Econometrics I: Univariate Time Series Econometrics (1)

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1 Econometrics I: Dipartimento di Economia Politica e Metodi Quantitativi University of Pavia

2 Overview of the Lecture 1 st EViews Session VI: Some Theoretical Premises 2

3 Overview of the Lecture 1 st EViews Session VI: Some Theoretical Premises 2 nd EViews Session VII: An AR model for the Italian Unemployment Rate 2-a

4 Overview of the Lecture 1 st EViews Session VI: Some Theoretical Premises 2 nd EViews Session VII: An AR model for the Italian Unemployment Rate 3 rd EViews Session VIII: Simulation of ARMA models 2-b

5 Overview of the Lecture 1 st EViews Session VI: Some Theoretical Premises 2 nd EViews Session VII: An AR model for the Italian Unemployment Rate 3 rd EViews Session VIII: Simulation of ARMA models 4 th EViews Session IX: The Box-Jenkins Empirical Analysis + Exercises 2-c

6 EViews Session VI: Some Theoretical Premises Autocovariance The j-th autocovariance of Y t is given by: Cov[Y t, Y t j ] γ t,t j E[(Y t E[Y t ])(Y t j E[Y t j ])], correspondingly the variance of Y t is defined as: V[Y t ] γ t,t E[(Y t E[Y t ]) 2 ]. Autocorrelation The j-th autocorrelation of Y t is given by: Corr[Y t, Y t j ] ρ t,t j Cov[Y t, Y t j ] V[Y t ] 1 2 V[Yt j ] 2 1 = γ t,t j γ 1 2 t,t γ 1 2 t j,t j 3

7 EViews Session VI: Some Theoretical Premises Covariance Stationarity A time series {Y t } t= is called covariance stationary, weakly stationary, or second order stationary if E[Y t ] = µ Y <, V[Y t ] = γ t,t = σ 2 Y <, Cov[Y t, Y t j ] = γ t,t j = γ j <, t t t, j Autocovariance Function The autocovariance function (ACVF) of a covariance stationary process {Y t } t= is the sequence of autocovariances γ j for all j = 0, 1, 2,.... Autocorrelation Function The autocorrelation function (ACF) of a covariance stationary process {Y t } t= is the sequence of autocorrelations ρ j for all j = 0, 1, 2,.... 4

8 EViews Session VI: Some Theoretical Premises Empirical Autocorrelation Function The empirical (or sample) autocorrelation function of a time series Y t is the sequence of sample autocorrelation coefficients ˆρ j for all j = 0, 1, 2,..., where, ˆρ j = ˆγ j / ˆγ 0 = T t=j+1 (Y t Ȳ )(Y t j Ȳ ) T t=1 (Y t Ȳ )2 with ˆγ j = 1 T Ȳ = 1 T T t=j+1 T t=1 Y t (Y t Ȳ )(Y t j Ȳ ) The graphical depiction of the empirical autocorrelation function is called autocorrelogram. 5

9 EViews Session VI: Some Theoretical Premises Partial Autocorrelation Function Consider the autoregressive process of order p: Y t = c + φ 1p Y t 1 + φ 2p Y t φ pp Y t p + ǫ t, where φ ip, i p is the coefficient on the lagged value Y t p in the AR(p) model. Then, the sequence of the last coefficients φ 11, φ 22,..., φ pp is called the partial autocorrelation function (PACF). The partial autocorrelation function is the second major explorative tool in time series analysis besides the autocorrelation function and is also used to describe the properties of linear time series processes. In order to understand its meaning let us consider a linear model that explains Y t by its j latent historical values: Y t = φ 1j Y t 1 + φ 2j Y t φ jj Y t j + u t (1) = X tφ j + u t with X t = (Y t 1, Y t 2,..., Y t j ) and φ j = (φ 1j, φ 2j,..., φ jj ). Without loss of generality we assume that Y t is centered so that no intercept occurs. 6

