stat 497 lecture note 12

STAT 497LECTURE NOTE 12

COINTEGRATION

1

Multivariate Unit Root Processes

• Generally we cannot reject the null hypothesis, that many time series have unit roots. For example,log consumption and log output are both non-stationary, but log consumption –log output is stationary. This situation is called cointegration. The practical problem is that when we have cointegration, asymptotics change completely. Furthermore, we really do not have enough data to definitively tell whether or not we have cointegrated series.

2


• In a univariate nonstationary time series Yt is said to be integrated of order d, I(d), if its (d1)th difference is nonstationary but d-th difference is stationary.

• If Yt is nonstationary but Yt=(1B)Yt is stationary, then Yt is integrated of order 1.

Yt~I(1) but Yt~I(0)

3


• In many time series, integrated processes are considered together and they form equilibrium relationships.– Short-term and long-term interest rates– Income and consumption• These leads to the concept of cointegration.• The idea behind the cointegration is that

although multivariate time series is integrated, certain linear transformations of the time series may be stationary.

4

Granger Causality Tests• According to Granger, causality can be further sub-

divided into long-run and short-run causality.• This requires the use of error correction models or

VECMs, depending on the approach for determining causality.

• Long-run causality is determined by the error correction term, whereby if it is significant, then it indicates evidence of long run causality from the explanatory variable to the dependent variable.

• Short-run causality is determined as before, with a test on the joint significance of the lagged explanatory variables, using an F-test or Wald test.

5

Granger Causality Tests• Before the ECM can be formed, there first has to

be evidence of cointegration, given that cointegration implies a significant error correction term, cointegration can be viewed as an indirect test of long-run causality.

• It is possible to have evidence of long-run causality, but not short-run causality and vice versa.

• In multivariate causality tests, the testing of long-run causality between two variables is more problematic, as it is impossible to tell which explanatory variable is causing the causality through the error correction term.

6

SPURIOUS REGRESSION• If we regress a y series with unit root on regressors

who also have unit roots the usual t tests on regression coefficients show statistically significant regressions, even if in reality it is not so.

• The Spurious Regression Problem can appear with I(0) series (see Granger, Hyung and Jeon (1998)). This is telling us that the problem is generated by using WRONG CRITICAL VALUES!!!!

• In a Spurious Regression the errors would be correlated and the standard t-statistic will be wrongly calculated because the variance of the errors is not consistently estimated. In the I(0) case the solution is:

7

1/2)ˆ of cerun varian-(long ˆ where,on distributi-t tˆˆˆ

SPURIOUS REGRESSION• How do we detect a Spurious Regression (between I(1)

series)?Looking at the correlogram of the residuals and also by testing for a unit root on them.

• How do we convert a Spurious Regression into a valid regression?By taking differences.

• Does this solve the SPR problem?It solves the statistical problems but not the economic interpretation of the regression. Think that by taking differences we are loosing information and also that it is not the same information contained in a regression involving growth rates than in a regression involved the levels of the variables.

8

SPURIOUS REGRESSIONTypical symptom: “High R2, t-values, F-value, but low DW”

1. Egyptian infant mortality rate (Y), 1971-1990, annual data, on Gross aggregate income of American farmers (I) and Total Honduran money supply (M)

Y ^ = 179.9 - .2952 I - .0439 M, R2 = .918, DW = .4752, F = 95.17 (16.63) (-2.32) (-4.26) Corr = .8858, -.9113, -.9445

2. US Export Index (Y), 1960-1990, annual data, on Australian males’ life expectancy (X)Y ^ = -2943. + 45.7974 X, R2 = .916, DW = .3599, F = 315.2

(-16.70) (17.76) Corr = .9570

9

SPURIOUS REGRESSION3. US Defense Expenditure (Y), 1971-1990, annual data, on Population

of South African (X)Y ^= -368.99 + .0179 X, R2 = .940, DW = .4069, F = 280.69

(-11.34) (16.75) Corr = .9694

4. Total Crime Rates in the US (Y), 1971-1991, annual data, on Life expectancy of South Africa (X)Y ^= -24569 + 628.9 X, R2 = .811, DW = .5061, F = 81.72

(-6.03) (9.04) Corr = .9008

5. Population of South Africa (Y), 1971-1990, annual data, on Total R&D expenditure in the US (X)Y ^= 21698.7 + 111.58 X, R2 = .974, DW = .3037, F = 696.96

(59.44) (26.40) Corr = .9873

10

SPURIOUS REGRESSION• Does it make sense a regression between two I(1) variables?

