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JMLR: Workshop and Conference Proceedings 12 (2011) 95–118Causality in Time Series
Causal Search in Structural Vector Autoregressive Models
Alessio Moneta
Max Planck Institute of Economics
Jena, Germany
moneta@econ.mpg.de
Nadine Chlaß
Friedrich Schiller University of Jena, Germany
nadine.chlass@uni-jena.de
Doris Entner
Helsinki Institute for Information Technology, Finland
doris.entner@cs.helsinki.fi
Patrik Hoyer
Helsinki Institute for Information Technology, Finland
patrk.hoyer@helsinki.fi
Editor: Florin Popescu and Isabelle Guyon
Abstract
This paper reviews a class of methods to perform causal inference in the framework of a
structural vector autoregressive model. We consider three different settings. In the first
setting the underlying system is linear with normal disturbances and the structural model is
identified by exploiting the information incorporated in the partial correlations of the esti-
mated residuals. Zero partial correlations are used as input of a search algorithm formalized
via graphical causal models. In the second, semi-parametric, setting the underlying system
is linear with non-Gaussian disturbances. In this case the structural vector autoregressive
model is identified through a search procedure based on independent component analysis.
Finally, we explore the possibility of causal search in a nonparametric setting by studying
the performance of conditional independence tests based on kernel density estimations.
Keywords: Causal inference, econometric time series, SVAR, graphical causal models,
independent component analysis, conditional independence tests
1. Introduction
1.1. Causal inference in econometrics
Applied economic research is pervaded by questions about causes and effects. For exam-
ple, what is the effect of a monetary policy intervention? Is energy consumption causing
growth or the other way around? Or does causality run in both directions? Are economic
fluctuations mainly caused by monetary, productivity, or demand shocks? Does foreign
aid improve living standards in poor countries? Does firms’ expenditure in R&D causally
influence their profits? Are recent rises in oil prices in part caused by speculation? These
are seemingly heterogeneous questions, but they all require some knowledge of the causal
process by which variables came to take the values we observe.
A traditional approach to address such questions hinges on the explicit use of a priori
economic theory. The gist of this approach is to partition a causal process in a determinis-
tic, and a random part and to articulate the deterministic part such as to reflect the causal
c ? 2011 A. Moneta, N. Chlaß, D. Entner & P. Hoyer.
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Moneta Chlaß Entner Hoyer
dependencies dictated by economic theory. If the formulation of the deterministic part is ac-
curate and reliable enough, the random part is expected to display properties that can easily
be analyzed by standard statistical tools. The touchstone of this approach is represented by
the work of Haavelmo (1944), which inspired the research program subsequently pursued
by the Cowles Commission (Koopmans, 1950; Hood and Koopmans, 1953). Therein, the
causal process is formalized by means of a structural equation model, that is, a system of
equations with endogenous variables, exogenous variables, and error terms, first developed
by Wright (1921). Its coefficients were given a causal interpretation (Pearl, 2000).
This approach has been strongly criticized in the 1970s for being ineffective in both policy
evaluation and forecasting. Lucas (1976) pointed out that the economic theory included
in the SEM fails to take economic agents’ (rational) motivations and expectations into
consideration. Agents, according to Lucas, are able to anticipate policy intervention and
act contrary to the prediction derived from the structural equation model, since the model
usually ignores such anticipations. Sims (1980) puts forth another critique which runs
parallel to Lucas’ one. It explicitly addresses the status of exogeneity which the Cowles
Commission approach attributes (arbitrarily, according to Sims) to some variables such that
the structural model can be identified. Sims argues that theory is not a reliable source for
deeming a variable as exogenous. More generally, the Cowles Commission approach with its
strong a priori commitment to theory, risks falling into a vicious circle: if causal information
(even if only about direction) can exclusively be derived from background theory, how do
we obtain an empirically justified theory? (Cfr. Hoover, 2006, p.75).
An alternative approach has been pursued since Wiener (1956) and Granger’s (1969)
work. It aims at inferring causal relations directly from the statistical properties of the
data relying only to a minimal extent on background knowledge. Granger (1980) proposes
a probabilistic concept of causality, similar to Suppes (1970). Granger defines causality in
terms of the incremental predictability (at horizon one) of a time series variable {Yt} (given
the present and past values of {Yt} and of a set {Zt} of possible relevant variables) when
another time series variable {Xt} (in its present and past values) is not omitted. More
formally:
{Xt} Granger-causes {Yt} if P(Yt+1|Xt,Xt−1,...,Yt,Yt−1,...,Zt,Zt−1,...) ?=
P(Yt+1|Yt,Yt−1,...,Zt,Zt−1,...)
As pointed out by Florens and Mouchart (1982), testing the hypothesis of Granger non-
causality corresponds to testing conditional independence. Given lags p, {Xt} does not
Granger cause {Yt}, if
Yt+1⊥ ⊥ (Xt,Xt−1,...,Xt−p) | (Yt,Yt−1,...,Yt−p,Zt,Zt−1,...,Zt−p)
To test Granger noncausality, researchers often specify linear vector autoregressive (VAR)
models:
Yt= A1Yt−1+ ... + ApYt−p+ ut,
in which Yt is a k × 1 vector of time series variables (Y1,t,...,Yk,t)?, where ()?is the
transpose, the Aj(j = 1,...,p) are k × k coefficient matrices, and utis the k × 1 vector
of random disturbances. In this framework, testing the hypothesis that {Yi,t} does not
Granger-cause {Yj,t}, reduces to test whether the (j,i) entries of the matrices A1,...,Ap
(1)
(2)
(3)
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are vanishing simultaneously. Granger noncausality tests have been extended to nonlinear
settings by Baek and Brock (1992), Hiemstra and Jones (1994), and Su and White (2008),
using nonparametric tests of conditional independence (more on this topic in section 4).
The concept of Granger causality has been criticized for failing to capture ‘structural
causality’ (Hoover, 2008). Suppose one finds that a variable A Granger-causes another
variable B. This does not necessarily imply that an economic mechanism exists by which
A can be manipulated to affect B. The existence of such a mechanism in turn does not
necessarily imply Granger causality either (for a discussion see Hoover 2001, pp. 150-155).
Indeed, the analysis of Granger causality is based on coefficients of reduced-form models,
like those incorporated in equation (3), which are unlikely to reliably represent actual eco-
nomic mechanisms. For instance, in equation (3) the simultaneous causal structure is not
modeled in order to facilitate estimation. (However, note that Eichler (2007) and White
and Lu (2010) have recently developed and formalized richer structural frameworks in which
Granger causality can be fruitfully analyzed.)
