumptions 1-7, uniformly on K(Ck ),
(i)
(ii)
PN
0 0
i Z i "i
i=1
p1
T
(iii)
p1
T
(iv)
1
T
PN
i=1
PN
i=1
PN
i=1
p
= Op (
N );
0"
Xi0 Xi
0 0
1X
pi i
i Z i Xi ( T )
T
q
= Op (
N
T
0 M Z
Z2i
Z0 M "
0
i 2i
0
) 1 2ipTi i
i (Z i Mi Z2i )(
T
"0i Mi Z i (
0 M Z
Z2i
i 2i
) 1 Z 0 i M i "i
T
);
q
= Op (
N
T
);
= Op ( N
);
T
q
0 M Z
Z2i
i 2i
N
N
0
1 0
"
M
Z
(
M
"
=
O
(
)
Z
)
+
O
(
);
p T
p
i i i
i=1 i i 0i
T
T
i
h
P "0i Mi Z0i Z2i0 Mi Z2i 1
Z0 M Z
Z0 M "
p
(vi) N
( T )
( 0i T i 0i ) 1 0ipTi i = Op ( N
):
i=1
T
T
(v)
1
T
PN
Proof of Lemma 2. (i) Under Assumption 3, for large N ,
!
N
N
X
X
0 0
0 0
V ar
Z
"
=
i
i i
i Z i ";i Z i i :
i=1
It can be shown equal to O(
i=1
N ) under
Assumptions 4 to 7, similar to the proof of Lemma 1
p
P
0 0
"
=
O
(
Z
in the Supplementary Appendix, implying N
i
p
N ) on K(Ck ).
i
i
i=1
The proofs of Lemma 2(ii)-(vi) are included in the Supplementary Appendix.
With these lemmas, we are ready to prove Theorem 1.
Proof of Theorem 1. To prove lim(N;T )!1 P (k^ = k0 ) = 1, it is equivalent to showing
> 0, for both large T and N , P (jk^ k0 j
P
show that P supK(Ck ) N
SVi (k0 )] 0 < ; or
i=1 [SVi (k)
that for any given
P
sup
N
X
J2i (k)
inf
K(Ck )
K(Ck ) i=1
By Lemma 1, it su¢ ces to show P (supK(Ck )
1
N
30
PN
N
X
J1i (k)
i=1
i=1
!
J2i (k)
1) < . It is su¢ cient to
< :
1)
< :For any k 2 K(Ck ),
N
X
N
X
J2i (k)
i=1
0
0
0
M i "i
Mi Z2i ) 1 Z2i
Mi Z2i )(Z2i
2 0i (Z0i
0
M i "i
2 0i Z0i
i=1
+
N
X
0
0
"0i Mi Z2i (Z2i
Mi Z2i ) 1 Z2i
Mi "i
0
0
"0i Mi Z0i (Z0i
Mi Z0i ) 1 Z0i
Mi "i :
i=1
Consider the …rst term, Z2i = Z0i + Z
N
X
i
for k < k0 ;
0
0
0
Mi Z2i )(Z2i
Mi Z2i ) 1 Z2i
Mi "i
2 0i (Z0i
0
M i "i
2 0i Z0i
i=1
=
N
X
i=1
N
X
2
0
0
Mi Z2i ) 1 Z2i
M i "i
2 0i (Z 0 i Mi Z2i )(Z2i
2 0i Z 0 i Mi "i
0 0
i Z i "i
i=1
N
2 X
+p
T i=1
0
i
N
2 X 0 0
Xi0 Xi
+p
)
i Z i Xi (
T
T i=1
(Z 0 i Mi Z2i )(
0
Mi Z2i
Z2i
)
T
0
1 Xi "i
p
T
0
1 Z2i Mi "i
p
T
:
By (i), (ii) and (iii) of Lemma 2, the …rst term
N
X
0
0
0
2 0i (Z0i
Mi Z2i )(Z2i
Mi Z2i ) 1 Z2i
Mi "i
0
2 0i Z0i
M i "i
i=1
Now consider the second term
N
X
0
0
"0i Mi Z2i (Z2i
Mi Z2i ) 1 Z2i
M i "i
p
= Op (
N ).