Yes if the regression errors are I(0).• Can this be possible?

The same question asked David Hendry to Clive Granger time ago. Clive answered NO WAY!!!!! but he also said that he would think about. In the plane trip back home to San Diego, Clive thought about it and concluded that YES IT IS POSSIBLE. It is possible when both variables share the same source of the I(1)’ness (co-I(1)), when both variables move together in the long-run (co-move), ... when both variables are COINTEGRATED!

11

COINTEGRATION

• An mx1 vector time series Yt is said to be cointegrated of order (d, b), CI(d,b) where 0<bd, if each of its component series Yit is I(d) but some linear combination of the series ’Yt is I(db) for some nonzero constant vector ’.

• ’ is the cointegrating vector or the long run parameter and it is not unique.

• The most common case is d=b=1.

12

COINTEGRATION

• More generally, if the mx1 vector series Yt contains more than two components, each being I(1), then there may exist k (<m) linearly independent 1xm vectors 1’, 2’,…, k’, such that ’Yt is a nonstationary kx1 vector process where

is a kxm cointegrating matrix.• The number of linearly independent cointegrating

vectors is called the cointegrating rank. Yt is cointegrated of rank k.

13

k,, 1

EXAMPLE• Consider the following system of processes

where the three error terms are uncorrelated white noise processes. Clearly, all those three processes are individually I(1). Let yt=(x1t,x2t,x3t)’ and =(1,1,2), then yt=a1t. which is a I(0) process. Another cointegrating relationship is between x2t and x3t. So, we can let *= (0, 1,−3), then *yt = a2t is also I(0).

tt,t

ttt

tttt

axx

axx

axxx

3133

2332

132211

14

VAR with Cointegration• Let Yt be mx1. Suppose we estimate VAR(p)

or • Let say we have a unit root. Then, we can

write

• This is like a multivariate version of the augmented Dickey- Fuller test

tptptt aYYY 11

.aYBY t1tt

BBB * 11

.aYYY t

p

iittt

11

15

VAR with Cointegration• Rearranging the equation

where Rank((1)I)<m. There are two cases:• (1)= I the n we have m independent unit roots, so

there is no cointegration, and we should run the VAR in differences.

• 0<Rank((1)I)=k<m, then we can write (1)I =’ where and are mxk. The equation becomes:

This is called a vector error correction model (VECM).

.B tt*

t aYYYt 1111

.B tt*

t aYYYt 11

16

VAR with Cointegration• Note that if you run OLS in differences, then the

modeled is misspecified and the results will be biased. What can you do?

(a)If you know the location of the unit roots and cointegration relations, then you can run the VECM by doing OLS of Yt on lags of Y and ’Yt1.

(b)If you know nothing, then you can either (i) run OLS in levels, or (ii) test (many times) to estimate cointegrating relations, and run VECM. The problem with this approach is that you are testing many times and you are estimating cointegrating relationships. This leads to poor finite sample properties.

17

Residual Based Tests of the Null of No Cointegration

• Procedures designed to distinguish a system without cointegration from a system with at least one cointegrating relationship; they do not estimate the number of cointegrating vectors (the k). Tests are conditional on a pretest for unit roots in each of the variables.

• When the cointegration vector is known: construct the hypothesized linear combination that is stationary, treat it as data, and apply a Dickey-Fuller unit root test to that linear combination. The null hypothesis is that there is a unit root, or no cointegration.

18


• When the cointegration vector is not known: Assume that, if there exists a cointegrating relation, the coefficient on Y1t is nonzero, allowing us to express the “static regression equation as

• You can apply a unit root test to the estimated OLS residual from estimation of the above equation, but– Include a constant in the static regression if the alternative

allows for a nonzero mean in ut

– Include a trend in the static regression if the alternative is stochastic cointegration, i.e., a nonzero trend for A’Yt.

t2t1t uY Y

19


• The first step in testing cointegration is to test the null hypothesis of a unit root in each component series Yit individually using the univariate unit root tests.

• If the hypothesis is rejected, then the next step is to test cointegration among the components, i.e., to test whether ’Yt is stationary.

20


• In practice, the cointegration vector is unknown. One way to test the existence of cointegration is the regression method (Engle&Granger, 1986, 1987).