1.2. The SVAR framework
Structural vector autoregressive (SVAR) models constitute a middle way between the Cowles
Commission approach and the Granger-causality approach. SVAR models aim at recovering
the concept of structural causality, but eschew at the same time the strong ‘apriorism’ of
the Cowles Commission approach. The idea is, like in the Cowles Commission approach,
to articulate an unobserved structural model, formalized as a dynamic generative model:
at each time unit the system is affected by unobserved innovation terms, by which, once
filtered by the model, the variables come to take the values we observe. But, differently
from the Cowles Commission approach, and similarly to the Granger-VAR model, the data
generating process is generally enough articulated so that time series variables are not dis-
tinguished a priori between exogenous and endogenous. A linear SVAR model is in principle
a VAR model ‘augmented’ by the contemporaneous structure:
Γ0Yt= Γ1Yt−1+ ... + ΓpYt−p+ εt. (4)
This is easily obtained by pre-multiplying each side of the VAR model
Yt= A1Yt−1+ ... + ApYt−p+ ut, (5)
by a matrix Γ0so that Γi= Γ0Ai, for i = 1,...,k and εt= Γ0ut. Note, however, that
not any matrix Γ0will be suitable. The appropriate Γ0will be that matrix corresponding
to the ‘right’ rotation of the VAR model, that is the rotation compatible both with the
contemporaneous causal structure of the variable and the structure of the innovation term.
Let us consider a matrix B0= I−Γ0. If the system is normalized such that the matrix Γ0has
all the elements of the principal diagonal equal to one (which can be done straightforwardly),
the diagonal elements of B0will be equal to zero. We can write:
Yt= B0Yt+ Γ1Yt−1+ ... + ΓpYt−p+ εt
(6)
from which we see that B0(and thus Γ0) determines in which form the values of a variable
Yi,twill be dependent on the contemporaneous value of another variable Yj,t. The ‘right’
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rotation will also be the one which makes εta vector of authentic innovation terms, which
are expected to be independent (not only over time, but also contemporaneously) sources
or shocks.
In the literature, different methods have been proposed to identify the SVAR model
(4) on the basis of the estimation of the VAR model (5). Notice that there are more
unobserved parameters in (4), whose number amounts to k2(p + 1), than parameters that
can be estimated from (5), which are k2p + k(k + 1)/2, so one has to impose at least
k(k − 1)/2 restrictions on the system. One solution to this problem is to get a rotation
of (5) such that the covariance matrix of the SVAR residuals Σε is diagonal, using the
Cholesky factorization of the estimated residuals Σu. That is, let P be the lower-triangular
Cholesky factorization of Σu (i.e. Σu = PP?), let D be a k × k diagonal matrix with
the same diagonal as P, and let Γ0= DP−1. By pre-multiplying (5) by Γ0, it turns out
that Σε= E[Γ0utu?tΓ?
P changes if the ordering of the variables (Y1t,...,Ykt)?in Ytand, consequently, the order
of residuals in Σu, changes. Since researchers who estimate a SVAR are often exclusively
interested on tracking down the effect of a structural shock εiton the variables Y1,t,...,Yk,t
over time (impulse response functions), Sims (1981) suggested investigating to what extent
the impulse response functions remain robust under changes of the order of variables.
Popular alternatives to the Cholesky identification scheme are based either on the use
of a priori, theory-based, restrictions or on the use of long-run restrictions. The former
solution consists in imposing economically plausible constraints on the contemporaneous
interactions among variables (Blanchard and Watson, 1986; Bernanke, 1986) and has the
drawback of ultimately depending on the a priori reliability of economic theory, similarly
to the Cowles Commission approach. The second solution is based on the assumptions that
certain economic shocks have long-run effect to other variables, but do not influence in the
long-run the level of other variables (see Shapiro and Watson, 1988; Blanchard and Quah,
1989; King et al., 1991). This approach has been criticized as not being very reliable unless
strong a priori restrictions are imposed (see Faust and Leeper, 1997).
In the rest of the paper, we first present a method, based on the graphical causal model
framework, to identify the SVAR (section 2). This method is based on conditional inde-
pendence tests among the estimated residuals of the VAR estimated model. Such tests
rely on the assumption that the shocks affecting the model are Gaussian. We then relax
the Gaussianity assumption and present a method to identify the SVAR model based on
independent component analysis (section 3). Here the main assumption is that shocks are
non-Gaussian and independent. Finally (section 4), we explore the possibility of extending
the framework for causal inference to a nonparametric setting. In section 5 we wrap up the
discussion and conclude by formulating some open questions.
0] = DD?, which is diagonal. A problem with this method is that
2. SVAR identification via graphical causal models
2.1. Background
A data-driven approach to identify the structural VAR is based on the analysis of the es-
timated residuals ˆ ut. Notice that when a basic VAR model is estimated (equation 3), the
information about contemporaneous causal dependence is incorporated exclusively in the
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residuals (being not modeled among the variables). Graphical causal models, as originally
developed by Pearl (2000) and Spirtes et al. (2000), represent an efficient method to recover,
at least in part, the contemporaneous causal structure moving from the analysis of the con-
ditional independencies among the estimated residuals. Once the contemporaneous causal
structure is recovered, the estimation of the lagged autoregressive coefficients permits us to
identify the complete SVAR model.
This approach was initiated by Swanson and Granger (1997), who proposed to test
whether a particular causal order of the VAR is in accord with the data by testing all
the partial correlations of order one among error terms and checking whether some partial
correlations are vanishing. Reale and Wilson (2001), Bessler and Lee (2002), Demiralp and
Hoover (2003), and Moneta (2008) extended the approach by using the partial correlations
of the VAR residuals as input to graphical causal model search algorithms.
In graphical causal models, the structural model is represented as a causal graph (a
Directed Acyclic Graph if the presence of causal loops is excluded), in which each node
represents a random variable and each edge a causal dependence. Furthermore, a set of
assumptions or ‘rules of inference’ are formulated, which regulate the relationship between
causal and probabilistic dependencies: the causal Markov and the faithfulness conditions
(Spirtes et al., 2000). The former restricts the joint probability distribution of modeled
variables: each variable is independent of its graphical non-descendants conditional on its
graphical parents. The latter makes causal discovery possible: all of the conditional inde-
pendence relations among the modeled variables follow from the causal Markov condition.
Thus, for example, if the causal structure is represented as Y1t→ Y2t→ Yt,3, it follows from
the Markov condition that Y1,t⊥ ⊥ Y3,t|Y2,t. If, on the other hand, the only (conditional)
independence relation among Y1,t,Y2,t,Y3,t is Y1,t ⊥ ⊥ Y3,t, it follows from the faithfulness
condition that Y1,t→ Y3,t<— Y2,t.
Constraint-based algorithms for causal discovery, like for instance, PC, SGS, FCI (Spirtes
et al. 2000), or CCD (Richardson and Spirtes 1999), use tests of conditional independence
to constrain the possible causal relationships among the model variables. The first step of
the algorithm typically involves the formation of a complete undirected graph among the
variables so that they are all connected by an undirected edge. In a second step, condi-
tional independence relations (or d-separations, which are the graphical characterization of
conditional independence) are merely used to erase edges and, in further steps, to direct
edges. The output of such algorithms are not necessarily one single graph, but a class of
Markov equivalent graphs.
