forecasting seasonal time series 1. introduction seasonal time series 1. introduction philip hans...
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Forecasting
Seasonal Time Series
1. Introduction
Philip Hans Franses
Econometric Institute
Erasmus University Rotterdam
SMU and NUS, Singapore, April-May 2004
1
1. Introduction
• How do seasonal time series (in macroeco-
nomics, finance and marketing) look like?
• What do we want to forecast?
• Why is seasonal adjustment often problem-
atic?
3
The contents of this tutorial are largely based
on
• Ghysels and Osborn (2001), The Econo-
metric Analysis of Seasonal Time Series,
Cambridge UP
• Franses and Paap (2004), Periodic Time
Series Models, Oxford UP.
4
These series are the running examples through-
out the lectures:
• Private consumption and GDP for Japan,
quarterly, 1980.1-2001.2 (source: econo-
magic.com)
• M1 for Australia, quarterly, 1975.2-2004.1
(source: economagic.com)
• Total industrial production for the USA,
monthly, 1919.01-2004.02 (source: econo-
magic.com)
• Decile portfolios (based on size), NYSE,
monthly, 1926.08-2002.12 (source: web-
site of Kenneth French) (ranked according
to market capitalization)
• Sales of instant decaf coffee, Netherlands,
weekly, 1994 week 29 to 1998 week 28
5
How do seasonal time series
(in macroeconomics, finance and market-
ing) look like?
Tools:
• Graphs (over time, or per season)
• Autocorrelations (after somehow removingthe trend, usually: ∆1yt = yt − yt−1)
• R2 of regression of ∆1yt on S seasonaldummies or on S (or less) sines and cosines
• Regression of squared residuals from ARmodel for ∆1yt on constant and S−1 sea-sonal dummies (to check for seasonal vari-ation in error variance)
• Autocorrelations per season (to see if thereis periodicity).
Note: these are all just first-stage tools, tosee in what direction one could proceed. Theyshould not be interpreted as final models.
6
The ”split seasonals” graph for consumption
indicates a rather constant pattern.
The R2 of the ”seasonal dummy regression”
for ∆1yt is 0.927 for log(consumption) and
log(income) it is 0.943.
A suitable model for both log(consumption)
and log(income) turns out to be
∆1yt =4∑
s=1
δsDs,t+ρ1∆1yt−1+εt+θ4εt−4 (1)
The R2 of these models is 0.963 and 0.975, re-
spectively. Hence, constant deterministic sea-
sonality accounts for the majority of trend-free
variation in the data.
9
The log of Australian M1 obeys a trend, andsome seasonality, although at first sight notvery regular.
A regression of the growth rates (differencesin logs) on seasonal dummies gives an R2 of0.203.
An AR(5) fits the data for ∆1yt (although theresiduals are not entirely ”white”). The re-gression of the squares of these residuals onan intercept and 3 seasonal dummies gives anF− value of 5.167, hence there is seasonal vari-ation in the variance.
If one fits an AR(2) model for each of theseasons, that is, regress ∆1yt on ∆1yt−1 and∆1yt−2 but allow for different parameters forthe seasons, then the estimation results forquarters 1 to 4 are (0.929, 0.235), (0.226,0.769), (0.070, -0.478), and (0.533, -0.203).This suggests that different models for differ-ent seasons might be useful.
12
The graphs suggest that the data might be
nonlinear and that there is a break in the vari-
ance, see lecture 4. For now, this is ignored.
A regression of ∆1 of log(production) on S =
12 seasonal dummies gives an R2 of 0.374.
Adding lags at 1, 12 and 13 to this auxiliary
regression model improves the fit to 0.524 (for
1008 data points!), with parameters 0.398, 0.280
and -0.290. There is no residual autocorrela-
tion. The sum of these parameters is 0.388.
Hence, for multi-step ahead forecasts, constant
seasonality dominates. There is also no sea-
sonal heteroskedasticity.
15
The returns (yt) of the 10 decile portfolios
(based on size), NYSE, monthly, 1926.08-2002.12,
can be best described by
yt =12∑
s=1
δsDs,t + εt (2)
εt = ρ1εt−1 + ut (3)
One might expect a ”January effect”, and more
so for the smaller stocks.
The graphs shows the estimates of δs for the
first two and last two deciles. Also the R2
values are higher for lower deciles.
16
Comparing the estimated parameters for the
decile models and their associated standard er-
rors suggests that only a few parameters are
significant.
This is even more so in case one has, say,
weekly data, which are common in marketing
and usually coined as store-level data.
One might then consider
yt =52∑
s=1
δsDs,t + εt, (4)
but that involves a large amount of param-
eters. One can then reduce this number by
deleting certain Ds,t variables, but this might
complicate the interpretation.
18
In that case, a more sensible model is
yt = µ +26∑
k=1
[αk cos(2πkt
52) + βk sin(
2πkt
52)] + εt,
(5)
where t = 1,2, .....
A cycle within 52 weeks (an annual cycle) then
corresponds with k = 1, and a cycle within 4
weeks corresponds with k = 13. Other reason-
able cycles would correspond with 2 weeks, 13
weeks and 26 weeks (k = 26, 4 and 2, respec-
tively). Note that sin(2πkt52 ) is equal to 0 for
k = 26, hence the µ.
