The Monetary Approach in the Presence of I(2) Components:
A Cointegration Analysis of the Official and Black Market for Foreign Currency
in Latin America
by
Panayiotis F. Diamandis1, Dimitris A. Georgoutsos2 and Georgios P. Kouretas3
June 2001
Abstract
This paper re-examines the long-run properties of the monetary exchange rate model in the
presence of a parallel or black market for U.S. dollars in two Latin American countries under
the twin hypotheses that the system contains variables that are I(2) and that a linear trend is
required in the cointegrating relations. Using the recent I(2) test by Rahbek et al. (1999) to
examine the presence of I(2) and I(1) components in a multivariate context we find that the
linear trend hypothesis could not be rejected and we find evidence that the system contains
two I(2) variables for each country namely, Chile and Mexico, and this finding is reconfirmed
by the estimated roots of the companion matrix (Juselius, 1995). The I(2) component led to
the transformation of the estimated model by imposing long-run but not short-run
proportionality between domestic and foreign money. Three statistically significant
cointegrating vectors were found and, by imposing linear restrictions on each vector as
suggested by Johansen and Juselius (1994) and Johansen (1995b), the order and rank
conditions for identification are satisfied while the test for overidentifying restrictions was
significant for either case. The main findings suggest that we reject the forward-looking
version of the monetary model for each country, but the unrestricted monetary model is still a
valid framework to explain the long-run movements of the parallel exchange rate in both
countries. Furthermore, we test for parameter stability using the tests developed by Hansen
and Johansen (1993) and it is shown that the dimension of the cointegration rank is sample
dependent while the estimated coeffficients do not exhibit instabilities in recursive estimations.
Key words: I(2) cointegration, monetary model, parallel foreign exchange market,
identification, temporal stability.
JEL Classification numbers: F31, F33, C32, C51, C52
Part of this paper was written while the third author was Visiting Fellow at the Department of
Economics, European University Institute, San Domenico di Fiesole, Italy. The hospitality of
EUI and the Robert Schuman Centre is gratefully acknowledged. The third author also
acknowledges generous financial support from EUI. An earlier version of this paper was
presented at the Twelfth World Congress of the International Economic Association, Buenos
Aires, Argentina, August 23-37, 1999 and thanks are due to conference participants for
numerous helpful comments. This paper has also benefited from comments by seminar
participants at Athens University of Economics and Business, European University Institute
and University of Crete. We would also like to thank without implicating Michael Artis, Richard
Baillie, Anindya Banerjee, Soren Johansen, Katarina Juselius, Kate Phylaktis, Aris Spanos,
Peter Schmidt and Jeffrey Wooldridge for many helpful comments and discussions.
1
Department of Business Administration, Athens University of Economics and Business, GR10434, Athens, Greece.
2
Department of International and European Economic Studies, Athens University of
Economics and Business, GR-10434, Athens Greece.
3
Department of Economics, University of Crete, University Campus, GR-74100, Rethymno,
Greece, Fax (0831) 77406, email: kouretas@econ.soc.uoc.gr (corresponding author).
1. Introduction
Recently there has been a growing recognition of the importance of parallel or black
markets for foreign currency. The evidence available suggests that black markets have
recently increased in size and sophistication in many countries, in relation to capital
movements, (Gupta 1981; Edwards 1989, 1999; Agenor 1992; Kiguel and O’Connell, 1995;
and Phylaktis 1996 among others provide an extensive theoretical and empirical analysis of
these markets as well as of the determinants of the black market premia in a variety of
countries).
The emergence of parallel or black markets is a well known feature of many
developing countries for several decades, with parallel exchange rates deviating, in some
cases, considerably from official rates. Parallel markets for foreign currency are the result of
direct and indirect government intervention in the foreign exchange market. When access to
the official foreign-exchange market is limited and there are various foreign-exchange
restrictions on international transactions of goods, services and assets, an excess demand
develops for foreign currency at the official rate, which encourages some of the supply of
foreign currency to be sold illegally, at a market price higher than the official rate. The size of
of the market as well the black market premium, i.e. the amount by which the parallel market
rate exceeds the official rate, varies from country to country and depends on the type of the
exchange and trade restrictions imposed along with the degree to which these restrictions are
implemented by the government agencies (see Edwards, 1989, 1999; Montiel, Agenor and
Haque, 1993).
The main determinants of the demand for foreign currency in the parallel markets are
the following. First, legal and illegal imports, the former resulting from the existence of
rationing of foreign currency in the official market, and the latter from the different types of
prohibitions of imports which give an incentive for smuggling when duties are greater than the
black market premium. Second, domestic residents travelling abroad and facing limits on the
amount of foreign currency they can buy. Third, portfolio diversification particularly in cases
where the inflation is high and there is great uncertainty in economic activity, leading domestic
residents to hold foreign currency as an efficient way of hedging against domestic inflation.1
Finally, capital flight in the presence of political instability.
1
The main sources of the supply of currency are the following. The most significant
source is smuggling and underinvoicing of exports; when there is an export tariff,
underinvoicing allows the exporter to avoid the tariff and to sell the foreign currency which has
been illegally obtained at a premium. When an export subsidy is considered, which is less
than the black market premium, the sale of foreign currency in the parallel market could
provide a compensation greater than the subsidy loss. Additional sources of supply of foreign
currency to the parallel market is overinvoicing of imports when the tariff rate on imported
goods is sufficiently lower than the premium, foreign tourists and diversion of remittances.
The purpose of this paper is to provide an insight on the relationship that exists
between the exchange rate and several key monetary variables when a parallel or black
market for dollars exists. The analysis is done by employing a popular model used to explain
the movements of the exchange rate, namely the monetary model to the exchange rate first
developed by Frenkel (1976).2 This model suggests that the exchange rate is considered to
be the price of relative monies and thus it should be explained by the movements of the
monetary aggregates in the two countries, the corresponding real outputs and the interest
rates. Blejer (1978) has extended the monetary approach to the exchange rate to emphasize
the role of monetary factors as the main determinants of the black market rates. The
importance of the monetary factors on the behaviour of the black market rate has been
verified by several studies in addition to the empirical results presented by Blejer (1978) for
Brazil, Chile and Colombia. Thus, Gupta (1980) and Biswas and Nandi (1986) have tested the
model for India; Olgun (1984) for Turkey; and within the cointegration context Van den Berg
and Jayanetti (1993) for Pakistan, Sri Lanka and India and recently Kouretas and Zarangas
(1998) for Greece.
Our analysis is applied to two Latin American countries, Chile and Mexico and covers
the recent period of floating exchange rates. Black market for foreign currency, and in
particular for U.S. dollars, has operated continuously in most of the Latin America countries
for the past decades. The experience of these countries with chronic high inflation rates and
corresponding current account deficits since 1970s has led to the emergence of a strong
black market for dollars, one that has become an integral part of the countries’ infrastructure.
Figure 1(a) gives a plot of the black market and official exchange rates for Chile. In January
2
1981 one U.S. dollar bought 41650 Chilean pesos on the black market and by January 1987
one U.S. dollar was buying 238100 pesos on the black market. At the same time in the official
market the official exchange rate was 39000 pesos per dollar in January 1981 and 206290 in
January 1987. Similar patterns emerge for the foreign exchange markets of Mexico and they
are plotted in figure 1(b). Accordingly, figures 2(a)-(b) show the evolution of the black market
premium in these Latin American countries. These plots show that there is significant
variation in both countries and over time with respect to premia.