There is nothing neither in the Markov or faithfulness condition, nor in the constraint-
based algorithms that limits them to linear and Gaussian settings. Graphical causal models
do not require per se any a priori specification of the functional dependence between vari-
ables. However, in applications of graphical models to SVAR, conditional independence is
ascertained by testing vanishing partial correlations (Swanson and Granger, 1997; Bessler
and Lee, 2002; Demiralp and Hoover, 2003; Moneta, 2008). Since normal distribution guar-
antees the equivalence between zero partial correlation and conditional independence, these
applications deal de facto with linear and Gaussian processes.
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2.2. Testing residuals zero partial correlations
There are alternative methods to test zero partial correlations among the error terms ˆ ut=
(u1t,...,ukt)?. Swanson and Granger (1997) use the partial correlation coefficient. That is,
in order to test, for instance, ρ(uit,ukt|ujt) = 0, they use the standard t statistics from a
least square regression of the model:
uit= αjujt+ αkukt+ εit,(7)
on the basis that αk= 0 ⇔ ρ(uit,ukt|ujt) = 0. Since Swanson and Granger (1997) impose
the partial correlation constraints looking only at the set of partial correlations of order
one (that is conditioned on only one variable), in order to run their tests they consider
regression equations with only two regressors, as in equation (7).
Bessler and Lee (2002) and Demiralp and Hoover (2003) use Fisher’s z that is incorporated
in the software TETRAD (Scheines et al., 1998):
z(ρXY.K,T) =1
2
?
T − |K| − 3 log
?|1 + ρXY.K|
|1 − ρXY.K|,
?
(8)
where |K| equals the number of variables in K and T the sample size. If the variables (for
instance X = uit, Y = ukt, K = (ujt,uht)) are normally distributed, we have that
z(ρXY.K,T) − z(ˆ ρXY.K,T) ∼ N(0,1) (9)
(see Spirtes et al., 2000, p.94).
A different approach, which takes into account the fact that correlations are obtained
from residuals of a regression, is proposed by Moneta (2008). In this case it is useful to
write the VAR model of equation (3) in a more compact form:
Yt= Π?Xt+ ut,(10)
where X?t= [Y?
has dimension (k×kp). In case of stable VAR process (see next subsection), the conditional
maximum likelihood estimate of Π for a sample of size T is given by
t−1, ...,Y?t−p], which has dimension (1 × kp) and Π?= [A1,...,Ap], which
ˆΠ?=
?T
t=1
?
YtX?
t
??T
?
t=1
XtX?
t
?−1
.
Moreover, the ith row ofˆΠ?is
ˆ π?
i=
?T
t=1
?
YitX?
t
??T
?
t=1
XtX?
t
?−1
,
which coincides with the estimated coefficient vector from an OLS regression of Yit on
Xt (Hamilton 1994: 293). The maximum likelihood estimate of the matrix of variance
and covariance among the error terms Σuturns out to beˆΣu= (1/T)?T
t=1ˆ utˆ u?t, where
ˆ ut= Yt−ˆΠ?Xt. Therefore, the maximum likelihood estimate of the covariance between uit
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and ujtis given by the (i,j) element ofˆΣu: ˆ σij= (1/T)?T
stacks the columns of a k × k matrix into a vector of length k2and vech, which vertically
stacks the elements of a k × k matrix on or below the principal diagonal into a vector of
length k(k + 1)/2. For example:
The process being stationary and the error terms Gaussian, it turns out that:
√T [vech(ˆΣu) − vech(Σu)]
where Ω = 2D+
matrix satisfying Dkvech(Ω) = vec(Ω), and ⊗ denotes the Kronecker product (see Hamilton
1994: 301). For example, for k = 2, we have,
ˆ σ22− σ22
Therefore, to test the null hypothesis that ρ(uit,ujt) = 0 from the VAR estimated residuals,
it is possible to use the Wald statistic:
t=1ˆ uitˆ ujt. Denoting by σijthe
(i,j) element of Σu, let us first define the following matrix transform operators: vec, which
vec
?
σ11σ12
σ21σ22
?
=
σ11
σ21
σ12
σ22
,
vech
?
σ11σ12
σ21σ22
?
=
σ11
σ21
σ22
.
d
−→ N(0, Ω),
k, Dkis the unique (k2×k(k+1)/2)
(11)
k(Σu⊗Σu)(D+
k)?, D+
k≡ (D?
kDk)−1D?
√T
ˆ σ11− σ11
ˆ σ12− σ12
d
−→ N
0
0
0
,
2σ2
2σ11σ12
2σ2
11
2σ11σ12
σ11σ22+ σ2
2σ12σ22
2σ2
2σ12σ22
2σ2
12
12
1222
T (ˆ σij)2
ˆ σiiˆ σjj + ˆ σ2
ij
≈ χ2(1).
The Wald statistic for testing vanishing partial correlations of any order is obtained by
applying the delta method, which suggests that if XTis a (r×1) sequence of vector-valued
random-variables and if [√T(X1T− θ1),...,√T(XrT− θr)]
are r real-valued functions of θ = (θ1,...,θr), hi : Rr→ R, defined and continuously
differentiable in a neighborhood ω of the parameter point θ and such that the matrix
B = ||∂hi/∂θj|| of partial derivatives is nonsingular in ω, then:
[√T[h1(XT) − h1(θ)],...,
(see Lehmann and Casella 1998: 61).
Thus, for k = 4, suppose one wants to test corr(u1t,u3t|u2t) = 0. First, notice that
ρ(u1,u3|u2) = 0 if and only if σ22σ13−σ12σ23= 0 (by definition of partial correlation). One
can define a function g : Rk(k+1)/2→ R, such that g(vech(Σu)) = σ22σ13− σ12σ23. Thus,
∇g?= (0, −σ23, σ22, 0, σ13, −σ12, 0, 0, 0, 0).
Applying the delta method:
√T[(ˆ σ22ˆ σ13− ˆ σ12ˆ σ23) − (σ22σ13− σ12σ23)]
d
−→ N(0,Σ) and h1,...,hr
√T[hr(XT) − hr(θ)]]
d
−→ N(0,BΣB?)
d
−→ N(0,∇g?Ω∇g).
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The Wald test of the null hypothesis corr(u1t,u3t|u2t) = 0 is given by:
T(ˆ σ22ˆ σ13− ˆ σ12ˆ σ23)2
∇g?Ω∇g
≈ χ2(1).
Tests for higher order correlations and for k > 4 follow analogously (see also Moneta, 2003).
This testing procedure has the advantage, with respect to the alternative methods, to be
straightforwardly applied to the case of cointegrated data, as will be explained in the next
subsection.
2.3. Cointegration case
A typical feature of economic time series data in which there is some form of causal depen-
dence is cointegration. This term denotes the phenomenon that nonstationary processes
can have linear combinations that are stationary. That is, suppose that each component
Yitof Yt= (Y1t,...,Ykt)?, which follows the VAR process
Yt= A1Yt−1+ ... + ApYt−p+ ut,
is nonstationary and integrated of order one (∼ I(1)). This means that the VAR process
Ytis not stable, i.e. det(Ik− A1z − Apzp) is equal to zero for some |z| ≤ 1 (L¨ utkepohl,
2006), and that each component ΔYit of ΔYt = (Yt− Yt−1) is stationary (I(0)), that
is it has time-invariant means, variances and covariance structure. A linear combination
between between the elements of Ytis called a cointegrating relationship if there is a linear
combination c1Y1t+ ... + ckYktwhich is stationary (I(0)).