(32)
0
0
"0i Mi Z0i (Z0i
Mi Z0i ) 1 Z0i
M i "i
i=1
N
N
0
0
Mi Z2i 1 0
Mi Z2i 1 0
1 X "0i Mi Z0i Z2i
Z2i
1 X 0
p
) Z i Mi "i + 2 p
(
) Z i M i "i
"i Mi Z i (
T i=1
T
T
T i=1
T
N
0
X
Mi Z2i
"0i Mi Z0i Z2i
p
+
(
)
T
T
i=1
1
(
0
Mi Z0i
Z0i
)
T
1
0
M"
Z0i
p i i :
T
Similarly, by (iv), (v) and (vi) of Lemma 2, the second term
N
X
0
0
Mi Z2i ) 1 Z2i
M i "i
"0i Mi Z2i (Z2i
0
0
Mi Z0i ) 1 Z0i
M i "i
"0i Mi Z0i (Z0i
i=1
r
N
N
= Op ( ) + Op (
):
T
T
(33)
Combining (32) and (33), we obtain
N
1 X
N
i=1
J2i (k) =
1
N
Under Assumption 2,
p
[Op (
1
N
r
r
N
N
1
N
N
1
)+Op (
)] = Op ( p )+ [Op ( )+Op (
)]:
N )+Op (
T
T
T
T
N
N
PN
i=1
J2i (k) vanishes for any k 2 K(Ck ), so does its maximum.
31
Compared with (8) of Model 1, (21) of Model 2 has the same form using transformed
~ i ; i = 1; :::; N g, except for the additional term Mw F i . The focus of the proof of
data fY~i ; X
Theorem 2 is on showing that Mw F
i
can be ignored asymptotically as (N; T ) ! 1.
~ i alone. Us] i be the sum of squared residuals of regressing Y~i on X
For i = 1; :::; N , let SSR
~ i Z~2i (k)]~i (k) with M
~i = I X
~ i (X
~ i0 X
~i) 1X
~ i0 ,
] i SSR
] i (k) = ~i (k)0 [Z~2i (k)0 M
ing the identity SSR
we obtain
k~ = arg min
1 k T 1
N
X
]
i SSRi (k) = arg max
1 k T 1
i=1
N
X
i=1
g
i SV i (k) = arg max
1 k T 1
gi (k) = ~i (k)0 [Z~2i (k)0 M
~ i Z~2i (k)]~i (k). Assume
where SV
i
N
X
g
i [SV i (k)
i=1
gi (k0 )]
SV
= 1=N , i = 1; :::; N for simplicity.
The rest proof proceeds in the same way as that of Theorem 1 using the new notations with
"~".
Partitioned regression gives
h
i
~i (k) = Z~2i (k)0 M
~ i Z~2i (k)
1
~ i Y~i :
Z~2i (k)0 M
~ i i + Z~0i i + ~"0 into the equation above, we obtain
Substituting Y~i = X
i
h
i 1
h
i 1
~i (k) = Z~2i (k)0 M
~ i Z~2i (k)
~ i Z~0i i + Z~2i (k)0 M
~ i Z~2i (k)
~ i~"0
Z~2i (k)0 M
Z~2i (k)0 M
i
and ~i (k0 ) =
~"0i = Mw F
i
i
0 ~ ~
Mi Z0i
+ Z~0i
1
~ i~"0 : Note that there is an additional term Mw F
Z~0i 0 M
i
i
in
+ ~"i .