The analysis is conducted within the context of cointegration and therefore we
examine the existence of a long-run relationship between the black market exchange rate, the
official exchange rate and the monetary variables. Our approach is novel in a number of
ways. First, we provide a new analysis for the determination of the order of integration of the
variables. Although testing for unit roots has become a standard procedure it has been made
clear that if the data are being determined in a multivariate framework, a univariate model is
at best a bad approximation of the multivariate counterpart, while at worst, it is completely
misspecified leading to arbitrary conclusions. Therefore, we employ the recently developed
testing methodology suggested by Johansen (1992a, 1995a, 1997) extended by Paruolo
(1996) and Rahbek et al. (1999) which allows us to reveal the existence of I(2) and I(1)
components in a multivariate context. This analysis is done by testing successively less and
less restricted hypotheses according to the Pantula (1989) principle. Additionally, we apply a
recent test developed by Juselius (1995) that is based on the roots of the companion matrix
and allows us to make firmer conclusions about the rank of the cointegration space. Second,
since in a multivariate framework, such as the one given by the monetary model, a vector
error correction model may contain multiple cointegrating vectors, a question arises as to
whether one can associate all of them with the monetary model or otherwise which vector is
identified with it and what is the interpretation given to the others. Thus, following Johansen
and Juselius (1994) and Johansen (1995b) we impose independent linear restrictions on the
coefficients of the accepted cointegrating vectors. Third, given that at least one statistically
significant cointegrating vector has been found we examine the stability of the long-run
relationships through time. Hansen and Johansen (1993) propose three tests for parameter
3
stability in cointegrated-VAR systems which allow us to provide evidence of the sample
independence of the cointegration rank as well as of parameter stability.
There are several important findings which stem from our estimation approach. First,
we find evidence of cointegration between the black market Chilean peso-dollar, and the
Mexican peso-dollar exchange rates and the corresponding official rates and the monetary
variables. Furthermore, in both cases we were able to establish the presence of a common
I(2) component which was assumed to be between the Chilean and U.S. money series and
the Mexican an U.S. money series, respectively. Second, given the presence of an I(2)
stochastic trend we adopted a data transformation that allows us to move to the I(1) model,
which can simplify the empirical analysis considerably. Therefore, for both cases we tested
whether long-run proportionality between domestic and foreign money could be imposed on
the data. Third, given that three cointegrating vectors were found to be statistically significant,
for both cases under investigation, we imposed independent linear restrictions so that we
associated one vector with the monetary model, the second with the uncovered interest parity
(UIP) condition and the third one was taken to describe a relationship between the official and
black exchange rates. This joint structure is shown to be overidentified and the joint
restrictions are rejected for both the Chilean peso-dollar and the Mexican peso-dollar
exchange rates. This result implies that the monetary model in its forward-looking solution
does not hold, an outcome which is attributed to the failure of the UIP condition in the longrun. Fourth, we find that the unconditional UIP condition version of the monetary model may
still be a valid framework to explain the long-run movements of the black and official
exchange rates in Chile and Mexico. Finally, the application of the recursive tests of Hansen
and Johansen (1993) show that the dimension of the cointegration space may be sample
dependent while the estimated coefficients do not exhibit instabilities in recursive estimations.
The plan of the remainder of the paper is as follows. Section 2 presents the monetary
model in the presence of a parallel market for foreign exchange. In section 3 we discuss the
econometric methodology for modelling and testing cointegration. The data used and the
multivariate cointegration results are presented in section 4. The final section presents our
concluding remarks.
4
2. The monetary exchange rate model
Since its conception in the 1970s the monetary exchange rate model has become the
dominant theoretical model of exchange rate determination. The monetary model class of
models is based on the assumption of perfect substitutability of non-money assets so that the
exchange rate is determined only by relative excess money supplies. However, although this
model is theoretically very appealing, its empirical validity has produced conflicting results.
Furthermore, Meese and Rogoff (1983) show that a random walk model outperforms the
monetary model in out-of-sample forecasting ability. Early studies for the recent floating
exchange rate experience has shown that the monetary model is plagued by unstable
regression coefficients in term of sign, magnitude and significance. Recently, attention has
shifted towards the ability of the monetary model to adequately characterize long run
movements in the exchange rate. In particular, following the work of Engle and Granger
(1987), studies have been conducted to test the long run properties of the monetary model
using cointegration analysis. Within this context, MacDonald and Taylor (1994), Kouretas
(1997) and Diamandis et al. (1998) among others, provide evidence for the long-run validity of
the model as well as its out-of-sample forecasting performance for a number of key
currencies. Additionally, Diamandis, et al. (2000) provide further evidence in favour of the
monetary model in the presence of variables that are I(2) processes, for the case of the
drachma/dollar and drachma/mark exchange rates.3
The basic monetary model was developed by Frenkel (1976) and combines domestic
and foreign money demand functions with purchasing power parity (PPP). Moreover, the UIP
condition is invoked to derive the forward-looking version of the monetary model, under which
the exchange rate depends on all the expected realizations of the forcing variables, that is,
the monetary aggregates and the output variables.
Under these assumptions a typical monetary reduced form equation is obtained (see
Baillie and MacMahon, 1989; and MacDonald and Taylor, 1992):
et = β 0 + β 1 mt + β 2 mt* + β 3 yt + β 4 yt* + β 5 it + β 6 it* + ut
5
(1)
where et is the spot exchange rate (home currency price of foreign currency); mt denotes the
domestic money supply; yt denotes domestic income; i t denotes the short-term domestic
interest rate; corresponding foreign magnitude are denoted by an asterisk; ut is a disturbance
error; and all variables apart from the interest rate terms, are expressed in natural logarithms.
The expected signs of the coefficients in (1) are:
β 4 > 0, β 5 > 0 , β 6 < 0 .
β 0 > 0, β 1 > 0, β 2 < 0, β 3 < 0,
The Keynesian (sticky-price) model assumes opposite signs for
the interest rates. Different signs of the interest rate coefficients in equation (1) will also be
produced under imperfect substitutability between the assets of the two countries. Associated
with equation (1) is a set of coefficients restrictions that are regularly imposed and tested. The
most important restriction is whether proportionality exists between the exchange rate and
relative monies ( β 1 = − β 2 = 1). Moreover, the assumptions that the income and interest rate
elasticities for money demand are equal in both countries, ( β 3 = − β 4 ) and ( β 5 = − β 6 ), are
often being tested.
Blejer (1978) extended the monetary approach to emphasize the role of monetary
factors as the main determinants of the black market rates. Blejer constructs a model of the
black market exchange rate by incorporating a flow black market for foreign exchange into a
monetary model in which the rate of devaluation of the official exchange rate is fixed by the
authorities according to some reaction function aimed at maximizing a government utility
function. In this section, we follow Phylaktis (1996) and Kouretas and Zarangas (1998) and
we provide a simplified version of the model.
As a starting point we consider that the black market exchange rate depends on: (1)
the underlying supply and demand for foreign currency, which according to PPP are in the
long-run driven by countries’ price levels, (2) the level of the official exchange rate, and (3) the
diverse set of policies and institutions that govern the legal exchange market, e.g., rationing
procedures, who is permitted to buy and sell there, and of course the severity and likelihood
of penalties for dealing in the black market. If this latter set of policies and institutions is
stable, we can then investigate whether there is a linear long-run equilibrium relationship
between the parallel and official exchange rate as well as between the parallel exchange rate
and the two price levels.