If it is the case that the VAR process is unstable with the presence of cointegrating rela-
tionships, it is more appropriate (L¨ utkepohl, 2006; Johansen, 2006) to estimate the following
re-parametrization of the VAR model, called Vector Error Correction Model (VECM):
ΔYt= F1ΔYt−1+ ... + Fp−1ΔYt−p+1− GYt−p+ ut,
where Fi= −(Ik− A1− ... − Ai), for i = 1,...,p − 1 and G = Ik− A1− ... − Ap. The
(k×k) matrix G has rank r and thus G can be written as HC with H and C?of dimension
(k × r) and of rank r. C ≡ [c1,...,cr]?is called the cointegrating matrix.
Let˜C,˜H, and˜Fi be the maximum likelihood estimator of C, H, F according to Jo-
hansen’s (1988, 1991) approach. Then the asymptotic distribution of˜Σu, that is the maxi-
mum likelihood estimator of the covariance matrix of ut, is:
√T [vech(˜Σu) − vech(Σu)]
which is equivalent to equation (11) (see it again for the definition of the various operators).
Thus, it turns out that the asymptotic distribution of the maximum likelihood estimator
˜Σuis the same as the OLS estimationˆΣufor the case of stable VAR.
Thus, the application of the method described for testing residuals zero partial corre-
lations can be applied straightforwardly to cointegrated data. The model is estimated as
a VECM error correction model using Johansen’s (1988, 1991) approach, correlations are
tested exploiting the asymptotic distribution of˜Σuand finally can be parameterized back
in its VAR form of equation (3).
(12)
d
−→ N(0, 2D+
k(Σu⊗ Σu)D+?
k),(13)
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2.4. Summary of the search procedure
The graphical causal models approach to SVAR identification, which we suggest in case of
Gaussian and linear processes, can be summarized as follows.
Step 1
specification tests about normality, zero autocorrelation of residuals, lags, and unit roots
(see L¨ utkepohl, 2006). If the hypothesis of nonstationarity is rejected, estimate the VAR
model via OLS (equivalent to MLE under the assumption of normality of the errors). If unit
root tests do not reject I(1) nonstationarity in the data, specify the model as VECM testing
the presence of cointegrating relationships. If tests suggest the presence of cointegrating
relationships, estimate the model as VECM. If cointegration is rejected estimate the VAR
models taking first difference.
Estimate the VAR model Yt = A1Yt−1+ ... + ApYt−p+ ut with the usual
Step 2
the Wald statistics on the basis of the asymptotic distribution of the covariance matrix of
ut. Note that not all possible partial correlations ρ(uit,ujt|uht,...) need to be tested, but
only those necessary for step 3.
Run tests for zero partial correlations between the elements u1t,...,uktof utusing
Step 3
which is equivalent to the causal structure among Y1t,...,Ykt (cfr. section 1.2 and see
Moneta 2003). In case of acyclic (no feedback loops) and causally sufficient (no latent
variables) structure, the suggested algorithm is the PC algorithm of Spirtes et al. (2000,
pp. 84-85). Moneta (2008) suggested few modifications to the PC algorithm in order to
make the orientation of edges compatible with as many conditional independence tests as
possible. This increases the computational time of the search algorithm, but considering
the fact that VAR models deal with a few number of time series variables (rarely more
than six to eight; see Bernanke et al. 2005), this slowing down does not create a serious
concern in this context. Table 1 reports the modified PC algorithm. In case of acyclic
structure without causal sufficiency (i.e. possibly including latent variables), the suggested
algorithm is FCI (Spirtes et al. 2000, pp. 144-145). In the case of no latent variables
and in the presence of feedback loops, the suggested algorithm is CCD (Richardson and
Spirtes, 1999). There is no algorithm in the literature which is consistent for search when
both latent variables and feedback loops may be present. If the goal of the study is only
impulse response analysis (i.e. tracing out the effects of structural shocks ε1t,...,εkton
Yt,Yt−1,...) and neither contemporaneous feedbacks nor latent variables can be excluded
a priori, a possible solution is to apply only steps (A) and (B) of the PC algorithm. If the
resulting set of possible causal structures (represented by an undirected graph) contains a
manageable number of elements, one can study the characteristics of the impulse response
functions which are robust across all the possible causal structures, where the presence of
both feedbacks and latent variables is allowed (Moneta, 2004).
Apply a causal search algorithm to recover the causal structure among u1t,...,ukt,
Step 4
Step 3 is a set of causal structures, run sensitivity analysis to investigate the robustness of
the conclusions under the different possible causal structures. Bootstrap procedures may
Calculate structural coefficients and impulse response functions. If the output of
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also be applied to determine which is the most reliable causal order (see simulations and
applications in Demiralp et al., 2008).
3. Identification via independent component analysis
The methods considered in the previous section use tests for zero partial correlation on
the VAR-residuals to obtain (partial) information about the contemporaneous structure
in an SVAR model with Gaussian shocks.
and independent shocks can be exploited for model identification by using the statistical
method of ‘Independent Component Analysis’ (ICA, see Comon (1994); Hyv¨ arinen et al.
(2001)). The method is again based on the VAR-residuals utwhich can be obtained as in
the Gaussian case by estimating the VAR model using for example ordinary least squares
or least absolute deviations, and can be tested for non-Gaussianity using any normality test
(such as the Shapiro-Wilk or Jarque-Bera test).
To motivate, we note that, from equations (3) and (4) (with matrix Γ0) or the Cholesky
factorization in section 1.2 (with matrix PD−1), the VAR-disturbances utand the structural
shocks εtare connected by
ut= Γ−1
In this section we show how non-Gaussian
0εt= PD−1εt
(14)
with square matrices Γ0and PD−1, respectively. Equation (14) has two important prop-
erties: First, the vectors utand εtare of the same length, meaning that there are as many
residuals as structural shocks. Second, the residuals utare linear mixtures of the shocks
εt, connected by the ‘mixing matrix’ Γ−1
0. This resembles the ICA model, when placing
certain assumptions on the shocks εt.
In short, the ICA model is given by x = As, where x are the mixed components, s the
independent, non-Gaussian sources, and A a square invertible mixing matrix (meaning that
there are as many mixtures as independent components). Given samples from the mixtures
x, ICA estimates the mixing matrix A and the independent components s, by linearly
transforming x in such a way that the dependencies among the independent components
s are minimized. The solution is unique up to ordering, sign and scaling (Comon, 1994;
Hyv¨ arinen et al., 2001).