As in the proof of Theorem 1, we suppress k in ~i (k) and Z~2i (k) for simplicity. Since
0 ~ ~ ~
gi (k) = ~0i (Z~2i
SV
Mi Z2i ) i
=
0 ~0 ~ ~
1 ~ 0 ~ ~
~0 ~ ~
i (Z0i Mi Z2i )(Z2i Mi Z2i ) (Z2i Mi Z0i ) i
0 ~ 0
1 ~0 ~ 0
0 ~ ~
0 ~ ~
~ ~ ~0 ~ ~
"i ;
Mi Z2i ) 1 Z~2i
Mi~"i + ~"00
+2 0i (Z~0i
Mi Z2i )(Z~2i
i Mi Z2i (Z2i Mi Z2i ) Z2i Mi ~
gi (k0 ) =
and SV
gi (k)
SV
0 ~ 0 ~ ~
i (Z0i Mi Z0i ) i
gi (k0 ) =
SV
0 ~ 0
1 ~0 ~ 0
~ ~ ~0 ~ ~
+ 2 0i Z~0i
Mi~"i + ~"00
"i , it follows that
0 ~0 ~ ~
1 ~ 0 ~ ~
~0 ~ ~
i [(Z0i Mi Z2i )(Z2i Mi Z2i ) (Z2i Mi Z0i )
0 ~ ~
0 ~ ~
0 ~ 0
+2 0i (Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 Z~2i
Mi~"i
1 ~0 ~ 0
~ ~ ~0 ~ ~
+~"00
"i
32
~ i Z~0i )]
(Z~0i 0 M
i
0 ~ 0
2 0i Z~0i
Mi~"i
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i :
De…ne
J~1i (k) =
0 ~ 0 ~ ~
i [(Z0i Mi Z0i )
0 ~ ~
0 ~ ~
~ i Z~0i )] i :
(Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 (Z~2i 0 M
In addition, de…ne
0 ~ ~
0 ~ ~
0 ~ 0
J~2i (k) = 2 0i (Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 Z~2i
Mi~"i
1 ~0 ~ 0
~ ~ ~0 ~ ~
+~"00
"i
gi (k)
Thus, SV
gi (k0 ) =
SV
k~ = arg max
1 k T 1
For each i, i = 1;
for k
N
X
i=1
0 ~ 0
2 0i Z~0i
Mi~"i
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i :
J~1i (k) + J~2i (k) and
gi (k)
[SV
gi (k0 )] = arg max
SV
1 k T 1
; N , de…ne Z~
i
= Z~2i (k)
k0 : Since Z0i = X0i R and Z2i = X2i R, Z~
"
N
X
i=1
J~1i (k) +
N
X
J~2i (k) :
i=1
Z~2i (k0 ) for k < k0 and Z~2i (k0 )
i
= Mw Z
As in the proof of Theorem 1, we assume that k~
i
#
Z~2i (k)
~ i R.
= Mw X i R = X
k0 is stochastically bounded, and only
consider the values of k belonging to the set K(Ck ) = fk : 1 jk k0 j < Ck ; aT < k < (1 a)T g.
P
0 ~0 ~
~
~
Let ~ 1 (k) be the minimum eigenvalue of N1 N
i=1 R (X i X i )R. De…ne 1 = mink2K(Ck ) 1 (k).
Under Assumption 14, ~ 1 (k) > 0 and ~ 1 > 0.
Lemma 3 Under Assumptions 1, 2, 8-15 (or 16), for all large N and T , with probability
tending to 1,
inf
K(Ck )
N
X
J~1i (k)
~1
N:
i=1
Lemma 3 can be shown in the same way as Lemma 1 using the transformed data or the
"~" notation. See the Supplementary Appendix.
Di¤erent from Model 1, there is an extra term Mw F
i
in the error ~"0i , thus in J~2i (k).
To examine the e¤ect of this extra term on the estimated k~ and ~bi , we introduce some new
matrix notation. Since xit =
0
i ft
+ vit in (3), we write
Xi
T p
= F
i
T mm p
33
+ Vi ;
T p
; viT )0 . Denote F0 = (0;
where Vi = (vi1;
; fT )0 and V0i = (0;
; 0; fk0 +1 ;
; 0; vi;k0 +1 ;
Thus,
X0i = (0;
= F0
; xi;T )0 = (0;
; 0; xi;k0 +1 ;
i
ut (k0 ) =
U (k0 ) =
0
i fk0 +1
+ vi;k0 +1 ;
;
0
i fT
+ vi;T )0
+ V0i :
For the error term (18), denote ut =
Thus, ut (k0 ) =
; 0;
PN
i=1
i uit (k0 )
0
0
;
;
8
>
>
<
"t +
0
0
;
0
i vit i
and
t = 1; :::; k0 ;
0
i vit R i
i=1
= ut +
i=1
vt
0
;
0
PN
>
>
:
PN
0
; t = k0 + 1; :::; T:
ut (k0 ): Denote U = (u1 ; :::; uT )0 and
PN
PN
0
0
i=1 i vi;k0 +1 R i
i=1 i vi;T R
;
;
0
0
i
0
:
Thus, stacking cross-sectional averages wt = C(k0 )0 ft + ut (k0 ), we obtain
W
T (p+1)
; wT )0
= (w1 ; :::; wk0 ; wk0 +1 ;
= (C10 f1 + u1 ; :::; C10 fk0 + uk0 ; C20 fk0 +1 + uk0 +1 ;
= F C1 + F0 (C2
Denote
F = (F; F0 ),
T 2m
C
2m (p+1)
C1 ) + U +
U (k0 ):
C1 )0 )0 and
= (C10 ; (C2
; C20 fT + uT )0
U
T (p+1)
=U +
U (k0 ). Therefore,
(34)
W = FC + U:
With this notation, we obtain a lemma, which can be proved similarly to Lemmas 1, 2 and
3 in Pesaran (2006).