6
The black market rate is determined by the interaction between demand for and
supply of foreign currency in the black market. The demand for foreign currency in this market
depends positively on the return from holding this asset. Furthermore, this return is a function
of the expected rate of appreciation of the foreign currency in the black market. If we assume
that economic agents form their expectations by comparing the movements of the exchange
rate with the movements of the ratio between domestic and foreign price level, then the
demand for foreign currency can be described as follows:
D b = β 0 + β 1 ( p − p* − eb ) ,
β 1 > 0,
(2)
*
where D b is the demand for foreign currency in the black market, p and p are respectively
the domestic and foreign price level, and eb is the black market exchange rate. Therefore, in
*
case that p rises faster than p and at the same time there is no corresponding increase in
the parallel market exchange rate, the economic agents expect a depreciation of the parallel
exchange rate by a percentage equal to the observed inflation rate differential.
The supply of foreign currency to the market is provided mainly through receipts from
the overinvoicing of imports and underinvoincing of exports as well receipts from tourism,
shipping and immigrants’ remittances. These activities are positively related to the differential
between the official and the black market exchange rates. As the differential increases, the
profit possibilities increase leading to higher incentive to divert foreign exchange to the
parallel market. The supply function of foreign currency to the black market is given as
follows:
S b = γ 0 + γ 1 (eb − e0 ) ,
γ 1 > 0,
(3)
where S b is the supply of foreign currency to the black market, and e0 is the official exchange
rate. Both exchange rates are defined as domestic currency per one unit of foreign currency.
All variables are in logarithms.
7
Equating the demand for and the supply of foreign currency in the black market and
solving for eb , we obtain
eb = α 0 + α 1e0 + α 2 p + α 3 p*
where
α1 =
γ1
γ 1 + β1
and α 2 =
(4)
β1
, and α 0 > 0 , α 1 > 0 , α 2 > 0 , and α 3 < 0 .
γ 1 + β1
The above formulation considers the black market exchange rate, is being a weighted
average of the official exchange rate, e0 , and the price differential, which essentially is the
PPP exchange rate. Absence of direct or indirect official intervention in the foreign exchange
market through the imposition of capital controls the official exchange rate will converge to the
PPP rate in the long-run while it will be equal to the black market rate leading to a gradual
elimination of the black market for foreign currency. In case though, that intervention of some
form exists, then the official exchange rate will be different from the PPP rate, and the black
rate will be a function of the official rate and the equilibrium rate implied by PPP.
Substituting equation (4) in (1) we obtain the monetary model relationship in the
presence of a black market rate
e pt = β 0 eot + β 1 mt + β 2 mt* + β 3 yt + β 4 yt* + β 5 it + β 6 it* + ut
(5)
Model (5) implies that an increase in the domestic money supply results in a domestic
money market disequilibrium. As economic agents get rid off their excess cash balances,
domestic prices rise. This creates expectations of exchange rate depreciation and an
increase in the demand for black market dollars. This in turn increases the differential
between the official and the black market rate, increasing the incentive to underinvoice
exports, to smuggle imports, or to divert remittances through the black market. Although this
increase in the supply of foreign currency in the black market will reduce the upward pressure
on the black market exchange rate, a higher stock of money will overall be associated with a
depreciation of the parallel market rate.
8
3. Econometric Methodology
Our cointegration analysis is based on the multivariate cointegration technique
developed by Johansen (1988, 1991) and extended by Johansen and Juselius (1990, 1992)
which is a Full Information Maximum Likelihood (FIML) estimation method. It makes use of
the information incorporated in the dynamic structure of the model and it also estimates the
entire space of the long-run relationships among a set of variables, without imposing a
normalization on the dependent variable a priori. Although the Johansen procedure is well
known we discuss it briefly in light of some recent extensions of the methodology that are
applied in this paper.
Consider a p-dimensional vector time series z t with an autoregressive representation
(AR) which in its error correction form is given by
k −1
∆z t = ∑ Γi ∆z t −i + Π z t −1 + γDt + µ 0 + µ1t + ε t ,
t = 1,...., T
(6)
i =1
where z t = [e p , e0 , m, m , y, y , i, i ]t as defined in section 2, z k +1 ,....., z 0 are fixed and
*
*
*
ε t ~ Niid p (0, Σ ) . The adjustment of the variables to the values implied by the steady state
relationship is not immediate due to a number of reasons like imperfect information or costly
arbitrage. Therefore, the correct specification of the dynamic structure of the model, as
expressed by the parameters (Γ1,.........,Γk−1,γ ) , is important in order that the equilibrium be
revealed. The matrix Π = αβ ' defines the cointegrating relationships, β , and the rate of
adjustment,
α,
of the endogenous variables to their steady state values. D t is a vector of
nonstochastic variables, such as centered seasonal dummies which sum to zero over a full
year by construction and are necessary to account for short-run effects which could otherwise
violate the Gaussian assumption, and/or intervention dummies;
µ
is a drift and T is the
sample size.
If we allow the
parameters of the model
θ = (Γ1 ,......, Γk −1, Π, γ , µ , Σ)
to vary
unrestrictedly, then model (6) corresponds to the I(0) model. The I(1) and I(2) models are
9
obtained if certain restrictions are satisfied. Thus, the higher-order models are nested within
the more general I(0).
It has been shown (Johansen, 1991) that if z t ~ I (1) , then that matrix Π has reduced
rank r < p, and there exist pxr matrices
α
and
β
such that Π = αβ ' . Furthermore,
κ
Ψ = α ⊥' (Γ) β ⊥ has full rank, where Γ = I − ∑ Γi and α ⊥ and β ⊥ are px( p − r) matrices
i =1
orthogonal to α and
β
, respectively.
Following this parameterization, there are
r
linearly-independent stationary relations
given by the cointegrating vectors β and p − r linearly-independent non-stationary relations.
These last relations define the common stochastic trends of the system and the contribute to
the various variables. By contrast the AR representation of model (6) is useful for the analysis
of the long-run relations of the data.
The I(2) model is defined by the first reduced rank condition of the I(1) model and that
Ψ = α ⊥' Γβ ⊥ = ϕη ' is of reduced rank s1 , where ϕ and η are ( p − r ) x s1 matrices and
s1 < ( p − r) .
Under these conditions we may re-write (6) as
k −2
∆2 z t = Π z t −1 − Γ∆z t −1 + ∑ Ψi ∆2 z t −1 + γDt + µ 0 + µ1t + ε t
(7)
i =1
where Ψi = −
k −1
∑Γ ,
j = i +1
i
i = 1,....., k − 2
Following Rahbek et. al (1999) we outline a representation of the restricted VAR (7)
which allows the observed process z t to have (at most) linear deterministic trends and some
or all components I(2). In general if z t ~ I ( 2) then the unrestricted linear regressor,
allows for cubic trends while the constant regressor,
µ 1t ,
µ 0 , allows for quadratic trends. Rahbek
et al. (1999) show that to guarantee linear trends in all linear combinations of z t we must
10
impose restrictions on both
spanned by
α
and
α⊥
µ1
and
µ0 .
First, the constant is decomposed into the spaces
respectively such that
−'
−
−
µ 0 = α α µ 0 + α ⊥ α ⊥' µ 0 ≡ ακ 0' + α ⊥ α ⊥' µ 0
(8)
Then, the restrictions required to guarantee the linear trends correspond to
µ1 = αβ 0'
where
(9)
β 0' = − β 'τ 1 , and
−
α ⊥' µ 0 = −ξη 0' − (α ⊥' Γ β ) β 0'
(10)
−
where
η 0' = −( β ⊥ η ) ' τ 1 = − β ⊥' τ 1 .