The next graph shows the fit of this model,
where εt is an AR(1) process, and where only
cycles within 2, 4, 13, 26 and 52 weeks are con-
sidered, and hence there are only 9 variables to
characterize seasonality. The fit is 0.139.
19
What do we want to forecast?
When one constructs forecasting models fordisaggregated data, one usually want to usethe models for out-of-sample forecasting ofsuch data. Hence, for seasonal data with sea-sonal frequency S, one usually considers fore-casting 1-step ahead, S
2-steps ahead or S-stepsahead.
One may also want to forecast 1 to S stepsahead, that is, say, a year. It is unknownwhether it pays off to use models for disaggre-gated data. Hence, if one wants to forecasta year ahead, it is uncertain whether a modelfor the monthly data does better than a modelfor annual data. Indeed, there are more data,but also more noise and more effort needed tomake a model.
If one aims to forecast a year ahead, one canalso use a model for seasonally adjusted data.(Otherwise : better not!)
21
Why is seasonal adjustment often prob-
lematic?
Seasonal adjustment is usually applied to macroe-
conomic series, in order to make the trend and
the cycle more visible. However, often only
seasonally adjusted data are officially released.
• Seasonally adjusted data are estimated data,
and one should better provide standard er-
rors (Koopman, Franses, OBES 2003)
• Estimated innovations do not concern true
innovations
• Uncertainty about the variable to be fore-
cast: what is it?
• Correlations across variables disturbed
• Why not just take ∆Syt? or ∆1yt after
removal of seasonal dummies?
• Season, trend and cycle can be related.
22
Conclusion for lecture 1
There is substantial seasonal variation in many
economic and business time series.
It seems wise to make a model for the seasonal
data, in order to forecast out of sample, but
also to construct multivariate models.
As we will see in the next lectures: making
models for seasonal data is not that difficult.
23
Forecasting
Seasonal Time Series
2. Basic Models
Philip Hans Franses
Econometric Institute
Erasmus University Rotterdam
SMU and NUS, Singapore, April-May 2004
1
2. Basic Models
• Constant seasonality (seasonal dummies,
sines and cosines)
• Seasonal random walk
• Airline model
• Basic structural model
3
Constant seasonality
Why could the seasonal pattern be constant?
• Weather: harvests, ice-free lakes and har-
bors, consumption patterns, but also: mood
(think of consumer survey data with sea-
sonality due to less good mood in October
and better mood in January)
• Calendar: festivals, holidays
• Institutions: Tax year, end-of-year bonus,
13-th month salary, children’s holidays
4
A general model for constant seasonality is
yt = µ +
S2∑
k=1
[αk cos(2πkt
S) + βk sin(
2πkt
S)] + ut,
(1)
where t = 1,2, .... and ut is some ARMA type
process.
For example, for S=4, one has
yt = µ+α1 cos(1
2πt)+β1 sin(
1
2πt)+α2 cos(πt)+ut,
(2)
where cos(12πt) is the variable (0,-1,0,1,0,-1,...)
and sin(12πt) is (1,0,-1,0,1,0,...) and cos(πt) is
(-1,1,-1,1,...).
If one considers yt + yt−2, which can be writ-
ten as (1 + L2)yt, then (1 + L2) cos(12πt) = 0
and (1 + L2) sin(12πt) = 0. Moreover, (1 +
L) cos(πt) = 0, and of course, (1− L)µ = 0.
5
This shows that deterministic seasonality can
be removed by applying the transformation (1−L)(1+L)(1+L2) = 1−L4. Note that this also
implies that the error term becomes (1−L4)ut.
Naturally, the relation with the often applied
seasonal dummies regression, that is
yt =4∑
s=1
δsDs,t + ut, (3)
is that µ =∑4
s=1 δs, that α1 = δ4 − δ2, β1 =
δ1 − δ3 and α2 = δ4 − δ3 + δ2 − δ1
6
The constant seasonality model seems appli-
cable to many marketing and tourism series, a
little less so in finance, but it is often not fully
descriptive for macroeconomic series.
Many macroeconomic series display some form
of changing seasonality.
Why could the seasonal pattern be changing?
• Changing consumption patterns (pay for
holidays in advance, eating ice in winter)
• Changing institutions: Tax year, end-of-
year bonus, 13-th month salary, children’s
holidays
• Different responses to exogenous shocks in
different seasons
• Shocks may occur more often in some sea-
sons.
7
Seasonal random walk
A simple model that allows the seasonal pat-tern to change over time is the seasonal ran-dom walk, given by
yt = yt−S + εt. (4)
To understand why seasonality may change,consider the S annual time series Ys,T . Theseasonal random walk implies that for theseannual series it holds that
Ys,T = Ys,T−1 + εs,T . (5)
Hence, each season follows a random walk, anddue to the innovations, the annual series mayintersect, such that ”summer becomes win-ter”.
From graphs it is not easy to discern whethera series is a seasonal random walk or not. Theobservable pattern depends on the starting val-ues of the time series, see the next graph.
8
0
20
40
60
80
100
-40
0
40
80
120
60 65 70 75 80 85 90 95
Y X
Y: starting values all zeroX: starting values 10,-8,14 and -12
9
So, due to widely varying starting values, the
observed pattern of a seasonal random walk
can be very regular. Of course, the autocorre-
lation function can be helpful, as the applica-
tion of the filter 1−LS to yt should result in a
white noise (uncorrelated) time series.