10 EViews Session VI: Some Theoretical Premises Assume now that a time series is generated by an AR(p) process, where p < j. In this case only the p lagged values of Y t are useful to predict Y t and the remaining j p coefficients would be zero. In particular the coefficient on the lagged dependent variable Y t j, φ jj, which is the j t h partial autocorrelation coefficient would be zero. Since this result holds for every integer j > p, all partial autocorrelation coefficients φ p+1,p+1, φ p+2,p+2,... would be zero. On the other hand if the series is generated by an AR(p) with p j the coefficient φ jj must be different from zero. However, if we consider an MA(p)-process, we ll see that the partial autocorrelations die out at infinity: hence φ jj j, would be different from zero. Moreover, the autocorrelations die out at p, instead. (see pag. 151 onwards of the book I gave you for some examples). 7

11 EViews Session VII: An AR model for the Italian Unemployment Rate a) Open the workfile UTS_mod.WF1. You find the following variables: Variable Description u Italian unemployment rate lwp logs of the real wage b) Estimate the following white noise model, u t = c + ε t, ε t n.i.d.(0, σ 2 ) where u t is normally and independently distributed over time with constant variance and constant mean. Is it an appropriate model for the Unemployment rate?... 8

12 EViews Session VII: An AR model for the Italian Unemployment Rate... NO! The white noise model does not fit actual data for u because it does not feature the time series most common characteristic: PERSISTENCE. In fact, the actual u is by far more persistent than the simple WN process under and above the natural rate of about 7%. In order to have an idea, generate a corresponding artificial series with same sample mean and standard deviation of the historical u. genr uaswn = * nrnd genr meanline = 6.94 plot u uaswn meanline 9

13 EViews Session VII: An AR model for the Italian Unemployment Rate 10

14 EViews Session VII: An AR model for the Italian Unemployment Rate c) Now consider another stochastic process, the AR(1) model, u t = c + αu t 1 + ε t, ε t n.i.d.(0, σ 2 ) where u t is not independently distributed over time because it depends on u t 1. We can estimate c and α by using the OLS method. What can we conclude? How are the diagnostic tests (no autocorrelation, no heteroschedasticity, normality)?... 11

15 EViews Session VII: An AR model for the Italian Unemployment Rate The estimate of α parameter is very close to one; The AR(1) model fits unemployment quite well Diagnostic tests: Under the null: no autocorrelation no heteroschedasticity normality AR(1) residuals rejected not rejected not rejected 12

16 EViews Session VII: An AR model for the Italian Unemployment Rate d) Expand the previous model by estimating the AR(2) model, u t = c + α 1 u t 1 + α 2 u t 2 + ε t, ε t n.i.d.(0, σ 2 ) where there is one more parameter, and the dynamics is extended to the second lag. What do you find? The residual tests are all fine (white noise errors); The AR(2) model equally fits well; The sum of the two a estimates is close to one. 13

17 EViews Session VII: An AR model for the Italian Unemployment Rate e) When we deal with real data, not all the economic series are untrended, and in case of trended variables we must introduce deterministic components in our statistical models in order to (potentially) give account of this further feature. Consider the initial white noise model and include a deterministic trend to model the logs of the real wage: lwp t = c + βt + ε t, ε t n.i.d.(0, σ 2 ) How are the residuals diagnostics? If you then include an AR(1) dynamics what are your findings? 14

18 EViews Session VII: An AR model for the Italian Unemployment Rate 1) Under the null: no autocorrelation no heteroschedasticity normality Only trend residuals rejected rejected not rejected 15

19 EViews Session VII: An AR model for the Italian Unemployment Rate 2) Under the null: no autocorrelation no heteroschedasticity normality Trend + AR(1) residuals not rejected not rejected not rejected 16

20 EViews Session VII: An AR model for the Italian Unemployment Rate Some final comments: Despite the inclusion of a deterministic trend, lwp persistence needs an AR dynamics. in general many economic series can be represented by AR models of different orders, with or without deterministic trends; The (sum of) AR parameter estimates is very often close to one... (evidence of non-stationarity)... 17