Lemma 4 Under Assumptions 1, 2, 8-15, uniformly on K(Ck ),
(i) ut = Op ( p1N );
(ii)
1 0
UU
T
ut (k0 ) = Op ( p1N );
= Op ( N1 );
1 0
FU
T
= Op ( pN1 T ),
1 0
V F
T i
(iii)
1 0
V U
T i
= Op ( N1 ) + Op ( pN1 T );
1 0
"U
T i
(iv)
1
Xi0 U
T
= Op ( N1 ) + Op ( pN1 T );
1
0
X0i
U
T
= Op ( p1T );
= Op ( N1 ) + Op ( pN1 T ),
1 0
V U
T 0i
= Op ( N1 ) + Op ( pN1 T ):
34
= Op ( N1 ) + Op ( pN1 T );
; vi;T )0 .
(i)
(ii)
1 0
FF
T
= Op (1);
1
Xi0 F
T
1 0
FF
T
= Op (1);
= Op (1);
1
X (k)0 F
T i
= Op (1):
Proof of Lemma 5. (i) is obvious by Assumption 8.
(ii) Since Xi = F
i +Vi
By (i) and Lemma 4 (iv),
= (F; F0 )( 0i ; 0)0 +Vi , T1 Xi0 F can be written as ( 0i ; 0)( T1 F0 F)+ T1 Vi0 F.
1
Xi0 F
T
= Op (1). Similarly,
1
X (k)0 F
T i
= Op (1):
With Lemmas 4 and 5, we are ready to establish the property of the T
m matrix
Mw F i , which will be frequently used in the derivations below. Denote
E
(p+1) (p+1)
f (E)
(p+1) (p+1)
1 0 0
1
1
C F U + U0 FC + U0 U;
T
T
T
1
P
1 0
1 0
1
1
( 1)k+1 [( C F0 FC) E]k ( C F0 FC) :
=
T
T
k=1
=
By Lemma 5 (v), E = Op ( N1 ) + Op ( pN1 T ), thus f (E) = Op ( N1 ) + Op ( pN1 T ). In addition,
denote
D1
2m m
=
0
F
0F F
Cf (E)C
+ C[(C
C)
T
T
0F
0
1
U0 F
+ f (E)]
T
(35)
and
D2
=
[(C
0F
(p+1) m
Since C = O(1),
F0 F
T
and
F0 F
T
0
F
C)
T
1
+ f (E)](C
U0 F
F
+
):
T
T
0F
0
are Op (1), f (E) = Op ( N1 ) + Op ( pN1 T ), and
(36)
U0 F
T
= Op ( pN1 T ),
1
1
1
1
1
) + Op ( p
)]Op (1) + Op (1)[Op (1) + Op ( ) + Op ( p
)]Op ( p
)
N
N
NT
NT
NT
1
1
= Op ( ) + Op ( p
):
N
NT
D1 = Op (1)[Op (
Similarly,
D2 = [Op (1) + Op (
1
1
1
) + Op ( p
)][Op (1) + Op ( p
)] = Op (1):
N
NT
NT
Mw F
i
= FD1
35
i
+ UD2 i :
By Lemma 4 (i) where each element of U is Op ( p1N ), each element of Mw F
i
is of order
Op ( p1N ):
Proof of Lemma 6. Plugging in (34), we obtain
1 0
1 0
1 0
1
1
1 0
W W = C F0 FC + C F0 U + U0 FC + U0 U = C F0 FC + E:
T
T
T
T
T
T
By Lemma 5 (i),
1 0 0
C F FC
T
is Op (1). Since E = Op ( N1 ) + Op ( pN1 T ), it could be very small
when both N and T are large. By Horn and Johnson (1985, p.335)