• If Yt=(Y1t,Y2t,…,Ymt) is cointegrated, ’Yt is stationary where =(1, 2,…, m). Then, (1/1) is also a cointegrated vector where 10.

21


• Consider the regression model for Y1t

and check whether t is I(1) or I(0).

• If t~I(1), then Yt is not cointegrated.

• If t~I(0), then Yt is cointegrated with a normalizing cointegrating vector

’=(1,1,…, m1) .22

tmtmtt YYY 1211


• In testing the error series for nonstationary,– Calculate the OLS estimate– Use the residual series for the test using the

standard ADF or PP .– if t~I(0).

– H0: =1 vs H1: <1 for the model

– H0: =0 vs H1: <1 for the model

23

.ˆ,,ˆ1,ˆ 1m1

t

ondistributit~ˆ.asympt

i

t

p

jjtjtt a

1

11

t

p

jjtjtt a

1

11


• t-statistic:

• The critical values are obtained by simulation (Engle&Granger, 1987).

Level of significance 1% 5% p=1 4.07 3.37 p>1 3.73 3.17

• If T<Critical Value, reject H0Cointegration exists.

24

ˆs

ˆT

The Johansen Trace and Maximal Eigenvalue Tests

• To test whether the variables are cointegrated or not, one of the well-known tests is the Johansen trace test. The Johansen test is used to test for the existence of cointegration and is based on the estimation of the ECM by the maximum likelihood, under various assumptions about the trend or intercepting parameters, and the number k of cointegrating vectors, and then conducting likelihood ratio tests.

25


• Assuming that the ECM errors are independent Nm[0, ] distribution, and given the cointegrating restrictions on the trend or intercept parameters, the maximum likelihood Lmax(k) is a function of the cointegration rank k.

• The trace test is based on the log-likelihood ratio ln[Lmax(k)/Lmax(k)], and is conducted sequentially for k = m-1,...,1,0. The name comes from the fact that the test statistics involved are the trace (the sum of the diagonal elements) of a diagonal matrix of generalized eigenvalues. This test examines the null hypothesis that the cointegration rank is less than or equal to k, against the alternative that the cointegration rank is greater than k. If the trace is greater than the critical value for a certain rank, then the null hypothesis that the cointegration rank is equal to k is rejected.

26


• Consider a non-stationary cointegrated VAR(p) model

where at are normally distributed with mean 0 and covariance matrix . In a series of influential papers, Johansen (1988, 1991), and Johansen and Juselius (1990) proposed practical full maximum likelihood estimation and testing approaches based on the error correction representation (ECM).

27

p1( B ... B ) p t tI x a


• Consider the ECM

where ,dt is a vector of deterministic variables, such as constant and seasonal dummy variables,are m×m, ,A and are m×k parameter matrices, the are i.i.d. Nm(0, ) errors, and

det( ) has all of its roots outside the unit circle.

28

1

01

p

t t j t j t p tj

x d x x a

1 t t tx x x

1 , j = 1, , p-1j jI

A

1

j

1

Bp

j

j

I


• This ECM is based on the Engle-Granger (1987) error correction representation theorem for cointegrated systems, and the asymptotic inference involved is related to the work of Sims, Stock, and Watson (1990).

• By step-wise concentrating all the parameter matrices in the likelihood function out except for the matrix A, Johansen shows that the maximum likelihood estimator of A can be derived as the solution of a generalized eigenvalue problem. Likelihood ratio tests of hypotheses about the number of cointegrating vectors can then be based on these eigenvalues. Moreover, Johansen (1988) also proposes likelihood ratio tests for linear restrictions on these cointegrating vectors.

29


• The Johansen test for the existence of cointegration is based on the estimation of the above ECM by the maximum likelihood and is used to test the hypothesis

, where k is less than m. This formulation shows that I(1) models form nested sequence models

where H(m) is the unrestricted VAR model or I(0) model, and H(0) corresponds to the restriction =0, which is the VAR model for indifferences. Since , it is equivalent to test that A and are of full column rank k, the number of independent cointegrating vectors that forms the matrix A. The test has been named the Johansen trace test because the likelihood ratio test statistic is the trace of a diagonal matrix of generalized eigenvalues from .