By comparing the ICA model x = As and equation (14), we see a one-to-one correspon-
dence of the mixtures x to the residuals utand the independent components s to the shocks
εt. Thus, to be able to apply ICA, we need to assume that the shocks are non-Gaussian
and mutually independent. We want to emphasize that no specific non-Gaussian distribu-
tion is assumed for the shocks, but only that they cannot be Gaussian.1For the shocks
to be mutually independent their joint distribution has to factorize into the product of the
marginal distributions. In the non-Gaussian setting, this implies zero partial correlation,
but the converse is not true (as opposed to the Gaussian case where the two statements
are equivalent). Thus, for non-Gaussian distributions conditional independence is a much
stronger requirement than uncorrelatedness.
Under the assumption that the shocks εtare non-Gaussian and independent, equation
(14) follows exactly the ICA-model and applying ICA to the VAR residuals ut yields a
unique solution (up to ordering, sign, and scaling) for the mixing matrix Γ−1
0
and the
1. Actually, the requirement is that at most one of the residuals can be Gaussian.
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Table 1: Search algorithm (adapted from the PC Algorithm of Spirtes et al. (2000: 84-85);
in bold character the modifications).
Under the assumption of Gaussianity conditional independence is tested by zero partial
correlation tests.
(A): (connect everything):
Form the complete undirected graph G on the vertex set u1t,...,ukt so that each
vertex is connected to any other vertex by an undirected edge.
(B)(cut some edges):
n = 0
repeat :
repeat :
select an ordered pair of variables uhtand uit that are adjacent in G
such that the number of variables adjacent to uhtis equal or greater
than n + 1. Select a set S of n variables adjacent to uht such that
uti/ ∈ S. If uht⊥ ⊥ uit|S delete edge uht— uitfrom G;
until all ordered pairs of adjacent variables uht and uit such that the
number of variables adjacent to uhtis equal or greater than n+1 and all sets
S of n variables adjacent to uhtsuch that uit/ ∈ S have been checked to see if
uht⊥ ⊥ uit|S;
n = n + 1;
until for each ordered pair of adjacent variables uht, uit, the number of adjacent
variables to uhtis less than n + 1;
(C)(build colliders):
for each triple of vertices uht,uit,ujtsuch that the pair uht,uitand the pair uit,ujtare
each adjacent in G but the pair uht,ujtis not adjacent in G, orient uht— uit— ujtas
uht−→ uit<— ujtif and only if uitdoes not belong to any set of variables S such
that uht⊥ ⊥ ujt|S;
(D)(direct some other edges):
repeat :
if uat −→ ubt, ubt and uct are adjacent, uat and uct are not adjacent and
ubtbelongs to every set S such that uat⊥ ⊥ uct|S, then orient ubt— uctas
ubt−→ uct;
if there is a directed path from uatto ubt, and an edge between uatand ubt,
then orient uat— ubtas uat−→ ubt;
until no more edges can be oriented.
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Moneta Chlaß Entner Hoyer
independent components εt(i.e. the structural shocks in our case). However, the ambiguities
of ICA make it hard to directly interpret the shocks found by ICA since without further
analysis we cannot relate the shocks directly to the measured variables.
Hence, we assume that the residuals ut follow a linear non-Gaussian acyclic model
(Shimizu et al., 2006), which means that the contemporaneous structure is represented
by a DAG (directed acyclic graph). In particular, the model is given by
ut= B0ut+ εt
(15)
with a matrix B0, whose diagonal elements are all zero and, if permuted according to the
causal order, is strictly lower triangular. By rewriting equation (15) we see that
Γ0= I − B0. (16)
From this equation it follows that the matrix B0describes the contemporaneous structure
of the variables Ytin the SVAR model as shown in equation (6). Thus, if we can identify the
matrix Γ0, we also obtain the matrix B0for the contemporaneous effects. As pointed out
above, the matrix Γ−1
0
(and hence Γ0) can be estimated using ICA up to ordering, scaling,
and sign. With the restriction of B0representing an acyclic system, we can resolve these
ambiguities and are able to fully identify the model. For simplicity, let us assume that the
variables are arranged according to a causal ordering, so that the matrix B0is strictly lower
triangular. From equation (16) then follows that the matrix Γ0is lower triangular with all
ones on the diagonal. Using this information, the ambiguities of ICA can be resolved in the
following way.
The lower triangularity of B0allows us to find the unique permutation of the rows of
Γ0, which yields all non-zero elements on the diagonal of Γ0, meaning that we replace the
matrix Γ0with Q1Γ0where Q1is the uniquely determined permutation matrix. Finding
this permutation resolves the ordering-ambiguity of ICA and links the shocks εt to the
components of the residuals utin a one-to-one manner. The sign- and scaling-ambiguity is
now easy to fix by simply dividing each row of Γ0(the row-permuted version from above)
by the corresponding diagonal element yielding all ones on the diagonal, as implied by
Equation (16). This ensures that the connection strength of the shock εton the residual ut
is fixed to one in our model (Equation (15)).
For the general case where B0is not arranged in the causal order, the above arguments
for solving the ambiguities still apply. Furthermore, we can find the causal order of the
contemporaneous variables by performing simultaneous row- and column-permutations on
Γ0 yielding the matrix closest to lower triangular, in particular˜Γ0 = Q2Γ0Q?
appropriate permutation matrix Q2. In case non of these permutations leads to a close to
lower triangular matrix a warning is issued.
Essentially, the assumption of acyclicity allows us to uniquely connect the structural
shocks εtto the components of utand fully identify the contemporaneous structure. Details
of the procedure can be found in (Shimizu et al., 2006; Hyv¨ arinen et al., 2010). In the
sense of the Cholesky factorization of the covariance matrix explained in Section 1 (with
PD−1= Γ−1
variables can be determined.
2with an
0), full identifiability means that a causal order among the contemporaneous
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In addition to yielding full identification, an additional benefit of using the ICA-based
procedure when shocks are non-Gaussian is that it does not rely on the faithfulness assump-
tion, which was necessary in the Gaussian case.
We note that there are many ways of exploiting non-Gaussian shocks for model identifi-
cation as alternatives to directly using ICA. One such approach was introduced by Shimizu
et al. (2009). Their method relies on iteratively finding an exogenous variable and regressing
out their influence on the remaining variables. An exogenous variable is characterized by
being independent of the residuals when regressing any other variable in the model on it.
Starting from the model in equation (15), this procedure returns a causal ordering of the
variables utand then the matrix B0can be estimated using the Cholesky approach.
One relatively strong assumption of the above methods is the acyclicity of the contem-
poraneous structure. In (Lacerda et al., 2008) an extension was proposed where feedback
loops were allowed. In terms of the matrix B0this means that it is not restricted to being
lower triangular (in an appropriate ordering of the variables). While in general this model
is not identifiable because we cannot uniquely match the shocks to the residuals, Lacerda
et al. (2008) showed that the model is identifiable when assuming stability of the generating
model in (15) (the absolute value of the biggest eigenvalue in B0is smaller than one) and
disjoint cycles.