21 EViews Session VIII: Simulation of ARMA models Eviews program for Simulating an AR(1) model Determine AR and constant coefficients!phi1=0.5!c=0.1 error term series u=nrnd initialization of time series smpl series ar1=!c/(1-!phi1) generation of time series smpl series ar1=!c+!phi1*ar1(-1)+ u smpl delete u 18

22 EViews Session VIII: Simulation of ARMA models Eviews program for Simulating a MA(1) model Determine constant and MA coefficients!mu=0.1!theta=-0.5 innovation term series z=nrnd initialization series eps=z series r=!mu generation of time series smpl eps=z r=!mu+!theta*eps(-1)+eps smpl delete z 19

23 EViews Session VIII: Simulation of ARMA models Eviews program for Simulating an ARMA(1,1) model Determine AR, MA and constant coefficients!phi1=0.6!theta1=0.4!c=0.1 error term series u1=nrnd initialization of time series smpl series eps1=u1 series arma1=!c/(1-!phi1) generation of time series smpl series arma1=!c+!phi1*arma1(-1)+!theta1*eps1(-1)+eps1 smpl delete u1 u2 eps1 eps2 20

24 EViews Session IX: The Box-Jenkins Empirical Analysis Model Identification 1. Identification of the appropriate order of differencing d in an ARIMA(p,d,q) 2. Identification of the lag order p and q of an ARMA model. Calculation of descriptive statistics of the the time series mean variance ACF PACF 3. Comparison of the descriptive statistics with their theoretical counterparts that would hold true if a certain model is adequate Model Estimation Estimation of the ARMA(p, q) models. Model Evaluation Evaluation of the estimation results and diagnostics: 21

25 EViews Session IX: The Box-Jenkins Empirical Analysis 1. Having estimated an ARMA-model it is common to check whether the selection of the order of the process is correct. In case of a dynamic misspecification the residuals e t should reveal some type autocorrelation. Testing for the white noise property of e t can be done by applying a portmanteau test to the residuals. Note, that we test for the absence of autocorrelation up to the chosen lag order k, actually. Thus, a rejection of the null, rejects that the residual time series is a white noise. In the case, where we cannot reject the null, further test are necessary to verify, whether the residual time series is a white noise. The Ljung-Box statistics for a test of absence of autocorrelation up to order k based on the residuals of an ARMA(p, q) model is 22

26 EViews Session IX: The Box-Jenkins Empirical Analysis Q LB (k) = T(T + 2) k j=1 ˆρ j (e) 2 T j a χ 2 (k p q), where ˆρ j (e) denotes the j-th empirical autocorrelation coefficient of the residual time series e t. Diagnostics based on residuals: Lagrange Multiplier test to test an AR(p) model against an AR(p + r) model (Godfrey, 1978). Auxiliary regression: e t = α 1 Y t α p Y t p + β 1 e t β r e t r + ν t, where ê t are the residuals of the AR(p) model. LM test-statistic tests joint significance of the parameter β 1,..., β r. Test statistic LM = T R 2 χ 2 (r) where R 2 is the (uncentered) coefficient of determination from (2). diagnostic testing for homoscedasticity of the residuals: White test, ARCH test (more details later) diagnostic testing for normality of the residuals (Jarque-Bera t.) (2) 23

27 EViews Session IX: The Box-Jenkins Empirical Analysis 2. Model selection by evaluating the in-sample fit. Principle: Comparing in-sample fit, measured by the based on residual variance, to the number of estimated parameters. Models are selected by minimzing the information criteria. Akaike Information Criterion (AIC): AIC(k) = T ln ˆσ 2 + 2k Bayes Information Criterion (BIC) or Schwartz (SIC) Information Criterion: BIC(k) = T ln ˆσ 2 + k ln T with ˆσ 2 = 1/T T t=1 e2 t. Respecification of the Model if Necessary Repeat all previous steps. Forecasting...in the following days 24

28 EViews Session IX: Exercises Do on your own the following exercises: 1. Simulate an estimate an AR(2), a MA(2) and an ARMA(2,2) model. 2. Find out the mysterious series contained in the workfile mystery.wf1! 25