Note that
κ 0' , β 0'
and
η 0'
are freely varying vectors of
dimension s , r and s respectively.
Finally, Rahbek et al., (1999) provide a likelihood ratio (LR) test to test whether the
linear trend enters the cointegrating vector significantly. Thus, under H r the hypothesis of no
linear trend in
β ' zt
and therefore in the polynomial or multicointegrating relations is given by
β 0 = 0 . The likelihood ratio test for this hypothesis is given by
^ β0
r
Qβ 0 = T ∑ ln{
i =1
(1 − λ i )
^
(1 − λ i )
}
(11)
^
where
λi
are the r largest eigenvalues solving the eigenvalue problem in (7) and likewise
^ β0
λi
*
are the r largest eigenvalues solving (7) with z t replaced by z t . The test statistic for
this likelihood ratio test is asymptotically
χ 2 (r ) distributed.
Johansen (1997) shows that the space spanned by the vector z t can be decomposed
into r stationary directions,
the
directions
β,
(β ⊥1 , β ⊥2 ),
β ⊥2 = β ⊥ ( β ⊥' β ⊥ ) −1η ⊥
and p − r nonstationary directions,
where
β ⊥1 = β ⊥η
is
of
β ⊥,
and the latter into
dimension
p x s1
and
is of dimension p x s2 and s1 + s2 = p − r . The properties of the
process are described by:
11
I (2):{ β ⊥2 z t } ,
I (1):{ β ' z t },{ β ⊥1' z t },
I (0):{ β ⊥1' ∆zt },{ β ⊥2' ∆2zt },{ β 'zt + ω '∆zt }
where
ω
is a p x r matrix of weights, designed to pick out the I(2) components of z t
β ' zt
(Johansen, 1995a). Thus, we have that the cointegrating vectors
are actually I(1) and
require a linear combination of the differenced process ∆z t to achieve stationarity, i.e. the
polynomial or multicointegration cointegration (Haldrup, 1998).
Johansen (1991) shows how the model can be written in moving average form, while
Johansen (1997) derives the FIML solution to the estimation problem for the I(2) model.
Furthermore, Johansen (1995a) provides an asymptotically equivalent two-step procedure
which computationally is simpler. It applies the standard eigenvalue procedure derived for the
I(1) model twice, first to estimate the reduced rank of the Π matrix and then, for given
estimates of
α
and
β,
^ '
to estimate the reduced rank of
^
α⊥Γβ
⊥,
(Juselius, 1994, 1995,
1998).
In a multivariate context, such as the one given by the monetary model, a vector error
correction model may contain multiple cointegrating vectors, and in such a case the individual
cointegrating vectors are underidentified in the absence of sufficient linear restrictions on
each of the vectors. The issue of identification in cointegrated systems has recently been
addressed by Johansen and Juselius (1994) and Johansen (1995b).
Consider again the long run matrix Π = αβ ' and let Φ be any r x r matrix of full
rank. Then Π = αΦ Φβ ' = α
−1
restrictions on α and
β
*
β * ' , where α* = αΦ−1 and β * = Φβ '
and without imposing
so that to limit the admissible matrices, Φ , the cointegrating vectors
are not unique. In fact given the normalization under which both α and
only the space spanned by the
β
β
are calculated,
vectors is uniquely determined. Thus, we need to impose
restrictions implied by economic theory, for example homogeneity and zero restrictions, so
that we are able to discriminate between them.
12
The necessary and sufficient conditions for identification in a cointegrated system in
terms of linear restrictions on the columns of
β
are analogous to the classic identification
problem that we face in the simultaneous equations problem. Thus, the order condition for
identification of each of the r cointegrating vectors is that we can impose at least r − 1 , just
identifying restrictions and one normalization on each vector without changing the likelihood
function. This is a necessary condition. The necessary and sufficient condition for
identification
of
the
ith
cointegration
rank( R i H 1 ,...., R i H k ) ≥ k , where i
'
vector,
the
Rank
condition,
is
that
the
and k = 1,...., r − 1 and k ≠ i (Johansen and
'
Juselius, 1994). The linear restrictions of the model are of the form R i β i = 0 , where R i is a
'
( p × k i ) matrix, or equivalently by R i' H i = 0, i = 1,..., r , where H i is a known ( p × si )
design matrix which satisfies
β i = H i'τ i
and τi is a ( si × 1) vector of freely varying
parameters ( k i + si = p) . For example, if there are three accepted cointegrating vectors
among the eight variables of our model, the exact identification, according to the order
condition requires one linear restriction on each cointegrating vector and the Rank condition is
satisfied if rank ( R i H j ) ≥ 1, i ≠ j . Johansen and Juselius (1994) provide a likelihood ratio
'
statistic to test for overidentifying restrictions that is distributed as a
ν = ∑ ( p − r + 1 − si ) , where p and r
are given by the dimension p x r of
β
χ2
with
, and si is the
i
number of freely estimated parameters
τ , in vector i
, which comply with
β i = H iτ i
.
An equally important issue, along with the existence of at least one cointegration
vector, is the issue of the stability of such a relationship through time as well as the stability of
the estimated coefficients of such a relationship. Thus, Septhon and Larsen (1991) have
shown that Johansen’s test may be characterised by sample dependency. Hansen and
Johansen (1993) have suggested methods for the evaluation of parameter constancy in
cointegrated VAR models, formally using estimates obtained from the Johansen FIML
technique. Three tests have been constructed under the two VAR representations. In the “Zrepresentation” all the parameters of model (8) are re-estimated during the recursions while
under the “R-representation” the short-run parameters Γi, i = 1...k, are fixed to their full sample
values and only the long-run parameters
α
and
13
β
are re-estimated.
The first test is called the Rank test and is used to examine the null hypothesis of
sample independency of the cointegration rank of the system. This is accomplished by first
estimating the model over the full sample, and the residuals corresponding to each recursive
subsample are used to form the standard sample moments associated with Johansen’s
reduced rank. The eigenvalue problem is then solved directly from these subsample moment
matrices. The obtained sequence of trace statistics is scaled by the corresponding critical
values, and we accept the null hypothesis that the chosen rank is maintained regardless of
the subperiod for which it has been estimated if it takes values greater than one.
A second test deals with the null hypothesis of constancy of the cointegration space
for a given cointegration rank. Hansen and Johansen propose a likelihood ratio test that is
constructed by comparing the likelihood function from each recursive subsample with the
likelihood function computed under the restriction that the cointegrating vector estimated from
the full sample falls within the space spanned by the estimated vectors of each individual
sample. The test statistic is a
χ2
distributed with ( p − r ) r degrees of freedom.
The third test examines the constancy of the individual elements of the cointegrating
vectors β . However, when the cointegration rank is greater than one, the elements of those
vectors can not be identified, except under restrictions. Fortunately, one can exploit the fact
that there is a unique relationship between the eigenvalues and the cointegrating vectors.
Therefore, when the cointegrating vectors have undergone a structural change, this will be
reflected in the estimated eigenvalues. Hansen and Johansen (1993) have derived the
asymptotic distribution as well as the asymptotic variance of the estimated eigenvalues.