Another restriction of the above model is that all relevant variables must be included in
the model (causal sufficiency). Hoyer et al. (2008b) extended the above model by allowing
for hidden variables. This leads to an overcomplete basis ICA model, meaning that there
are more independent non-Gaussian sources than observed mixtures. While there exist
methods for estimating overcomplete basis ICA models, those methods which achieve the
required accuracy do not scale well. Additionally, the solution is again only unique up to
ordering, scaling, and sign, and when including hidden variables the ordering-ambiguity
cannot be resolved and in some cases leads to several observationally equivalent models,
just as in the cyclic case above.
We note that it is also possible to combine the approach of section 2 with that described
here. That is, if some of the shocks are Gaussian or close to Gaussian, it may be advan-
tageous to use a combination of constraint-based search and non-Gaussianity-based search.
Such an approach was proposed in Hoyer et al. (2008a). In particular, the proposed method
does not make any assumptions on the distributions of the VAR-residuals ut. Basically, the
PC algorithm (see Section 2) is run first, followed by utilization of whatever non-Gaussianity
there is to further direct edges. Note that there is no need to know in advance which shocks
are non-Gaussian since finding such shocks is part of the algorithm.
Finally, we need to point out that while the basic ICA-based approach does not require
the faithfulness assumption, the extensions discussed at the end of this section do.
4. Nonparametric setting
4.1. Theory
Linear systems dominate VAR, SVAR, and more generally, multivariate time series models
in econometrics. However, it is not always the case that we know how a variable X may
cause another variable Y . It may be the case that we have little or no a priori knowledge
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Moneta Chlaß Entner Hoyer
about the way how Y depends on X. In its most general form we want to know whether
X is independent of Y conditional on the set of potential graphical parents Z, i.e.
H0: Y ⊥ ⊥ X | Z,
(17)
where Y,X,Z is a set of time series variables. Thereby, we do not per se require an a priori
specification of how Y possibly depends on X. However, constraint based algorithms typi-
cally specify conditional independence in a very restrictive way. In continuous settings, they
simply test for nonzero partial correlations, or in other words, for linear (in)dependencies.
Hence, these algorithms will fail whenever the data generation process (DGP) includes non-
linear causal relations.
In search for a more general specification of conditional independency, Chlaß and Moneta
(2010) suggest a procedure based on nonparametric density estimation. Therein, neither
the type of dependency between Y and X, nor the probability distributions of the variables
need to be specified. The procedure exploits the fact that if two random variables are in-
dependent of a third, one obtains their joint density by the product of the joint density of
the first two, and the marginal density of the third. Hence, hypothesis test (17) translates
into:
H0:f(Y,X,Z)
f(XZ)
=f(Y Z)
f(Z).(18)
If we define h1(∙) := f(Y,X,Z)f(Z), and h2(∙) := f(Y Z)f(XZ), we have:
H0: h1(∙) = h2(∙).(19)
We estimate h1and h2using a kernel smoothing approach (see Wand and Jones, 1995, ch.4).
Kernel smoothing has the outstanding property that it is insensitive to autocorrelation
phenomena and, therefore, immediately applicable to longitudinal or time series settings
(Welsh et al., 2002).
In particular, we use a so-called product kernel estimator:
??n
ˆh2(x,y,z;b) =
N2bm+d
b
KZ
ˆh1(x,y,z;b) =
1
N2bm+d
1
i=1K
?
Xi−x
Xi−x
b
?
K
?
Yi−y
b
Zi−z
b
?
K
?
Zi−z
b
????n
i=1Kp
Yi−y
b
?
Zi−z
b
Zi−z
b
??
??n
i=1K
???????n
i=1K
??
Kp
???
,
(20)
where Xi, Yi, and Zi are the ithrealization of the respective time series, K denotes the
kernel function, b indicates a scalar bandwidth parameter, and Kp represents a product
kernel2.
So far, we have shown how we can estimate h1and h2. To see whether these are different,
we require some similarity measure between both conditional densities. There are different
ways to measure the distance between a product of densities:
2. I.e. Kp((Zi − z)/b) =
kernel: K(u) = (3 − u2)φ(u)/2, with φ(u) the standard normal probability density function. We use a
“rule-of-thumb” bandwidth: b = n−1/8.5.
?d
j=1K((Zji− zj)/b). For our simulations (see next section) we choose the
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Causal Search in SVAR
(i) The weighted Hellinger distance proposed by Su and White (2008):
dH=1
n
n
?
i=1
?
1 −
?
h2(Xi,Yi,Zi)
h1(Xi,Yi,Zi)
?2
a(Xi,Yi,Zi),(21)
where a(∙) is a nonnegative weighting function. Both the weighting function a(∙), and
the resulting test statistic are specified in Su and White (2008).
(ii) The Euclidean distance proposed by Szekely and Rizzo (2004) in their ‘energy test’:
dE=1
n
n
?
i=1
n
?
j=1
||h1i− h2j|| −
1
2n
n
?
i=1
n
?
j=1
||h1i− h1j|| −
1
2n
n
?
i=1
n
?
j=1
||h2i− h2j||, (22)
where h1i= h1(Xi,Yi,Zi), h2i= h2(Xi,Yi,Zi), and || ∙ || is the Euclidean norm.3
Given these test statistics and their distributions, we compute the type-I error, or p-value
of our test problem (19). If Z = ∅, the tests are available in R-packages energy and cramer.
The Hellinger distance is not suitable here, since one can only test for Z ?= ∅.
For Z ?= ∅, our test problem (19) requires higher dimensional kernel density estimation.
The more dimensions, i.e. the more elements in Z, the scarcer the data, and the greater
the distance between two subsequent data points. This so-called Curse of dimensionality
strongly reduces the accuracy of a nonparametric estimation (Yatchew, 1998). To circum-
vent this problem, we calculate the type-I errors for Z ?= ∅ by a local bootstrap procedure,
as described in Su and White (2008, pp. 840-841) and Paparoditis and Politis (2000, pp.
144-145). Local bootstrap draws repeatedly with replacement from the sample and counts
how many times the bootstrap statistic is larger than the test statistic of the entire sample.
Details on the local bootstrap procedure ca be found in appendix A.
Now, let us see how this procedure fares in those time series settings, where other testing
procedures failed - the case of nonlinear time series.
4.2. Simulation Design
Our simulation design should allow us to see how the search procedures of 4.1 perform
in terms of size and power. To identify size properties (type-I error), H0(19) must hold
everywhere. We call data generation processes for which H0holds everywhere, size-DGPs.
We induce a system of time series {V1,t,V2,t,V3,t}n
autoregressive process AR(1) with a1 = 0.5 and error term et ∼ N(0,1), for instance,
V1,t= a1V1,t−1+ eV1,t. These time series may cause each other as illustrated in Fig. 1.
t=1whereby each time series follows an
Therein, V1,t⊥ ⊥ V2,t|V1,t−1, since V1,t−1d-separates V1,tfrom V2,t, while V2,t⊥ ⊥ V3,s, for any
t and s. Hence, the set of variables Z, conditional on which two sets of variables X and
3. An alternative Euclidean distance is proposed by Baringhaus and Franz (2004) in their Cramer test.
This distance turns out to be dE/2. The only substantial difference from the distance proposed in (ii)
lies in the method to obtain the critical values (see Baringhaus and Franz 2004).