1
1 0
1 0
1 0
1
1
( W 0 W ) 1 = ( C F0 FC)
( C F0 FC + E)
( C F0 FC)
T
T
T
T
1 0 0
1 0 0
1
1
1
1 1 0 0
= ( C F FC)
[I + ( C F FC) E] ( C F FC)
T
T
T
1
P
1
1
k 1 0 0
k+1 1 0 0
=
( 1) [( C F FC) E] ( C F FC) = f (E):
T
T
k=1
1
This yields
1
( W 0W )
T
1 0
= ( C F0 FC)
T
1
1
+ f (E):
It follows that
1
W ( W 0W )
T
1 0
1
1
(FC + U)[( C F0 FC) + f (E)] (FC + U)0 ]F
T
T
1
1
1
0 0
1
0
1
= [IT (FC)(C F FC) (FC)0 ]F (FC)ff (E)( FC)0 + [( C F0 FC) + f (E)] U0 gF
T
T
T
1
1 0
1 0 0
1
U[( C F FC) + f (E)]( FC + U) F;
T
T
T
Mw F = [IT
1
1 0
W ]F = [IT
T
As discussed in Pesaran (2006), MFC = IT
0
1
(FC)(C F0 FC) (FC)0 = IT
F(F0 F) 1 F under
the rank assumption. This implies that the …rst term is 0. Therefore, plugging in (35) and
(36), we obtain
Mw F
i
= FD1
i
+ UD2 i :
The following lemma collects terms involving Mw F i .
Lemma 7 Under Assumptions 1, 2, 8-15 (or 16), uniformly on K(Ck ),
36
(37)
(i)
(ii)
PN
(iii)
1
T
1
T
PN
i=1
1
T
b. T1
b.
1
T
c.
1
T
PN
PN
(vi)a.
1
T
Op ( pN1 T ):
PN
PN
p
= Op (
N );
q
q
N
N
=
O
(
)
+
O
(
);
p
p
i
N
T
q
q
~0 M
~ ~
Z
0 ~0 ~ ~
1 ~0 ~
2i i Z2i
N
(
Z
M
Z
)(
)
Z
M
M
F
=
O
(
)
+
O
(
p
p
i
i
2i i w
i i 2i
T
N
~ 0X
~
0 ~0 ~ X
1 ~0
i i
i Z i Xi ( T ) Xi Mw F
PN
i=1
i=1
i
0
0 0
~ ~ Z~2i M~ i Z~2i ) 1 Z~ 0 M
~ "i
i F Mw Mi Z0i (
i i~
T
h
N
T
);
= Op ( T1 );
i
= Op ( T1 );
= Op ( p1T );
= Op ( p1N ) + Op ( p1T );
0
0 0
~ ~ Z~2i M~ i Z~2i ) 1 Z~ 0 M
~
i F Mw Mi Z0i (
i i Mw F
T
PN
PN
i
~0 M
~ ~
Z
0 0
0 ~ ~
2i i Z2i
~ i Mw F
) 1 Z~ 0 i M
i F Mw Mi Z i (
T
0
~ i Z~0i ( Z~2i M~ i Z~2i ) 1 Z~ 0 M
~
"0i M
i i Mw F
i=1 ~
T
i=1
p1
T
i
0
~ i Z~ i ( Z~2i M~ i Z~2i ) 1 Z~ 0 M
~
"0i M
i i Mw F
i=1 ~
T
i=1
1
T
(v)a.
i=1
PN
(iv)a.
b.
0 ~0
i Z i Mw F
i=1
i
i
= Op ( p1N ) + Op ( p1T );
~0 M
~ ~
~0 M
~ Z
~
Z
Z
0 ~
2i i Z2i
) 1 ( 0i T i 0i ) 1 Z~0i
Mi Mw F i = Op ( pN1 T ) + Op ( T1 );
T
h ~0 ~ ~
i
~0 M
~ ~
Z2i Mi Z2i
Z
0 0
1
1 ~0 ~
0i i Z0i
~
~
F
M
M
Z
(
Z0i Mi Mw F i = Op ( N1 )+Op ( T1 )+
)
(
)
w
i 0i
i
T
T
~ Z
~
~
"0i M
pi 0i
i=1
T
(
Proof of Lemma 7.