4. Empirical results
The monthly data for this study, relating to the Chilean peso-dollar and Mexican pesodollar official exchange rates and Chilean, Mexican and US macroeconomic variables, are all
taken from the International Monetary’s Fund International Financial Statistics CD-ROM while
the data for the black market exchange rates were taken from the monthly series in various
issues of the World Currency Yearbook and the relevant time periods are Chile (1973.101993.12) and Mexico (1976.09-1993.12). In particular, the black and official exchange rates
are expressed in units of home currency per foreign currency and they are end-of-month
14
quotations; The money stock is M1 (line 34 for Chile and Mexico and line 59 for the U.S. and
is seasonally unadjusted). Real output is proxied by manufacturing output (Chile; line 66) or
industrial production (Mexico and U.S.; lines 66 and 66c, respectively). For the U.S. the
interest rate is the Treasury bill rate (line 60c). Because sufficient interest rate date do not
exist for Chile and Mexico, we measured the cost of holding money as the annualized threemonth rate of consumer price inflation (line 64).4,5
4.1 Determination of the cointegration rank and the order of integration
The first step in the analysis is the determination of the cointegration rank index,
r,
and the order of integration of the variables. As a first check for the statistical adequacy of
model (8) we report some multivariate and univariate misspecification tests in Table 1, in
order to investigate that the estimated residuals do not deviate from being Gaussian white
noise errors. A structure of three lags for each bilateral exchange rate was chosen based on
these misspecification tests.
Specifically, the multivariate LB test for serial correlation up to the 42nd order and the
multivariate LM tests for first and fourth order residual autocorrelations are not significant,
whereas multivariate normality is clearly violated. Normality can be rejected as a result of
skewness (third moment) or excess kurtosis (fourth moment). Since the properties of the
cointegration estimators are more sensitive to deviations from normality due to skewness than
to excess kurtosis we report the univariate Jarque-Bera test statistics together with the third
and fourth moment around the mean. It turns out that the rejection of normality is essentially
due to excess kurtosis, and hence not so serious for the estimation results. The presence of
non-normality may be attributed to the fact that both the Chilean peso-dollar and the Mexican
peso-dollar official exchange rates were administratively determined throughout the period
under consideration as well as to the short-term interest rates, signifying both the high
volatility of money stock in both countries. The ARCH(3) tests for third order autoregressive
heteroscedasticity and is rejected for all equations. Again cointegration estimates are not very
2
sensitive to ARCH effects.6 The R measures the improvement in explanatory power relative
to the random walk (with drift) hypothesis, i.e. ∆xt =
µ + εt .
They show that with this
information set we can explain quite a large proportion of the variation in the exchange rates
15
and the money supply, but to a much lesser extent the variation in the output and the interest
rates.
The Johansen-Juselius multivariate cointegration technique, as explained in section
3, is applicable only in the presence of variables that are realizations of I(1) processes only or
a mixture of I(1) and I(0) processes, in systems used for testing for the order of cointegration
rank. Until recently the order of integration of each series was determined via the standard
unit root tests. However, it has be made clear by now that if the data are being determined in
a multivariate framework, a univariate model is at best a bad approximation of the multivariate
counterpart, while at worst, it is completely misspecified leading to arbitrary conclusions.
Thus, in the presence of I(1) series, Johansen and Juselius (1990) developed a multivariate
stationarity test which has become the standard tool for determining the order of integration of
the series within the multivariate context.
Additionally, when the data are I(2) one also has to determine the number of I(2)
trends, s2 , among the p − r common trends. The two-step procedure discussed in section 3
is used to determine the order of integration and the rank of the two matrices. The hypothesis
that the number of I(1) trends = s1 and the rank =
r
is tested against the unrestricted H 0
model based on a likelihood ratio test procedure discussed in Paruolo (1996) and Rahbek
et.al (1999).
Table 2(a) reports the evidence from the application of the two step procedure
discussed in section 3. The numbers refer to the value of the trace test statistics for all values
of
r
and s1 = p − r − s2 , under the assumption that the data contain linear but no quadratic
trends. The 95% critical test values reported in italics below the calculated test values are
taken from the asymptotic distributions reported in Rahbek et.al (1999, Table 1). Starting from
the most restricted hypothesis {r=0, s1 = 0, s2 = 8} and testing successively less and less
restricted hypotheses according to the Pantula (1989) principle, it is shown that the case in
favour of one I(2) component can not be rejected in both cases. Specifically, we are unable to
reject the hypothesis {r=3, s1 = 4, s2 = 1} for both the Chilean peso – dollar and the Mexican
peso – dollar case.7,8.
In addition to the formal test, Juselius (1995) offers further insight into the I(2) and I(1)
analysis as well as the correct cointegration rank. She argues that the results of the trace and
16
maximum eigenvalue test statistics of the I(1) analysis, i.e. from the estimation of the model
without allowing for I(2) trends, should be interpreted with some caution for two reasons. First,
the conditioning on intervention dummies and weakly exogenous variables is likely to change
the asymptotic distributions to some (unknown) extent. Second, the asymptotic critical values
may not be very close approximations in small samples. Juselius (1995) suggests the use of
the additional information contained in the roots of the characteristic polynomial. Table 2(b)
provides the pxk roots of the companion matrix. If there are I(2) components in the vector
process, then the number of unit roots in the characteristic polynomial is s1 + 2s2 . The results
of this test are consistent with the estimated roots of the companion matrix since for both the
Chilean peso-dollar case and the Mexican peso-dollar case there are six unit roots in the
process, four of which are I(1) components and one of which is the I(2) component, and given
that we have a system of eight variables three additional smaller roots are left in the process
associated with the three stationary long-run relationships9.
Finally, we allow for the presence of a linear trend following the work by Dornbusch
(1989) who suggests that due to both differing productivity trends in the tradeable and nontradeable goods sectors and inter-country differences in consumption patterns, a decline in
domestic prices relative to foreign prices could appear as a linear trend in the purchasing
power parity (PPP) relationship underlying the monetary model. We tested for the significance
of the deterministic trend in the multicointegrating relation by applying the likelihood ratio
statistic discussed in (12). The test statistic in the Chilean peso – dollar case is and in the
Mexican peso – dollar case is 13.69 and 15.21 respectively, with a p-value (0.00) and thus we
reject the null hypothesis that the linear trend does not enter significantly in the cointegration
vector of the multicointegrating relation.
4.2. A data transformation from I(2) to I(1)
Since the statistical inference of the I(2) model is quite complicated relative to that of
I(1) model, a data transformation that allows us to move to the I(1) model will simplify the
empirical analysis considerably without any loss of substance, and this transformation is
needed for both the Chilean peso-dollar and the Mexican peso–dollar cases. A possible
hypothesis which could be extracted from presence of an I(2) component in the system is that
17
the variable { mt − mt } is a first-order nonstationary process.10 If accepted, the implication is
*
that the domestic and foreign money aggregates are cointegrating from I(2) to I(1), and use of
the transformed data vector z t = [e p , eo , mt − mt , ∆mt , y t , it , it ] , would then allow us to
'
*
*
*
move to the I(1) model. The validity of this transformation is based on the assumption that
{mt − mt* } ~ I (1) , {e p , eo , y t , y t* , it , it* } ~ I (1) , and that {mt − mt* } is a valid restriction on
the long-run structure, but not necessarily on the short-run structure.
To test whether long-run proportionality between the domestic and foreign money
could be imposed on the data and the test statistic which is asymptotically distributed as
χ 2 (1) , is equal to 0.64 for the Chilean peso – dollar case and 0.89 for the Mexican peso –
dollar case and hence was not significant. Therefore, long-run proportionality between the
Chilean and US money stock and the Mexican and US money stock could not be rejected.
Furthermore, the I(2) test confirmed that this transformation removes all signs of the I(2)
components from the data.