Even though the realizations of a seasonal ran-
dom walk can show substantial variation, the
out-of-sample forecasts are constant. Indeed,
at time n, the forecasts are
yn+1 = yn+1−S (6)
yn+2 = yn+2−S (7)
: (8)
yn+S = yn (9)
yn+S+1 = yn+1 (10)
yn+S+2 = yn+2 (11)
10
A subtle form of changing seasonality can also
be described by a time-varying seasonal dummy
parameter model, that is, for S = 4,
yt =4∑
s=1
δs,tDs,t + ut, (12)
where
δs,t = δs,t−S + εs,t. (13)
When the variance of εs,t = 0, then the con-
stant parameter model appears. The amount
of variation depends on the variance of εs,t.
The next slide gives a graphical example (for
simulated data). Canova and Hansen (JBES,
1995) take this model to test for constant sea-
sonality.
11
-30
-20
-10
0
10
20
-20
-10
0
10
20
60 65 70 75 80 85 90 95
Y X
X: constant parametersY: parameter for season 1 is random walk
12
Airline model
An often applied model, popularized by Box
and Jenkins (1970) and named after its appli-
cation to monthly airline passenger data, is the
Airline model, given by
(1−L)(1−LS)yt = (1+θ1L)(1+θSLS)εt. (14)
Note that
(1− L)(1− LS)yt = yt − yt−1 − yt−S + yt−S−1
(15)
It can be proved (Bell, JBES 1987) that as
θS = −1, the model reduces to
(1− L)yt =S∑
s=1
δsDs,t + (1 + θ1L)εt. (16)
13
Strictly speaking, the airline model assumes
S + 1 unit roots. This is due to the fact that
for the characteristic equation
(1− z)(1− zS) = 0 (17)
there are S+1 solutions on the unit circle. For
example, if S = 4 the solutions are (1, 1, -1, i,
-i).
This implies a substantial amount of random
walk like behavior, even though it is corrected
to some extent by the (1 + θ1L)(1 + θSLS)εt
part of the model. In terms of forecasting, it
assumes very wide confidence intervals around
the point forecasts.
Some estimation results for the airline model
are given next.
14
Dependent Variable: LC-LC(-1)-LC(-4)+LC(-5)Method: Least SquaresDate: 04/07/04 Time: 09:48Sample(adjusted): 1981:2 2001:2Included observations: 81 after adjusting endpointsConvergence achieved after 10 iterationsBackcast: 1980:1 1981:1
Variable Coefficient Std. Error t-Statistic Prob.
C -0.000329 0.000264 -1.246930 0.2162MA(1) -0.583998 0.088580 -6.592921 0.0000
SMA(4) -0.591919 0.090172 -6.564320 0.0000
R-squared 0.444787 Mean dependent var -5.01E-05Adjusted R-squared 0.430550 S.D. dependent var 0.016590S.E. of regression 0.012519 Akaike info criterion -5.886745Sum squared resid 0.012225 Schwarz criterion -5.798061Log likelihood 241.4132 F-statistic 31.24325Durbin-Watson stat 2.073346 Prob(F-statistic) 0.000000
Inverted MA Roots .88 .58 .00+.88i -.00 -.88i -.88
15
Dependent Variable: LM-LM(-1)-LM(-4)+LM(-5)Method: Least SquaresDate: 04/07/04 Time: 09:50Sample(adjusted): 1976:3 2004:1Included observations: 111 after adjusting endpointsConvergence achieved after 16 iterationsBackcast: 1975:2 1976:2
Variable Coefficient Std. Error t-Statistic Prob.
C -0.000192 0.000320 -0.599251 0.5503MA(1) 0.353851 0.088144 4.014489 0.0001
SMA(4) -0.951464 0.016463 -57.79489 0.0000
R-squared 0.647536 Mean dependent var -3.88E-06Adjusted R-squared 0.641009 S.D. dependent var 0.032112S.E. of regression 0.019240 Akaike info criterion -5.036961Sum squared resid 0.039981 Schwarz criterion -4.963730Log likelihood 282.5513 F-statistic 99.20713Durbin-Watson stat 2.121428 Prob(F-statistic) 0.000000
Inverted MA Roots .99 .00+.99i -.00 -.99i -.35 -.99
16
Dependent Variable: LI-LI(-1)-LI(-12)+LI(-13)Method: Least SquaresDate: 04/07/04 Time: 09:52Sample(adjusted): 1920:02 2004:02Included observations: 1009 after adjusting endpointsConvergence achieved after 11 iterationsBackcast: 1919:01 1920:01
Variable Coefficient Std. Error t-Statistic Prob.