(i) Consider the term
Z~ 0 i Mw F
i
= Z 0 i Mw F
i
= Z 0 i FD1
i
+ Z 0 i UD2 i :
k0 j non-zero elements in Z i , Z 0 i F is equal to the sum of jk
k0 j
elements, and is …nite on K(Ck ). Similarly, Z 0 i U is also equal to the sum of jk
k0 j
Since there are jk
elements, each is Op ( p1N ). Due to the fact that D1 = Op ( N1 ) + Op ( pN1 T ) and D2 = Op (1),
Z 0 i Mw F
i
= Op (
1
1
1
1
) + Op ( p
) + Op ( p ) = Op ( p ):
N
NT
N
N
Therefore,
N
X
i=1
0 ~0
i Z i Mw F
i
p
= Op ( N
p
1
p
)O
(
)
=
O
(
p
p
N
N
N ):
The rest of this lemma can be proved similarly. See the Supplementary Appendix for a
detailed proof.
Lemma 2 in Model 1 can be extended to Model 2 using transformed data as Lemma 8.
37
(i)
(ii)
PN
i=1
(iv)
(v)
(vi)
PN
p
= Op (
N );
q
~ 0~
~ 0X
~
"
0 ~0 ~ X
1X
N
i i
pi i = Op (
);
i=1 i Z i Xi ( T )
T
T
q
h
PN 0 ~ 0 ~ ~ Z~2i0 M~ i Z~2i 1 Z~2i0 M~ i~"i i
N
p1
p
(
Z
M
Z
)(
=
O
(
)
);
p
i i 2i
i=1 i
T
T
T
T
PN 0 ~ ~ Z~2i0 M~ i Z~2i 1 ~ 0 ~
1
"i Mi Z i ( T ) Z i Mi~"i = Op ( N
);
i=1 ~
T
T
p1
T
(iii)
0 ~0
"i
i Z i~
q
PN
~0 M
~ Z
~
~ Z
~
Z
~
"0i M
~0 M
~ "i = Op ( N ) + Op ( N );
pi 0i ( 2i i 2i ) 1 Z
i i~
i=1
T
T
T
T
i ~0 ~
0
PN ~"0i M~ i Z~0i h Z~2i0 M~ i Z~2i 1
~
~
~
Z M~
"
Z MZ
p
( T )
( 0i T i 0i ) 1 0ipTi i = Op ( N
):
i=1
T
T
p1
T
When the error ~"i is replaced with ~"0i = Mw F
i
+ ~"i , terms involving Mw F
i
in Lemma
7 are used to obtain the following lemma.
(i)
(ii)
PN
(iii)
(iv)
(v)
(vi)
i=1
0 ~0
"0i
i Z i~
PN
p
= Op (
N );
~ 0X
~
X
i i
q
q
~ ~
X
"
) 1 piTi = Op ( NN ) + Op ( TN );
i=1
T
q
PN 0 ~ 0 ~ ~ Z~2i0 M~ i Z~2i 1 Z~2i0 M~ i~"0i
p1
p
)
= Op ( NN ) +
i=1 i (Z i Mi Z2i )(
T
T
T
PN 00 ~ ~ Z~2i0 M~ i Z~2i 1 ~ 0 ~ 0
1
"i Mi Z i ( T ) Z i Mi~"i = Op ( N
);
i=1 ~
T
T
p1
T
0 ~0 ~
i Z i Xi (
0 0
q
Op (
N
T
q
PN
);
~
~0 M
~ Z
~ Z
~
Z
~
"00
i M
~ "0 = Op ( N ) + Op ( N ) + Op ( p1 );
~0 M
p i 0i ( 2i i 2i ) 1 Z
i
i i~
i=1
T
T
T
T
N
i ~0 ~ 0
0
PN ~"00i M~ i Z~0i h Z~2i0 M~ i Z~2i 1
~0i
~ M
~ iZ
Z
Z
M
~
"
i
N
p
( T )
( 0i T ) 1 0ipT i = Op ( T ) + Op ( N1 ) +
i=1
T
p1
T
Proof of Lemma 9. (i) Plugging in ~"0i = Mw F
N
X
0 ~0
"0i
i Z i~
=
i=1
N
X
i
0 ~0
i Z i Mw F
i
i=1
0 ~0
"i
i Z i~
i=1
+ ~"i , we obtain
+
N
X
0 ~0
"i
i Z i~
i=1
p
Lemma 7(i) shows that the …rst term is Op (
N
X
Op ( pN1 T ):
N ).
p
= Op (
By Lemma 8 (i), the second term
N)
uniformly on K(Ck ). It follows that
N
X
i=1
0 ~0
"0i
i Z i~
p
= Op (
p
)
+
O
(
p
N
p
)
=
O
(
p
N
N ):
The proofs of Lemma 9(ii)-(vi) are included in the Supplementary Appendix.