The remaining analysis for Chile and Mexico will be performed in the I(1) model,
containing long-run but not short-run proportionality between the domestic and foreign money,
based on the vector [e p , eo , m − m , ∆m, y , y , i, i ] . Alternatively we could have chosen to
*
*
*
analyze the vector [e p , eo , m − m , ∆m , y, y , i, i ] as it corresponds to the same likelihood
*
*
*
*
function. Since, we are interested in how the exchange rate reacts to disequilibrium positions
in the domestic money we choose the first alternative.
To assess the statistical properties of the chosen variables for both cases the test
statistics reported in Table 3 are useful. The test of long-run exclusion is a check of the
adequacy of the chosen measurements and show that none of the variables can be excluded
from the cointegration space. The test for stationarity indicate that none of the variables can
be considered stationary under any reasonable choice of r . Finally, the test of weak
exogeneity shows that the output and possibly the domestic interest variables can be
considered weakly exogenous for the long-run parameters
β.
All three tests are
χ2
distributed and are constructed following Johansen and Juselius (1990, 1992). Furthermore,
table 3 presents diagnostics on the residuals from the cointegrated VAR model which indicate
18
that they are i.i.d. processes since no evidence of serial correlation or non-normality was
detected. This provides further support for the hypothesis of a correctly specified model.
4.3. The empirical analysis of the transformed I(1) model
All results discussed in this section are based on the analysis of model (2) with the
reduced rank condition on Π imposed for k = 3 and r = 3 applied to the transformed vector
for Chile and Mexico z t = [e p , eo , m − m , ∆m, y, y , i, i ] .
*
*
*
Table 4 reports the unrestricted estimates of the normalized cointegrating vectors
which are based upon eigenvectors obtained from an eigenvalue problem resulting from
Johansen’s reduced rank regression approach. The estimated parameters, in both cases,
carry signs which are in line with those that the monetary model predicts in (1) (expressed in
implicit form that the estimates correspond to the elements of an eigenvector).
Given the presence of three cointegrating vectors we continue now with the economic
identification of our system. On the first cointegrating vector we impose five restrictions,
namely proportionality between the exchange rate and relative monies and exclusion of the
official exchange rate, the growth of domestic money as well as of the two interest rates. This
long-run relationship is necessary to hold in the forward looking solution of the monetary
model when the variables are I(1) processes, the UIP condition is invoked and no bubbles are
present in the foreign exchange market. In fact, the imposition of these five restrictions
overidentifies this relationship. Identification of the second cointegration vector requires a set
of restrictions that is independent of the one imposed on the first one. This implies that from
the accepted cointegrated vectors only one can possibly describe the long-run monetary
model and this is in variance with the cointegrating results on the monetary model which other
researchers report (e.g. MacDonald and Taylor, 1992, 1994), where they conclude that as
many four vectors can be considered as possibly explaining the monetary model, but in line
with the recent results of Kouretas (1997) and Diamandis et al. (1998, 2000). The second
vector can be interpreted as a particular variant of the UIP condition for countries like Chile
and Mexico, which has been suffering from chronic budget deficits and have adopted a policy
19
of high interest rates to finance these deficits with increasing capital inflows while at the same
time the Central Bank of Chile and the Central Bank of Mexico had been using the exchange
rate as a target for the monetary policy in an effort to combat double digit inflation rates,
(Edwards, 1988, 1989; Edwards and Montiel, 1989). During the period under examination
Latin American countries have experienced serious financial imbalances and a quite
contrasting behaviour of net capital flows. In late 1970s those capital flows were associated
mainly with foreign direct investments while in early 1990s there was a tremendous surge in
portfolio funds following the market oriented reforms adopted by almost all the countries in the
region. In the meantime, the 1980s, the area experienced a drying up of private international
financing which resulted to significantly negative net resource transfers.
A common feature for the majority of Latin American countries had been the
restrictions on international capital mobility through a variety of means like administrative
controls, outright prohibitions etc. However, the true degree of capital mobility was
substantially higher than what the legal restrictions would imply. This has been clearly
documented either by examining the historical events following the 1982 debt crisis and the
ensuing massive private capital outflows and/or from several recent papers (Edwards, 1994,
1998 and 1999; Phylaktis, 1991). The foreign exchange restrictions initially were adopted in
order to defend the domestic currency from devaluation pressures. In fact we observed a
significant increase in capital controls prior to the abandonment of the fixed peg and a
substantial increase of the black market premium (Edwards, 1998; Montiel and Reinhart,
1999). In the early 1990s several Latin American countries- with the exception of Mexicoresorted to exchange controls in order to prevent the real appreciation of their currencies.
This appreciation was the outcome that the capital inflows had on the monetary base with a
resulting negative impact on inflation. Most countries tried to deal with this situation through
the imposition of controls on capital inflows and sterilized interventions. The latter mechanism
has been used by almost all countries in the region although its effectiveness in the medium
to long run is very doubtful due to the high cost it imposes on the central bank and the higher
interest rates it generates. Chile has been an exception to this situation through the adoption
20
of a policy towards higher exchange rate flexibility based on a crawling band system which
helped it to maintain the real appreciation of peso to controlled levels.
Finally, on the third vector we impose the proportionality hypothesis between the
black and official exchange rates and zero restrictions on all other coefficients, as well as on
the constant term, and this set up provides a direct test of long-run informational efficiency
between the two markets (Moore and Phylaktis, 1996).
Imposing the above restrictions on the transformed vector for Chile and Mexico, the
matrix of the linear and homogeneous restrictions is the following.
1 0 − 1 0
β = 0 0 0 β 3
1 − 1 0
0
where
β3
β5
β8 0
0
0
0
0
0
1 − 1
0 0
(12)
is expected to be negative.
The results of the estimated restricted vectors along with the likelihood test for the
acceptance of the overidentifying restrictions, for both the Chilean peso-dollar and the
Mexican peso-dollar exchange rates, are given in Table 4. According to the evidence we
reject the joint restrictions for both cases which implies that for both countries we reject the
forward-looking version of the monetary model.
In order to uncover which of the three structures, the monetary model in its forwardlooking version (i. e. the interest rates are excluded) or the UIP condition in the long-run (i.e
the interest rate differential is stationary) or the proportionality hypothesis between the black
and official exchange rates, is responsible for the afore-mentioned result, we tested each one
of them separately. This can be accomplished by imposing the same restrictions on all three
cointegrating vectors (Johansen and Juselius, 1992) and the test statistics is distributed as
χ2
with ( p − s)xr degrees of freedom. The test results imply that we are unable to reject the
coefficient restrictions implied by the monetary model given in equation (1). On the contrary in
both cases we rejected the result that the UIP condition is encompassed in the cointegrating
space we have estimated. This finding may be attributed to the extensive foreign exchange
controls which still exist in both countries and cause a continuous deviations from the UIP
21
condition. Finally, long-run informational efficiency holds in both countries implying the black
and official exchange markets have the ability to process information efficiently.
Figures 3-5 present the evidence from the Hansen-Johansen (1993) recursive
analysis on the sample independence of the Johansen procedure results. The overall
conclusion drawn from the three tests is mixed and it may suggest that there is evidence of
sample dependency of the cointegration results. Specifically, Figures 3(a) and 3(b) show that
the rank of the cointegration space depends on the sample size from which it has been
estimated, since the null hypothesis of a constant rank is rejected for both the Chilean pesodollar and the Mexican peso-dollar cases. Figures 4(a) and 4(b) clearly indicate that we are
always unable to reject the null hypothesis for the sample independence of the cointegration
space for a given cointegration rank for both cases. Finally, the last two figures 5(a) and 5(b)
in each case provide overwhelming evidence in favour of the constancy of the cointegrating
vectors since no substantial drift was detected on the time paths of the eigenvalues. The last
finding seems to indicate that the maximum likelihood estimates do not display considerable
instabilities in recursive estimates. These results further reinforce our conclusion that the
unrestricted monetary model of exchange rate determination is a valid framework to analyze
movements of the Chilean and Mexican black market exchange rates from a long-run
perspective.