C 2.38E-05 0.000196 0.121473 0.9033MA(1) 0.378388 0.029006 13.04509 0.0000
SMA(12) -0.805799 0.016884 -47.72483 0.0000
R-squared 0.522387 Mean dependent var -0.000108Adjusted R-squared 0.521437 S.D. dependent var 0.031842S.E. of regression 0.022028 Akaike info criterion -4.790041Sum squared resid 0.488142 Schwarz criterion -4.775422Log likelihood 2419.576 F-statistic 550.1535Durbin-Watson stat 1.839223 Prob(F-statistic) 0.000000
Inverted MA Roots .98 .85+.49i .85 -.49i .49+.85i .49 -.85i .00 -.98i -.00+.98i -.38 -.49 -.85i -.49+.85i -.85 -.49i -.85+.49i -.98
17
Basic structural model
A structural time series model for a quarterly
time series can look like
yt = µt + st + wt, wt ∼ NID(0, σ2w) (18)
(1− L)2µt = ut, ut ∼ NID(0, σ2u) (19)
(1 + L + L2 + L3)st = vt, vt ∼ NID(0, σ2v )
(20)
where the error processes wt, ut and vt are also
mutually independent, and where NID denotes
normally and independently distributed.
18
These three equations together imply that yt
can be described by
(1− L)(1− L4)yt = ζt (21)
where ζt is a moving average process of order
5 [MA(5)].
The autocovariances γk, k = 0,1,2, . . ., for ζt
are
γ0 = 4σ2u + 6σ2
v + 4σ2w (22)
γ1 = 3σ2u − 4σ2
v − 2σ2w (23)
γ2 = 2σ2u + σ2
v (24)
γ3 = σ2u + σ2
w (25)
γ4 = −2σ2w (26)
γ5 = σ2w (27)
γj = 0 for j = 6,7, . . . . (28)
19
Conclusion for lecture 2
There are various ways one can describe a time
series (and use that description for forecasting)
with constant or changing seasonal variation.
The next lecture will add even more to this
list.
To make a choice, one needs to test models
against each other. A key distinguishing fea-
ture of the models is the amount of ”random
walk” behavior, or better, the amount of unit
roots (seasonal and non-seasonal). Part of lec-
ture 3 will be devoted to the choice process.
An informal look at graphs and autocorrelation
functions is not sufficient. One needs to make
a decision based on formal tests on parameter
values in auxiliary regressions.
20
Forecasting
Seasonal Time Series
3. Advanced Models
Philip Hans Franses
Econometric Institute
Erasmus University Rotterdam
SMU and NUS, Singapore, April-May 2004
1
3. Advanced Models
• Seasonal unit roots
• Seasonal cointegration
• Periodic models
• Unit roots in periodic time series
• Periodic cointegration
3
A time series variable has a unit root if the au-
toregressive polynomial (of the autoregressive
model that best describes this variable), con-
tains the component 1 − L, and the moving-
average part does not, where L denotes the
familiar lag operator defined by Lkyt = yt−k,
for k = . . . ,−2,−1,0,1,2, . . . .
For example, the model yt = yt−1 + εt has a
first-order autoregressive polynomial 1−L, as it
can be written as (1−L)yt = εt, and hence data
that can be described by this model, which
is coined the random walk model, are said to
have a unit root. The same holds of course for
the model yt = µ + yt−1 + εt, which is called a
random walk with drift.
4
Solving this last model to the first observation,
that is,
yt = y0 + µt + εt + εt−1 + · · ·+ ε1 (1)
shows that such data display a trend. Due to
the summation of the error terms, it is possible
that data diverge from the overall trend µt for
a long time, and hence at first sight one would
conclude from a graph that there are all kinds
of temporary trends. Therefore, such data are
sometimes said to have a stochastic trend.
The unit roots in seasonal data, which can be
associated with changing seasonality, are the
so-called seasonal unit roots, see Hylleberg et
al (JofE, 1990).
5
For quarterly data, these roots are −1, i, and
−i. For example, data generated from the
model yt = −yt−1 + εt would display season-
ality, but if one were to make graphs with the
split seasonals, then one could observe that the
quarterly data within a year shift places quite
frequently.
Similar observations hold for the model yt =
−yt−2+εt, which can be written as (1+L2)yt =
εt, where the autoregressive polynomial 1+L2
corresponds to the seasonal unit roots i and −i,
as these two values solve the equation 1+z2 =
0. Hence, when a model for yt contains an au-
toregressive polynomial with roots −1 and/or
i, −i, the data are said to have seasonal unit
roots.
6
Testing for seasonal unit roots
The HEGY method for quarterly data amounts
to a regression of ∆4yt on deterministic terms
like an intercept, seasonal dummies, a trend
and seasonal trends on (1 + L + L2 + L3)yt−1,
(−1+ L−L2 + L3)yt−1, −(1+ L2)yt−1, −(1+
L2)yt−2, and on lags of ∆4yt.
The t1-test is for the significance of the pa-
rameter for (1 + L + L2 + L3)yt−1, the t2-test
for (−1 + L − L2 + L3)yt−1 and the joint F34
test for −(1+L2)yt−1 and −(1+L2)yt−2 . An
insignificant test value indicates the presence
of the associated root(s), which are 1, −1, and
the pair i, −i, respectively. Asymptotic theory
for the tests is developed in various studies.
7
The preferable approach is where alternating
dummy variables D1,t−D2,t+D3,t−D4,t, D1,t−D3,t, and D2,t − D4,t are included. Under the
null hypothesis that there is a unit root −1,
the parameter for the first alternating dummy
should also be zero. These joint tests extend
the work of Dickey and Fuller (1981).
When models are created for panels of time se-
ries or for multivariate series, these restrictions
on the deterministics (based on the sine-cosine
notation) are very important.