38
With these lemmas, we are ready to prove Theorem 2.
Proof of Theorem 2. To prove lim(N;T )!1 P (k~ = k0 ) = 1, it is equivalent to show that
> 0, for both large T and N , P (jk~ k0 j
i
P hg
g
that P supK(Ck ) N
SV
(k)
SV
(k
)
0 < ; or
i
i 0
i=1
for any given
P
N
X
sup
J~2i (k)
inf
K(Ck )
K(Ck ) i=1
N
X
1) < . It is su¢ cient to show
!
J~1i (k)
i=1
< :
By Lemma 3, it is su¢ cient to show
N
1 X ~
P ( sup
J2i (k)
K(Ck )
N
~1) < :
i=1
For any k 2 K(Ck ),
N
X
J~2i (k)
i=1
N h
X
0 ~ ~
0 ~ ~
0 ~ 0
2 0i (Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 Z~2i
Mi~"i
0 ~ 0
2 0i Z~0i
Mi~"i
i=1
+
N h
X
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
i
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
i=1
Consider the …rst term, with Z~2i = Z~0i
N h
X
2
0 ~0
"0i
i Z i~
i=1
N
2 X
+p
T i=1
0
i
"
:
Z~ i ,
0 ~ ~
0 ~ ~
0 ~ 0
2 0i (Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 Z~2i
Mi~"i
i=1
N
X
i
N
~
~ 0X
2 X 0 ~0 ~ X
i i
)
Z
+p
X
(
i
i i
T
T i=1
0 ~ 0
2 0i Z~0i
Mi~"i
~ 0 "0
i
1 Xi ~
i
p
T
#
~0 M
~ i~"0
~0 M
~ i Z~2i
Z
Z
~ i Z~2i )( 2i
(Z~ 0 i M
) 1 2ip i :
T
T
By (i), (ii) and (iii) of Lemma 9, the …rst term
N h
X
0 ~ ~
0 ~ ~
0 ~ 0
Mi~"i
2 0i (Z~0i
Mi Z2i )(Z~2i
Mi Z2i ) 1 Z~2i
i=1
39
0 ~ 0
2 0i Z~0i
Mi~"i
i
p
= Op (
N ):
(38)
Now consider the second term
N h
X
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
i=1
N
0 ~ ~
1 X 00 ~ ~ Z~2i
Mi Z2i 1 ~ 0 ~ 0
~"i Mi Z i (
) Z i Mi~"i
T i=1
T
i
N
~ ~ ~0 ~ ~
1 X ~"00
i Mi Z0i Z2i Mi Z2i
~ i~"0
p
+2 p
(
) 1 Z~ 0 i M
i
T
T i=1
T
"
#
N
0 ~ ~
0 ~ ~
0 ~ 0
X
~
~
~
~
~"00
M
Z
Z
M
Z
Z
M
Z
Z~0i
M ~"
i
0i
i
2i
i
0i
i
p
p i i :
+
( 2i
) 1 ( 0i
) 1
T
T
T
T
i=1
By (iv), (iv) and (vi) of Lemma 9, the second term
N h
X
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
i=1
r
1
N
N
) + Op ( p ):
= Op ( ) + Op (
T
T
N
1 ~0 ~ 0
~ ~ ~0 ~ ~
~"00
"i
i
Therefore, we get
r
N
N
1
) + Op (
) + Op ( p )]
N ) + Op (
T
T
N
N i=1
N
r
1
N
1
N
1
1
= Op ( p ) +
[Op ( ) + Op (
)] +
Op ( p ):
T
T
N
N
N
N
N
p
1 X ~
1
J2i (k) =
[Op (
Compared with the proof of Theorem 1, there is an extra term 1 Op ( p1N ) = op (1). But,
N
P
N
1
~
i=1 J2i (k) vanishes for any k 2 K(Ck ) as (N; T ) ! 1, so does its maximum.
N
40
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44

Estimation of Heterogeneous Panels with Structural Breaks"

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