5. Conclusions
In this paper we have examined the long-run properties of the monetary exchange
rate model modified to incorporate the existence of a substantial black market for U.S. dollars
for two Latin America countries, Chile and Mexico under the twin hypotheses that the system
contains variables that are I(2) and that a linear trend is required in the cointegrating relations.
The data used are monthly and are Chile (1973.10-1993.12) and Mexico (1976.09-1993.12).
Several recent developments in the econometrics of non-stationarity and cointegration were
applied and a number of novel results stem from our analysis. First, this paper makes use of
the recently developed testing methodology developed by Johansen (1992a, 1995a, 1997)
and extended by Paruolo (1986) and Rahbek et al. to test for the existence of I(2) and I(1)
components in a multivariate context. Additionally, we estimated the roots of the companion
22
matrix as suggested by Juselius (1995) in order to make firmer conclusions about the rank of
the cointegration space. The joint hypothesis of three cointegration vectors and one I(2)
component could not be rejected for both countries an outcome that led to the transformation
of the basic monetary model to contain I(1) variables and in which the rate of growth of
domestic money plays a significant role. Second, given that three cointegration vectors were
accepted, we formally imposed independent linear restrictions on each vector as suggested
by Johansen and Juselius (1994) and Johansen (1995) in order to identify our system. Based
on a likelihood ratio test for overidentifying restrictions (Johansen and Juselius, 1994) we
rejected the joint restriction that the system represents the forward looking version of the
monetary model for either case. Given this negative result we then tested whether
independently the unrestricted version of the monetary model, the UIP condition and the
proportionality between the black and official exchange rates could be considered and the
results show that the UIP condition is not valid as a long-run relationship while the
unrestricted version of the monetary model can still be a valid framework to investigate the
long-run movements of the Chilean and Mexican black market exchange rates. There is also
evidence of long-run informational efficiency in the black market which implies that this market
processes information efficiently to the official market. Finally, we tested for parameter
stability and it is shown that the dimension of the cointegration rank is sample dependent
while the estimated coefficients do not exhibit instabilities in recursive estimations.
23
Footnotes
1. Gulati (1988) estimated that during the period 1977-83 underinvoicing of exports as a
percentage of official exports was 20% for Argentina, 13% for Brazil and 34% for Mexico.
2. Apart for the monetary class of models, two other group of models have been developed to
explain the behaviour of black market exchange rates. One group of models evolved from the
theory of international trade and emphasize the transactions demand for foreign currency,
(see, for example, Sheikh, 1976; Pitt, 1984). Another class of models, the portfolio balance of
models, combines the characteristics of real trade models, by taking into account the flow
considerations for black market dollars with the characteristics of monetary approach models
by emphasizing the role of asset composition in the determination of the black market
exchange rates (see, for example, Dornbusch, et al., 1983; Phylaktis, 1991).
3. The important link between exchange rates and fundamentals and the relevance of the
monetary model to the exchange rate determination was again discussed in a series of recent
papers Rogoff (1999), Flood and Rose (1999) and MacDonald (1999).
4. Availability of data is a major problem with all Latin American countries and this fact
restricts our choices of measures. For Mexico, there is a treasury bill rate series available that
begins in January 1978 and it could be used as a proxy for short-term interest rate. However
when this series is compared to the inflation rate series the latter is smoother, which may be
the result of continuous intervention of the Central Bank of Mexico. Furthermore, the treasury
bill market in Mexico was substantially thin for most of the period. Similarly, for Chile a deposit
rate series exist from January 1977 but there is also doubt about its usefulness. Finally, we
note that we need to use as much as long data series following Hakkio and Rush (1991) who
demonstrate the difficulties of detecting cointegration over short periods.
5. For an early justification of inflation as a measure of the cost holding money see Cagan
(1956) and Wong (1977).
6. Gonzalo (1994) shows that the performance of the maximum likelihood estimator of the
cointegrating vectors is little affected by non-normal errors. Lee and Tse (1996) have shown
similar results when conditional heteroskedasticity is present.
24
7. The calculations of all tests as well as the estimation of the eigenvectors have been
performed using the program CATS 1.1 in RATS 4.20 developed by Katarina Juselius and
Henrik Hansen, Estima Inc. Illinois, 1995.
8. A small sample adjustment has been made to the Trace test statistics, Q r , for the I(1)
analysis
k
^
−2lnQ= −(T − kp) ∑ln(1− λ i ) as suggested by Reimers (1992)
i = r0+1
9. Madhavi and Zhou (1994) have shown that the Mexican peso-dollar official exchange rate
is I(2) and McNown and Wallace (1994) have shown that the Chilean peso-dollar official
exchange rate and the Chilean money stock are also I(2) variables. Both these works have
use univariate tests to reach their conclusion and we have already discussed how
inappropriate these tests are.
10. The assertion that the domestic and foreign money are I(2) comes from recent empirical
work on modeling money demand functions which suggest that nominal money stocks are
I(2), (see Johansen, 1992c; Haldrup, 1994; Paruolo, 1996; and Rahbek et al. (1999) for UK
monetary data and Juselius, 1994 for Danish data).
25
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29
Table 1. Residual misspecification tests of the model with k = 3
Eq.
∆eb
∆eo
∆m
∆m*
∆y
∆y*
∆i
∆i*
σε
0.0023
0.0018
0.0044
0.0056
0.0012
0.0145
0.0033
0.0125
Chilean Peso-Dollar
LB(42)
ARCH(3)
42.34
3.35
44.32
2.89
54.67
4.45
44.82
5.56
46.98
3.46
61.22
0.99
53.45
2.32
37.28
1.98
NORM(3)
6.98
12.34
5.57
4.56
8.67
1.23
22.34
5.97
R2
0.69
0.55
0.77
0.62
0.73
0.45
0.46
0.42
Mexican Peso-Dollar
σε
LB(42)
ARCH(3)
NORM(3)
Eq.
R2
0.0037
47.00
2.68
7.89
0.72
∆eb
0.0015
43.00
1.99
13.55
0.54
∆eo
0.0032
53.93
2.31
9.23
0.80
∆m
0.0127
55.42
2.42
3.38
0.84
∆m*
0.0013
34.56
4.28
9.76
0.85
∆y
0.0009
51.28
1.34
0.98
0.50
∆y*
0.0022
26.78
0.21
19.34
0.41
∆i
0.0008
33.24
2.47
5.78
0.49
∆i*
Notes: LB is the Ljung-Box test statistic for residual autocorrelation, ARCH is the test for
heteroscedastic residuals, and NORM the Jarque-Bera test for normality. All test statistics are
distributed as
Case
χ 2 with the degrees of freedom given in parentheses.
Multivariate Residuals Diagnostics
L-B(2938)
LM(1)
LM(4)
χ2
(14)
CP/USD
1933.29(0.20)
64.09(0.07)
59.79(0.14)
35.82(0.00)
MP/USD
2009.33(0.15)
66.22(0.06)
62.78(0.12)
37.67(0.00)
Notes: L-B is the multivariate version of the Ljung-Box test for autocorrelation based on the
estimated auto- and cross - correlations of the first [T/4=51] lags with 2938 degrees of
freedom. LM(1) and LM(4) are the tests for first and fourth-order autocorrelation distributed as
2
2
a χ with 24 degrees of freedom and χ is a normality test which is a multivariate version
of the Shenton-Bowman test with 14 degrees of freedom.