8
Seasonal cointegration
In case two or more seasonally observed time
series have seasonal unit roots, one may be
interested in testing for common seasonal unit
roots, that is, in testing for seasonal cointegra-
tion. If these series have such roots in com-
mon, they will have common changing sea-
sonal patterns. Engle et al (1993) propose a
method.
Suppose that two time series y1,t and y2,t are
to be differenced using the ∆4 filter. When
y1,t and y2,t have a common non-seasonal unit
root, then the series ut defined by
ut = (1+L+L2+L3)y1,t−α1(1+L+L2+L3)y2,t
(2)
does not need the (1 − L) filter to become
stationary.
9
Seasonal cointegration at the annual frequency
π, corresponding to unit root −1, implies that
vt = (1−L+L2−L3)y1,t−α2(1−L+L2−L3)y2,t
(3)
does not need the (1 + L) differencing filter.
Seasonal cointegration at the annual frequency
π/2, corresponding to the unit roots ±i, means
that
wt = (1− L2)y1,t (4)
− α3(1− L2)y2,t − α4(1− L2)y1,t−1
−α5(1− L2)y2,t−1 (5)
does not have the unit roots ±i.
10
In case all three ut, vt and wt series do not
have the relevant unit roots, the first equation
of a simple version of a seasonal cointegration
model is
∆4y1,t = γ1ut−1+γ2vt−1+γ3wt−2+γ4wt−3+ε1,t,
(6)
where γ1 to γ4 are error correction parameters.
The test method proposed in EGHL is a two-
step method, similar to the Engle-Granger ap-
proach to non-seasonal time series.
11
Seasonal cointegration in a multivariate time
series Yt can also be analyzed using an ex-
tension of the Johansen approach (see Lee
(1992), Johansen and Schaumburg (1999), Kunst
and Franses (OBES 1999). It amounts to test-
ing the ranks of matrices that correspond to
variables which are transformed using the fil-
ters to remove the roots 1, -1 or ± i.
More precise, consider the (m× 1) vector pro-
cess Yt, and assume that it can be described
by the VAR(p) process
Yt = ΘDt + Φ1Yt−1 + · · ·+ ΦpYt−p + et, (7)
where Dt is the (4 × 1) vector process Dt =
(D1,t, D2,t, D3,t, D4,t)′ containing the seasonal
dummies, and where Θ is an (m×4) parameter
matrix.
12
Similar to the Johansen approach and condi-
tional on the assumption that p > 4, the model
can be rewritten as
∆4Yt = ΘDt + Π1Y1,t−1 (8)
+ Π2Y2,t−1 + Π3Y3,t−2 + Π4Y3,t−1 (9)
+ Γ1∆4Yt−1 + · · ·+ Γp−4∆4Yt−(p−4) + et,
where
Y1,t = (1 + L + L2 + L3)Yt
Y2,t = (1− L + L2 − L3)Yt
Y3,t = (1− L2)Yt.
This is a multivariate extension of the univari-
ate HEGY model. The ranks of the matrices
Π1, Π2, Π3 and Π4 determine the number of
cointegration relations at a certain frequency.
13
Periodic models
Consider a univariate time series yt, which is
observed quarterly for N years, that is, t =
1,2, . . . , n. It is assumed that n = 4N . A peri-
odic autoregressive model of order p [PAR(p)]
can be written as
yt = µs + φ1syt−1 + · · ·+ φpsyt−p + εt, (10)
or
φp,s(L)yt = µs + εt (11)
with
φp,s(L) = 1− φ1sL− · · · − φpsLp, (12)
where µs is a seasonally-varying intercept term.
14
The φ1s, . . . , φps are autoregressive parameters
up to order ps which may vary with the season
s, where s = 1,2,3,4. εt is assumed to be
a standard white noise process with constant
variance σ2. This assumption may be relaxed
by allowing εt to have seasonal variance σ2s .
As some φis, i = 1,2, . . . , p, can take zero val-
ues, the order p is the maximum of all ps, where
ps denotes the AR order per season s.
15
Multivariate representation
In general, the PAR(p) process can be rewrit-
ten as an AR(P ) model for the (4 × 1) vec-
tor process YT = (Y1,T , Y2,T , Y3,T , Y4,T )′, T =
1,2, . . . , N , where Ys,T is the observation of yt
in season s of year T , s = 1,2,3,4. The model
is then
Φ0YT = µ+Φ1YT−1+· · ·+ΦPYT−P +εT , (13)
or
Φ(L)YT = µ + εT (14)
with
Φ(L) = Φ0 −Φ1L− · · · −ΦPLP , (15)
16
µ = (µ1, µ2, µ3, µ4)′, εT = (ε1,T , ε2,T , ε3,T , ε4,T )′,
and εs,T is the observation on the error process
εt in season s of year T . Note that the lag op-
erator L applies to data at frequencies t and to
T , that is, Lyt = yt−1 and LYT = YT−1. The
Φ0,Φ1, . . . ,ΦP are 4 × 4 parameter matrices
with elements
Φ0[i, j] =
1 i = j,0 j > i,−φi−j,i i < j,
Φk[i, j] = φi+4k−j,i,
(16)
for i = 1,2,3,4, j = 1,2,3,4, and k = 1,2, . . . , P .
17
For the model order P it holds that P = 1 +
[(p−1)/4], where [ · ] is again the integer func-
tion. Hence, when p is less than or equal to 4,
the value of P is only 1.