30

A

kRank:H 0

mHkHH 0

Sequential tests:i. H0: k=0, cannot be rejected →stop

(at most zero coint) rejected →next testii. H0: k<=1, cannot be rejected →stop→k=1

(at most one coint) rejected →next testiii. H0: k2, cannot be rejected →stop→k=2

(at most two coint) rejected →next test

31



(i) Rank k = m: all variables in x are I(0), not an interesting case to start with.

(ii) Rank k = 0: there are no linear combinations of x that are I(0), no cointegration exists, and is full of zeros. Model on differenced series

(iii) Rank k (m-1): up to (m-1) cointegration relationships ´xt-k.i.e. k (m-1) rows of form k linearly independent combinations of variables in x, each of which is I(0); alternatively (m-k) nonstationary vectors forming I(1) stochastic trends.

32


• Under some regularity conditions, we can write the cointegrated process as an Error Correction Model (ECM):

where is the difference operator , the at's are i.i.d. N(0, ).

33

1 1 1 1... t t p t p t p tx x x x a


• We can write this ECM as

where , , ,• The likelihood ratio statistic for hypothesis is given by

where denotes the eigenvalues of and are ordered by

34

0 1 t t pt tZ Z Z a

0 t tZ x 1 1 1( ,..., ) t t t pZ x x pt t pZ x 1 1( ,..., ) p

0H : A

i pppp S wrtSSS 01

000

m

kii

ˆlnnln1

12

.ˆˆm 01


Where

• If the test statistics are greater than the critical value for rank k, then the null hypothesis that the cointegration rank is equal to k is rejected.

35

ijiijij MMMMS 1111

.p,,j,i;ZZnMn

tjtitij 10

1

1


• The statistic ln has the following limiting distribution which can be expressed in terms of a mk dimensional Brownian motion as

• The percentiles of the asymptotic distribution for the trace statistic are tabulated in Johansen (1988, Table 1) using simulation analysis.

36

11 1 1

0 0 0

tr d dY Y YY dt Y Y


• An alternative LR statistic, given by

and called the maximal eigenvalue statistic, examines the null hypothesis of k cointegrating vectors versus the alternative k+1 cointegrating vectors. The asymptotic distribution of this statistic is given by the maximum eigenvalue of the stochastic matrix in

37

11 1 1

0 0 0

d dt dY Y YY Y Y

112 kˆlnnln

Analysis of U.S. Economic Variables (From SAS Online Doc)

• Consider the following four-dimensional system of U.S. economic variables. Quarterly data for the years 1954 to 1987 are used (Lütkepohl 1993, Table E.3.). The following statements plot the series and proceed with the VARMAX procedure.

38

SAS Codesymbol1 v=none height=1 c=black;

symbol2 v=none height=1 c=black;

title 'Analysis of U.S. Economic Variables';

data us_money;

date=intnx( 'qtr', '01jan54'd, _n_-1 );

format date yyq. ;

input y1 y2 y3 y4 @@;

y1=log(y1);

y2=log(y2);

label y1='log(real money stock M1)' y2='log(GNP in bil. of 1982 dollars)' y3='Discount rate on 91-day T-bills' y4='Yield on 20-year Treasury bonds';

datalines;

... data lines omitted ... ;

legend1 across=1 frame label=none;

39

SAS Code (Contd.)proc gplot data=us_money;

symbol1 i = join l = 1;


axis2 label = (a=-90 r=90 " ");

plot y1 * date = 1 y2 * date = 2 / overlay vaxis=axis2 legend=legend1;

run;

proc gplot data=us_money;



axis2 label = (a=-90 r=90 " ");

plot y3 * date = 1 y4 * date = 2 / overlay vaxis=axis2 legend=legend1;

run;

proc varmax data=us_money;

id date interval=qtr;

model y1-y4 / p=2 lagmax=6 dftest print=(iarr(3)) cointtest=(johansen=(iorder=2)) ecm=(rank=1 normalize=y1); cointeg rank=1 normalize=y1 exogeneity;

run;

40

SAS Output

• This example performs the Dickey-Fuller test for stationarity, the Johansen cointegrated test integrated order 2, and the exogeneity test. The VECM(2) fits the data. From the outputs shown below, you can see that the series has unit roots and is cointegrated in rank 1 with integrated order 1. The fitted VECM(2) is given as