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Moneta Chlaß Entner Hoyer
V1,t−1
V2,t−1
V3,t−1
V1,t
V2,t
V3,t
Figure 1: Time series DAG.
Y are independent of each other, contains zero elements, i.e. V2,t⊥ ⊥ V3,t−1, contains one
element, i.e. V1,t⊥ ⊥ V2,t|V1,t−1, or contains two elements, i.e. V1,t⊥ ⊥ V2,t|V1,t−1,V3,t−1.
In our simulations, we vary two aspects.
the causal dependency. To systematically vary nonlinearity and its impact, we charac-
terize the causal relation between, say, V1,t−1 and V2,t, in a polynomial form, i.e.
V2,t= f(V1,t−1) + e, where f =?p
set only bp?= 0. An additive error term e completes the specification.
The second aspect is the number of variables in Z conditional on which X and Y can
be independent. Either zero, one, but maximally two variables may form the set Z =
{Z1,...,Zd} of conditioned variables; hence Z has cardinality #Z = {0,1,2}.
To identify power properties, H0must not hold anywhere, i.e. X ⊥ ⊥ / Y |Z. We call data
generation processes where H0does not hold anywhere, power-DGPs. Such DGPs can be
induced by (i) a direct path between X and Y which does not include Z, (ii) a common
cause for X and Y which is not an element of Z, or (iii) a “collider” between X and Y be-
longing to Z.4As before, we vary the functional form f of these causal paths polynomially
where again, only bp?= 0. Third, we investigate different cardinalities #Z = {0,1,2} of set
Z conditional on which X and Y become dependent.
The first aspect is the functional form of
via
j=0bjVj
1,t−1. Herein, j reflects the degree of nonlinearity,
while bjwould capture the impact nonlinearity exerts. For polynomials of any degree, we
4.3. Simulation Results
Let us start with #Z = 0, that is, H0:= X ⊥ ⊥ Y . Table 2 reports our simulation results for
both size and power DGPs. Rejection frequencies are reported for three different tests, for
a theoretical level of significance of 0.05, and 0.1.
Take the first line of Table 2. For size DGPs, H0 holds everywhere. A test performs
accurately if it rejects H0in accordance with the respective theoretical significance level. We
see that the energy test rejects H0slightly more often than it should (0.065 > 0.05;0.122 >
0.1), whereas the Cramer test does not reject H0often enough (0.000 < 0.05,0.000 < 0.1).
In comparison to the standard parametric Fisher’s z, we see that the latter rejects H0much
more often than it should. The energy test keeps the type-I error most accurately. Contrary
to both nonparametric tests, the parametric procedure leads us to suspect a lot more causal
relationships than there actually are, if #Z = 0.
How well do these tests perform if H0does not hold anywhere? That is, how accurately
do they reject H0 if it is false (power-DGPs)? For linear time series, we see that the
nonparametric energy test has nearly as much power as Fisher’s z. For nonlinear time
4. An example of collider is displayed in Figure 1: V2,t forms a collider between V1,t−1 and V2,t−1.
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Causal Search in SVAR
Table 2: Proportion of rejection of H0(no conditioned variables)
EnergyCramer
level of significance 5%
FisherEnergy
level of significance 10%
Cramer Fisher
Size DGPs
S0.1 (ind. time series)
Power DGPs
P0.1 (time series linear)
P0.2 (time series quadratic)
P0.3 (time series cubic)
P0.3 (time series quartic)
Note: length series (n) = 100; number of iterations = 1000
0.0650.000
0.1510.1220.0000.213
0.959
0.986
1
1
0.308
0.255
0.905
0.781
0.999
0.432
1
0.647
0.981
0.997
1
1
0.462
0.452
0.975
0.901
1
0.521
1
0.709
series, the energy test clearly outperforms Fisher’s z5. As it did for size, Cramer’s test
generally underperforms in terms of power. Interestingly, its power appears to be higher for
higher degrees of nonlinearity. In summary, if one wishes to test for marginal independence
without any information on the type of a potential dependence, one would opt for the
energy test. It has a size close to the theoretical significance level, and has power similar to
a parametric specification.
Let us turn to #Z = 1, where H0:= X ⊥ ⊥ Y |Z, for which the results are shown in Table 3.
Starting with size DGPs, tests based on Hellinger and Euclidian distance slightly underreject
H0whereas for the highest polynomial degree, the Hellinger test strongly overrejects H0.
The parametric Fisher’s z slightly overrejects H0in case of linearity, and for higher degrees,
starts to underreject H0.
Table 3: Proportion of rejection of H0(one conditioned variable)
Hellinger Euclid
level of significance 5%
FisherHellinger
level of significance 10%
EuclidFisher
Size DGPs
S1.1 (time series linear)
S1.2 (time series quadratic)
S1.3 (time series cubic)
S1.4 (time series quartic)
Power DGPs
P1.1 (time series linear)
P1.2 (time series quadratic)
P1.3 (time series cubic)
P1.4 (time series quartic)
Note: n = 100; number of iterations = 200; number of bootstrap iterations (I) = 200
0.035
0.040
0.010
0.13
0.035
0.020
0.010
0
0.062
0.048
0.050
0.023
0.090
0.065
0.020
0.2
0.060
0.035
0.015
0.1
0.103
0.104
0.093
0.054
0.875
0.905
0.990
0.84
0.910
0.895
1
0.995
0.999
0.416
1
0.618
0.925
0.940
1
0.91
0.950
0.950
1
0.995
1
0.504
1
0.679
Turning to power DGPs, Fisher’s z suffers a dramatic loss in power for those polynomial
degrees which depart most from linearity, i.e. quadratic, and quartic relations. Nonpara-
5. For cubic time series, Fisher’s z performs as well as the energy test does. This may be due to the fact
that a cubic relation resembles more to a line than other polynomial specifications do.
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Moneta Chlaß Entner Hoyer
metric tests which do not require linearity have high power in absolute terms, and nearly
twice as much as compared to Fisher’s z. The power properties of the nonparametric proce-
dures indicate that our local bootstrap succeeds in mitigating the Curse of dimensionality.
In sum, nonparametric tests exhibit good power properties for #Z = 1 whereas Fisher’s z
would fail to discover underlying quadratic or quartic relationships in some 60%, and 40%
of the cases, respectively.
The results for #Z = 2 are presented in Table 4. We find that both nonparametric tests
have a size which is notably smaller than the theoretical significance level we induce. Hence,
both have a strong tendency to underreject H0. Turning to power DGPs, we find that the
Euclidean test still has over 90% power to correctly reject H0. For those polynomial degrees
which depart most from linearity, i.e. quadratic and quartic, the Euclidean test has three
times as much power as Fisher’s z. However, the Hellinger test performs even worse than
Fisher’s z. Here, it may be the Curse of dimensionality which starts to show an impact.