As Φ0 is a lower triangular matrix, model (13)
is a recursive model. This means that the ob-
servations in the fourth quarter, that is, Y4,T ,
depend on the preceding observations in the
same year, that is Y3,T , Y2,T , and Y1,T , and
observations in past years. Similarly, Y3,T de-
pends on Y2,T and Y1,T , and Y2,T on Y1,T and
observations in past years.
18
As an example, consider the PAR(2) process
yt = φ1syt−1 + φ2syt−2 + εt, (17)
which can be written as
Φ0YT = Φ1YT−1 + εT , (18)
with
Φ0 =
1 0 0 0−φ12 1 0 0−φ23 −φ13 1 0
0 −φ24 −φ14 1
and
Φ1 =
0 0 φ21 φ110 0 0 φ220 0 0 00 0 0 0
. (19)
19
A useful representation is based on the pos-
sibility of decomposing a non-periodic AR(p)
polynomial as (1−α1L)(1−α2L) · · · (1−αpL).
Note that this can only be done when the so-
lutions to the characteristic equation for this
AR(p) polynomial are all real-valued. Similar
results hold for the multivariate representation
of a PAR(p) process, and it can be useful to
rewrite (13) as
P∏
i=1
Ξi(L)YT = µ + εT , (20)
where the Ξi(L) are 4 × 4 matrices with ele-
ments which are polynomials in L.
20
A simple example is the PAR(2) process
Ξ1(L)Ξ2(L)YT = εT , (21)
with
Ξ1(L) =
1 0 0 −β1L−β2 1 0 00 −β3 1 00 0 −β4 1
,
Ξ2(L) =
1 0 0 −α1L−α2 1 0 00 −α3 1 00 0 −α4 1
, (22)
21
The PAR(2) can be written as
(1− βsL)(1− αsL)yt = µs + εt. (23)
This expression equals
yt−αsyt−1 = µs+βs(yt−1−αs−1yt−2)+εt, (24)
as, and this is quite important, the lag operator
L also operates on αs, that is, Lαs = αs−1 for
all s = 1,2,3,4 and with α0 = α4.
22
The characteristic equation is
|Ξ1(z)Ξ2(z)| = 0, (25)
which is equivalent to
(1− β1β2β3β4z)(1− α1α2α3α4z) = 0, (26)
and hence the PAR(2) model has one unit root
when either β1β2β3β4 = 1 or α1α2α3α4 = 1,
and has at most two unit roots when both
products equal unity. Obviously, the maximum
number of unity solutions to the characteristic
equation of a PAR(p) process is equal to p.
23
The analogy of a univariate PAR process with
a multivariate time series process can also be
exploited to derive explicit formulae for one-
and multi-step ahead forecasting. It should
be noted that the one-step ahead forecasts
concern one-year ahead forecasts for all four
Ys,T series. For example, for the model YT =
Φ−10 Φ1YT−1+ωT , where ωT = Φ−1
0 εT , the fore-
cast for N + 1 is YN+1 = Φ−10 Φ1YN .
24
Fitting non-periodic models to otherwise peri-
odic data increases the lag order. For example,
a PAR(1) model can be written as
yt = αs+3αs+2αs+1αsyt−4 + εt + αs+3εt−1
+ αs+3αs+2εt−2 + αs+3αs+2αs+1εt−3. (27)
As αs+3αs+2αs+1αs is equal for all seasons,
the AR parameter at lag 4 in a non-periodic
model is truly non-periodic. The MA part of
this model is of order 3. If if one estimates
a non-periodic MA model for these data, the
MA parameter estimates will attain some kind
of an average value of αs+3, αs+3αs+2, and
αs+3αs+2αs+1 across the seasons.
25
In other words, one might end up considering
an ARMA(4,3) model for PAR(1) data. And,
if one decides not to include an MA part in the
model, one usually needs to increase the order
of the autoregression so as to whiten the er-
rors. This suggests that higher-order AR mod-
els might fit to low-order periodic data. Note
that when αs+3αs+2αs+1αs = 1, this implies a
higher-order AR model for the ∆4 transformed
time series.
Hence, there seems to be a trade-off between
seasonality in parameters and short lags, and
no seasonality in parameters and longer lags.
26
Some further issues:
• Testing for unit roots
• Periodic cointegration
• Forecasting results (Lof, Franses IJF (2002)
analyze periodic and seasonal cointegra-
tion models for bivariate quarterly observed
time series in an empirical forecasting study,
as well as a VAR model in first differences,
with and without cointegration restrictions,
and a VAR model in annual differences.
The VAR model in first differences without
cointegration is best if one-step ahead fore-
casts are considered. For longer forecast
horizons, the VAR model in annual differ-
ences is better. When comparing periodic
versus seasonal cointegration models, the
seasonal cointegration models tend to yield
better forecasts. Finally, there is no clear
indication that multiple equations methods
improve on single equation methods.)
27
Conclusion for lecture 3
Tests for periodic variation in the parameters
and for unit roots allows one to make a choice
between the various models for seasonality.
Models and methods for data with frequencies
higher than 12 can become difficult to use in
practice.
Also, neglected structural breaks can inappro-
priately suggest seasonal unit roots and also
periodicity, see next lecture.