41

SAS Output

42

SAS OutputDickey-Fuller Unit Root Tests

Variable Type Rho Pr < Rho Tau Pr < Tau

y1 Zero Mean 0.05 0.6934 1.14 0.9343

Single Mean -2.97 0.6572 -0.76 0.8260

Trend -5.91 0.7454 -1.34 0.8725

y2 Zero Mean 0.13 0.7124 5.14 0.9999

Single Mean -0.43 0.9309 -0.79 0.8176

Trend -9.21 0.4787 -2.16 0.5063

y3 Zero Mean -1.28 0.4255 -0.69 0.4182

Single Mean -8.86 0.1700 -2.27 0.1842

Trend -18.97 0.0742 -2.86 0.1803

y4 Zero Mean 0.40 0.7803 0.45 0.8100

Single Mean -2.79 0.6790 -1.29 0.6328

Trend -12.12 0.2923 -2.33 0.4170

43

SAS Output

44

Cointegration Rank Test for I(2)

r\k-r-s 4 3 2 1 Traceof I(1)

5% CV of I(1)

0 384.60903 214.37904 107.93782 37.02523 55.9633 47.21

1 219.62395 89.21508 27.32609 20.6542 29.38

2 73.61779 22.13279 2.6477 15.34

3 38.29435 0.0149 3.84

5% CV I(2) 47.21000 29.38000 15.34000 3.84000

SAS Output

45

Long-Run ParameterBeta Estimates When

RANK=1

Variable 1

y1 1.00000

y2 -0.46458

y3 14.51619

y4 -9.35520

Adjustment CoefficientAlpha Estimates When

RANK=1

Variable 1

y1 -0.01396

y2 -0.02811

y3 -0.00215

y4 0.00510

Diagnostic Checks

46

Schematic Representation of Cross Correlationsof Residuals

Variable/Lag 0 1 2 3 4 5 6

y1 ++.. .... ++.. .... +... ..-- ....

y2 ++++ .... .... .... .... .... ....

y3 .+++ .... +.-. ..++ -... .... ....

y4 .+++ .... .... ..+. .... .... ....

+ is > 2*std error, - is < -2*std error, . is between

Portmanteau Test for Cross Correlationsof Residuals

Up To Lag DF Chi-Square Pr > ChiSq

3 16 53.90 <.0001

4 32 74.03 <.0001

5 48 103.08 <.0001

6 64 116.94 <.0001

Diagnostic Checks

47

Univariate Model ANOVA Diagnostics

Variable R-Square StandardDeviation F Value Pr > F

y1 0.6754 0.00712 32.51 <.0001

y2 0.3070 0.00843 6.92 <.0001

y3 0.1328 0.00807 2.39 0.0196

y4 0.0831 0.00403 1.42 0.1963

Univariate Model White Noise Diagnostics

Variable DurbinWatson

Normality ARCH

Chi-Square

Pr > ChiSq F Value Pr > F

y1 2.13418 7.19 0.0275 1.62 0.2053

y2 2.04003 1.20 0.5483 1.23 0.2697

y3 1.86892 253.76 <.0001 1.78 0.1847

y4 1.98440 105.21 <.0001 21.01 <.0001

Diagnostic Checks

48

Univariate Model AR Diagnostics

VariableAR1 AR2 AR3 AR4

F Value Pr > F F Value Pr > F F Value Pr > F F Value Pr > F

y1 0.68 0.4126 2.98 0.0542 2.01 0.1154 2.48 0.0473

y2 0.05 0.8185 0.12 0.8842 0.41 0.7453 0.30 0.8762

y3 0.56 0.4547 2.86 0.0610 4.83 0.0032 3.71 0.0069

y4 0.01 0.9340 0.16 0.8559 1.21 0.3103 0.95 0.4358

Diagnostic Checks

49

Testing Weak Exogeneity ofEach Variables

Variable DF Chi-Square Pr > ChiSq

y1 1 6.55 0.0105

y2 1 12.54 0.0004

y3 1 0.09 0.7695

y4 1 1.81 0.1786

Whether each variable is the weak exogeneity of other variables. The variable y1 is not the weak exogeneity of other variables, y2, y3, and y4; the variable y2 is not the weak exogeneity of other variables, y1, y3, and y4.

If a variable can be taken as "given" without losing information for the purpose of statistical inference, it call weak exogenous.

Weak exogeneityLong-run noncausality

stat 497 lecture note 12

Documents

evidence of long run

multivariate causality

multivariate time series

granger causality testsbefore

i0 series

i1 series

evidence of cointegration

y series