Table 4: Proportion of rejection of H0(two conditioned variables)
HellingerEuclid
level of significance 5%
Size DGPs
S2.1 (time series linear)0.006
S2.2 (time series quadratic) 0.000
S2.3 (time series cubic)0
S2.4 (time series quartic) 0.006
Power DGPs
P2.1 (time series linear)0.28
P2.2 (time series quadratic)0.170
P2.3 (time series cubic)0.667
P2.4 (time series quartic)0.086
Note: n = 100; number of iterations = 150; number of bootstrap iterations (I) = 100
FisherHellinger
level of significance 10%
Euclid Fisher
0.020
0.010
0.007
0
0.050
0.035
0.056
0.031
0.033
0.000
0
0.013
0.046
0.040
0.007
0
0.102
0.087
0.109
0.067
0.92
0.960
1
0.946
10.4
0.250
0.754
0.133
0.973
0.980
1
0.966
1
0.338
1
0.597
0.411
1
0.665
To sum up, we can say that both marginal independencies, and higher dimensional con-
ditional independencies, i.e. (#Z = 1,2) are best tested for using Euclidean tests. The
Hellinger test seems to be more affected by the Curse of dimensionality. We see that our
local bootstrap procedure mitigates the latter, but we admit that the number of variables
our nonparametric procedure can deal with is very small. Here, it might be promising to
opt for semiparametric (Chu and Glymour, 2008), rather than nonparametric procedures
which combine parametric and nonparametric approaches.
5. Conclusions
The difficulty of learning causal relations from passive, that is non-experimental, obser-
vations is one of the central challenges of econometrics. Traditional solutions involve the
distinction between structural and reduced form model. The former is meant to formal-
ize the unobserved data generating process, whereas the latter aims to describe a simpler
transformation of that process. The structural model is articulated hinging on a priori
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Causal Search in SVAR
economic theory. The reduced form model is formalized in such a way that it can be es-
timated directly from the data. In this paper, we have presented an approach to identify
the structural model which minimizes the role of a priori economic theory and emphasizes
the need of an appropriate and rich statistical model of the data. Graphical causal models,
independent component analysis, and tests of conditional independence are the tools we
propose for structural identification in vector autoregressive models. We conclude with an
overview of some important issues which are left open in this domain.
1. Specification of the statistical model. Data driven procedures for SVAR identification
depend upon the specification of the (reduced form) VAR model. Therefore, it is important
to make sure that the estimated VAR model is an accurate description of the dynamics of
the included variables (whereas the contemporaneous structure is intentionally left out, as
seen in section 1.2). The usual criterion for accuracy is to check that the model estimates
residuals conform to white noise processes (although serial independence of residuals is not
a sufficient criterion for model validation). This implies stable dependencies captured by
the relationships among the modeled variables, and an unsystematic noise. It may be the
case, as in many empirical applications, that different VAR specifications pass the model
checking tests equally well. For example, a VAR with Gaussian errors and p lags may fit the
data equally well as a VAR with non-Gaussian errors and q lags and these two specifications
justify two different causal search procedures. So far, we do not know how to adjudicate
among alternative and seemingly equally accurate specifications.
2. Background knowledge and assumptions. Search algorithms are based on different as-
sumptions, such as, for example, causal sufficiency, acyclicity, the Causal Markov Condition,
Faithfulness, and/or the existence of independent components. Maybe, background knowl-
edge could justify some of these assumptions and reject others. For example, institutional
or theoretical knowledge about an economic process might inform us that Faithfulness is
a plausible assumption in some contexts rather than in others, or instead, that one should
expect feedback loops if data are collected at certain levels of temporal aggregation. Yet,
if background information could inform us here, this might again provoke a problem of
circularity mentioned at the outset of the paper.
3. Search algorithms in nonparametric settings. We have provided some information on
which nonparametric test procedures might be more appropriate in certain circumstances.
However, it is not clear which causal search algorithms are most efficient in exploiting the
nonparametric conditional independence tests proposed in Section 4. The more variables
the search algorithm needs to be informed about at the same point of the search, the higher
the number of conditioned variables, and hence, the slower, or the more inaccurate, the
test.
4. Number of shocks and number of variables. To conserve degrees of freedom, SVARs
rarely model more than six to eight time series variables (Bernanke et al., 2005, p.388). It
is an open question how the procedures for causal inference we reviewed can be applied to
large scale systems such as dynamic factor models. (Forni et al., 2000)
5. Simulations and empirical applications.
SVARs, equivalent or similar to the search procedures described in section 2, have been
applied to several sets of macroeconomic data (Swanson and Granger, 1997; Bessler and
Lee, 2002; Demiralp and Hoover, 2003; Moneta, 2004; Demiralp et al., 2008; Moneta, 2008;
Hoover et al., 2009). Demiralp and Hoover (2003) present Monte Carlo simulations to
Graphical causal models for identifying
113
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Moneta Chlaß Entner Hoyer
evaluate the performance of the PC algorithm for such an identification. There are no
simulation results so far about the performance of the alternative tests on residual partial
correlations presented in section 2.2. Moneta et al. (2010) applied an independent compo-
nent analysis as described in section 3, to microeconomic US data about firms’ expenditures
on R&D and performance, as well as to macroeconomic US data about monetary policy
and its effects on the aggregate economy. Hyv¨ arinen et al. (2010) assess the performance of
independent component analysis for identifying SVAR models. It is yet to be established
how independent component analysis applied to SVARs fares compared to graphical causal
models (based on the appropriate conditional independence tests) in non-Gaussian settings.
Nonparametric tests of conditional independence, as those proposed in section 4, have been
applied to test for Granger non-causality (Su and White, 2008), but there are not yet any
applications where these test results inform a graphical causal search algorithm. Overall,
there is a need for more empirical applications of the procedures described in this paper.
Such applications will be useful to test, compare, and improve different search procedures,
to suggest new problems, and obtain new causal knowledge.
6. Appendix
6.1. Appendix 1 - Details of the bootstrap procedure from 4.1.
(1) Draw a bootstrap sampling Z∗
ˆf(z) = n−1b−d?n
(2) For t = 1,...,n, given Z∗
densityˆf(x|Z∗
(3) Using X∗
defined above.
t(for t = 1,...,n) from the estimated kernel density
t=1Kp((Zt− z)/b).
t, draw X∗
t) andˆf(y|Z∗
t, and Z∗
tand Y∗
tindependently from the estimated kernel
t) respectively.
t, Y∗
t, compute the bootstrap statistic S∗
nusing one of the distances
(4) Repeat steps (1) and (2) I times to obtain I statistics {S∗
(5) The p-value is then obtained by:
ni}I
i=1.
p ≡
?I
i=11{S∗
ni> Sn}
I
,
where Snis the statistic obtained from the original data using one of the distances
defined above, and 1{∙} denotes an indicator function taking value one if the expression
between brackets is true and zero otherwise.
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