28
Forecasting
Seasonal Time Series
4. Further topics
Philip Hans Franses
Econometric Institute
Erasmus University Rotterdam
SMU and NUS, Singapore, April-May 2004
1
4. Further topics
• Seasonality in panels of time series
• Nonlinear models for seasonal time series
• Further work
3
Seasonality in panels of time series
Seasonal patterns are quite prominent is manyeconomic data and the search for common sea-sonal patterns can lead to a dramatic reduc-tion in the number of parameters, see Engleand Hylleberg (1996).
Before one can say something about (com-mon) deterministic seasonality, one first has todecide on the number of seasonal unit roots.The basic test regression to test for seasonalunit roots is
Φpi(L)∆4yi,t = µi,t+ρi,1S(L)yi,t−1+ρi,2A(L)yi,t−1
+ ρi,3∆2yi,t−1 + ρi,4∆2yi,t−2 + εt, (1)
where
µi,t = µi + α1,i cos(πt) + α2,i cos(πt
2)
+ α3,i cos(π(t− 1)
2) + δit, (2)
and where ∆k is the k−th order differencingfilter, S(L)yi,t = (1 + L + L2 + L3)yi,t andA(L)yi,t = −(1− L + L2 − L3)yi,t.
4
The model assumes that each series yi,t can
be described by a (pi + 4)−th order autore-
gression. Franses and Kunst (1999) show that
an appropriate test for a seasonal unit root at
the bi-annual frequency is now given by a joint
F−test for ρi,2 and α1,i. An appropriate test
for the two seasonal unit roots at the annual
frequency is then given by a joint F−test for
ρ3,i, ρ4,i, α2,i and α3,i.
Franses and Kunst (1999) consider these F -
tests in a model where the autoregressive pa-
rameters are pooled over the equations. It is
seen that the power of this panel test proce-
dure is rather large.
Additionally, they examine if, once one has
taken care off seasonal unit roots, two or more
series have the same deterministic fluctuations.
This can be done by testing for cross-equation
restrictions.5
Periodic GARCH
Periodic models might also be useful to model
financial time series. They can be used not
only to describe the so-called day-of-the-week
effects, but also to describe the differences in
volatility across the day of the week.
Bollerslev-Ghysels (JBES 1996) propose a pe-
riodic generalized autoregressive conditional het-
eroskedasticity (PGARCH) model.
6
A PAR(p)–PGARCH(1,1) model for a daily ob-
served financial time series yt, t = 1, . . . , n =
5N , can be represented by
xt = yt −5∑
s=1
µs +
p∑
i=1
φisyt−i
Ds,t
=√
htηt
with ηt ∼ N(0,1), (3)
and
ht =5∑
s=1
(ωs + ψsx2t−1)Ds,t + γht−1, (4)
where the xt denotes the residual of the PAR
model for yt, and where Ds,t denote the usual
seasonal dummies but now for the days of the
week, that is, s = 1,2,3,4,5.
7
In order to investigate the properties of the
conditional variance model, it is useful to de-
fine zt = x2t − ht, and to write it as a PARMA
process
x2t =
5∑
s=1
(ωs+(ψs+γ)x2t−1)Ds,t+zt−γzt−1. (5)
This ARMA process for x2t contains time-varying
parameters ψs + γ and hence strictly speaking,
it is not a stationary process.
To investigate the stationarity properties of x2t ,
it is more convenient to write (5) in a param-
eter time-invariant VD representation.
8
Models of seasonality and nonlinearity
A change in the deterministic trend properties
of a time series yt is easily mistaken for the
presence of a unit root.
In a similar vein, if a change in the deter-
ministic seasonal pattern goes undetected, one
might well end up imposing seasonal unit roots,
see Ghysels (JBES, 1994), Smith and Otero
(EL, 1997), Franses, Hoek and Paap (JoE,
1997) and Franses and Vogelsang (RESTAT,
1998)
Changes in deterministic seasonal patterns usu-
ally are modelled by means of one-time abrupt
and discrete changes. However, when sea-
sonal patterns shift due to changes in technol-
ogy, institutions and tastes, for example, these
changes may materialize only gradually.
9
This suggests that a plausible description of
time-varying seasonal patterns is
φ(L)∆1yt =4∑
s=1
δ1,sDs,t(1−G(t; γ, c))
+4∑
s=1
δ2,sDs,tG(t; γ, c) + εt, (6)
where G(t; γ, c) is the logistic function
G(st; γ, c) =1
1 + exp{−γ(st − c)}, γ > 0.
(7)
10
As st increases, the logistic function changes
monotonically from 0 to 1, with the change
being symmetric around the location parame-
ter c, as G(c − z; γ, c) = 1 − G(c + z; γ, c) for
all z. The slope parameter γ determines the
smoothness of the change. As γ → ∞, the
logistic function G(st; γ, c) approaches the in-
dicator function I[st > c], whereas if γ → 0,
G(st; γ, c) → 0.5 for all values of st.
Hence, by taking st = t, the model takes an
“intermediate” position in between determin-
istic seasonality and stochastic trend seasonal-
ity.
11
A nonlinear model with smoothly changing de-
terministic seasonality is
∆1yt = [µ1(1−G(∆4yt−d; γ2, c2))+
µ2G(∆4yt−d; γ2, c2)](1−G(t; γ1, c1))
+ [µ3(1−G(∆4yt−d; γ2, c2))+
µ4G(∆4yt−d; γ2, c2)]G(t; γ1, c1)
+3∑
s=1
δ1,sD∗s,t(1−G(t; γ1, c1))
+3∑
s=1
δ2,sD∗s,tG(t; γ1, c1)
+ (φ1,1∆1yt−1 + · · ·+ φ1,p∆1yt−p)
(1−G(∆4yt−d; γ2, c2))
+ (φ2,1∆1yt−1 + · · ·+ φ2,p∆1yt−p)
G(∆4yt−d; γ2, c2) + εt, (8)
The seasonal pattern is allowed to vary over
time, but it is assumed to be identical in ex-
pansions and recessions at all times.
12
Finally, a time-varying smooth transition model
reads as
∆1yt = ([µ1 +3∑
s=1
δ1,sD∗s,t + φ1,1∆1yt−1
+ · · ·+ φ1,p∆1yt−p](1−G(∆4yt−d; γ2, c2))
+ [µ2 +3∑
s=1
δ2,sD∗s,t + φ2,1∆1yt−1
+· · ·+φ2,p∆1yt−p]G(∆4yt−d; γ2, c2))(1−G(t; γ1, c1))
([µ3 +3∑
s=1
δ3,sD∗s,t + φ3,1∆1yt−1
+ · · ·+ φ3,p∆1yt−p](1−G(∆4yt−d; γ2, c2))
+ [µ4 +3∑
s=1
δ4,sD∗s,t + φ4,1∆1yt−1
+· · ·+φ4,p∆1yt−p]G(∆4yt−d; γ2, c2))G(t; γ1, c1)+εt.
(9)
13
Franses and van Dijk (2004) examine the fore-
casting performance of various models for sea-
sonality and nonlinearity for quarterly industrial
production series of 18 OECD countries. They
find that the accuracy of point forecasts varies
widely across series, across forecast horizons
and across seasons. However, in general, linear
models with fairly simple descriptions of sea-
sonality outperform at short forecast horizons,
whereas nonlinear models with more elaborate
seasonal components dominate at longer hori-
zons. Simpler models are also preferable for in-
terval and density forecasts at short horizons.
Finally, none of the models is found to render
efficient forecasts and hence, forecast combi-
nation is worthwhile.
14
Conclusion for lecture 4
Economic time series (can) have trends, sea-
sonality, nonlinearity, outliers and conditional
volatility.
To fully capture all these features is not an
easy task. Various features may be related.
15
Conclusion for lectures 1-4
Forecasting studies show that model specifi-
cation efforts pay off in terms of performance.
Simple models for seasonally differenced data
forecast well for one or a few steps ahead.
For longer horizons, more involved models are
much better.
Remaining research issues:
• How to achieve parsimony?
• How to date turning points in case trends,
cycles and seasonality are related?
• How to interpret seasonally adjusted data
when trends, cycles and seasonality are re-
lated?
16
Further-reading.bib (April 7, 2004)
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FM89 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .M. Faig.Seasonal fluctuations and the demand for money.Quarterly Journal of Economics, 99:847–861, 1989.
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FP91 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .P. H. Franses.Model Selection and Seasonality in Time Series.PhD thesis, Tinbergen Institute, Amsterdam, 1991.
FP91c . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .P. H. Franses.A multivariate approach to modeling univariate seasonal time series.Econometric Institute Report 9101, Erasmus University Rotterdam, 1991.
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FP93 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .P. H. Franses.Periodically integrated subset autoregressions for Dutch industrial production and money stock.Journal of Forecasting, 12:601–613, 1993.
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FP96 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .P. H. Franses.Periodicity and Stochastic Trends in Economic Time Series.Oxford University Press, Oxford, 1996.
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FT00 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .P. H. Franses and A. M. R. Taylor.Determining the order of differencing in seasonal time series processes.Econometrics Journal, 3:250–264, 2000.
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FR86 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .K. R. French and R. Roll.Stock return variances: The arrival of information and the reaction of traders.Journal of Financial Economics, 17:5–26, 1986.
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GM78 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .M. Gersovitz and J. G. MacKinnon.Seasonality in regression: An application of smoothness priors.Journal of the American Statistical Association, 73:264–273, 1978.
GE88 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. Ghysels.A study towards a dynamic theory of seasonality for economic time series.Journal of the American Statistical Association, 83:168–172, 1988.
GE90 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. Ghysels.Unit root tests and the statistical pitfalls of seasonal adjustment: The case of US postwar real gross nationalproduct.Journal of Business and Economic Statistics, 8:145–152, 1990.
GE93 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. Ghysels.On scoring asymmetric periodic probability models of turning point forecasts.Journal of Forecasting, 12:227–238, 1993.
GE94 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. Ghysels.On the periodic structure of the business cycle.Journal of Business and Economic Statistics, 12:289–298, 1994.
GE00 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. Ghysels.Time-series model with periodic stochastic regime switching – part I: Theory.Macroeconomic Dynamics, 4:467–486, 2000.
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GE61 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .E. G. Gladyshev.Periodically correlated random sequences.Soviet Mathematics, 2:385–388, 1961.
GN70 . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .D. M. Grether and M. Nerlove.Some properties of optimal seasonal adjustment.Econometrica, 38:682–703, 1970.
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