30
Table 2. Testing the Rank of the I(2) and the I(1) Model
Testing the joint hypothesis H ( s1 ∩ r)
Chilean Peso-Dollar
Q ( s1 ∩ r / H 0 )
p-r r
8
0
7
1
6
2
5
3
4
4
3
5
2
6
1
7
s2
Qr
1152.9
441.5
942.4
397.4
895.4
351.6
795.0
356.5
725.7
311.2
691.8
269.2
676.4
317.9
586.8
274.0
528.1
233.8
504.2
198.2
576.8
283.3
469.9
241.2
401.9
202.8
362.75
167.9
322.4
137.0
489.4
252.3
378.2
211.6
304.1
174.9
253.0
142.2
209.5
113.0
204.9
86.7
425.1
225.6
309.2
186.1
219.3
151.3
169.3
119.8
125.8
92.2
107.7
68.2
99.3
47.6
8
7
6
5
4
3
2
368.3
202.2
252.1
164.6
156.2
130.9
93.5
101.5
70.2
75.3
47.3
53.2
39.6
34.4
45.1
19.9
1
350.6
182.6
239.5
146.8
146.7
115.4
79.2
87.2
58.6
62.8
30.6
42.7
16.0
25.4
4.4
12.5
Mexican Peso - Dollar
Q ( s1 ∩ r / H 0 )
p-r r
8
0
7
1
6
2
5
3
4
4
3
5
2
6
1
7
s2
Qr
989.4
441.5
813.5
397.4
710.6
351.6
688.3
356.5
564.2
311.2
519.1
269.2
566.7
317.9
445.4
274.0
401.4
233.8
345.9
198.2
471.8
283.3
362.9
241.2
313.0
202.8
237.5
167.9
206.2
137.0
392.7
252.3
285.0
211.6
241.8
174.9
166.0
142.2
116.3
113.0
128.5
86.7
334.7
225.6
231.1
186.1
193.7
151.3
117.5
119.8
78.1
92.2
53.5
68.2
89.2
47.6
8
7
6
5
4
3
2
284.2
202.2
194.2
164.6
148.1
130.9
94.1
101.5
55.8
75.3
32.4
53.2
31.3
34.4
28.7
19.9
1
265.4
182.6
177.3
146.8
124.7
115.4
85.3
87.2
48.2
62.8
25.0
42.7
14.0
25.4
5.4
12.5
Notes: p is the number of variables, r is the rank of the cointegration space, s1 is the number
of I(1) components and s2 is the number of I(2) components. The numbers in italics are the
95% critical values (Rahbek, et al., 1999, Table 1). For all tests a structure of three lags for
both black exchange rates was chosen according to a likelihood ratio test, corrected for the
degrees of freedom (Sims, 1980) and the Ljung-Box Q statistic for detecting serial correlation
in the residuals of the equations of the VAR. A model with an unrestricted constant in the VAR
equation is estimated for all three cases according to the Johansen (1992b) testing
methodology.
31
Table 2. Continues
Modulus of 9 largest roots
Chilean peso
Unrestricted model
r=3
0.98 0.98 0.97 0.97 0.95 0.88 0.71 0.55 0.42
1.00 1.00 1.00 1.00 1.00 0.94 0.65 0.48 0.33
Mexican peso
Unrestricted model
r=3
0.99 0.99 0.95 0.95 0.90 0.90 0.81 0.72 0.62
1.00 1.00 1.00 1.00 1.00 0.96 0.70 0.60 0.33
32
Table 3. Tests for long-run exclusion, stationarity, and weak exogeneity
Long-run exclusion
Stationarity
Weak exogeneity
Variable
CP/USD
MP/USD
CP/USD
MP/USD
CP/USD
MP/USD
11.34*
9.46*
33.87*
32.01*
15.45*
13.11*
ep
11.55*
13.97*
43.15*
31.93*
40.22*
41.58*
eo
22.45*
21.08*
21.56*
25.45*
11.23*
53.88*
m-m*
35.67*
24.29*
33.23*
20.45*
16.45*
20.20*
∆m
8.99*
7.99*
21.44*
20.49*
6.68
7.11
y
10.23*
8.87*
25.23*
19.71*
3.24
2.32
y*
9.67*
9.64*
17.77*
17.60*
17.56*
16.45*
i
19.67*
21.03*
19.02*
23.75*
14.22*
13.67*
i*
Notes: eo, ep, (m-m*), ∆m, y and i are respectively the spot exchange rate, the relative
monies, the first difference of the domestic money supply, the real output and the short-term
interest rate, with the U.S. magnitudes denoted with an asterisk. The long-run exclusion
2
restriction and the weak exogeneity tests are χ distributed with three degrees of freedom
2
and the 5% critical level is 7.81, and the stationarity test is a χ distributed with six degrees
of freedom and the 5% critical level is 12.59.
Case
Multivariate Residuals Diagnostics
L-B(3072)
LM(1)
LM(4)
χ2
(16)
GRD/USD
1696.12(0.26)
52.20(0.35)
61.12(0.09)
726.29(0.00)
GRD/DM
1453.60(0.28)
58.71(0.09)
63.50(0.08)
288.12(0.00)
Notes: L-B is the multivariate version of the Ljung-Box test for autocorrelation based on the
estimated auto- and cross - correlations of the first [T/4=51] lags with 3072 degrees of
freedom. LM(1) and LM(4) are the tests for first and fourth-order autocorrelation distributed as
2
2
a χ with 64 degrees of freedom and χ is a normality test which is a multivariate version
of the Shenton-Bowman test with 16 degrees of freedom.
33
Table 4. Estimated Coefficients and Hypothesis Testing
eb = β 1eo + β 2 (m − m * ) + β 3 ∆m + β 4 y + β 5 y * + β 6 i + β 7 i * + β 0 + γ 1t
CP/USD
1.00
1.00
-1.46
-6.55
3.45
11.24
-93.82
-86.46
-7.88
77.87
5.34
-33.24
0.03
0.01
-0.15
-0.11
-0.23
0.34
-1.34
0.02
1.00
-1.14
10.98
-34.56
22.56
-12.55
0.02
-0.16
0.09
0.23
MP/USD
1.00
-1.22
4.33
-3.32
1.00
-3.52
9.23
-6.22
1.00
-1.15
4.66
-4.77
Notes: The eigenvectors have been normalized
the black market exchange rate
-8.01
4.21
0.04
-0.11
-0.79 -1.45
7.21
-3.89
0.03
-0.17
1.25
0.11
-6.22
-2.55
0.19
-0.24
-011
-0.55
with respect to the estimated coefficient on
A. Tests for overidentifying restrictions
CP/USD
0
5.58(2.16) − 3.65(1.22) 0 0 − 6.6(1.34) − 3.56(2.56)
1 0 − 1
0
0
1 − 1 0.03(0.013) 0.23(0.89)
β = 0 0 0 − 29.16(3.22)
1 − 1 0
0
0
0
0 0
0
0.88(1.01)
Q(7)=43.79[0.00]
MP/USD
0
3.87(0.93) − 2.77(1.08) 0 0 − 4.6(1.1) − 2.89(3.01)
1 0 − 1
0
0
1 − 1 0.07(0.03) − 0.78(1.23)
β = 0 0 0 − 24.35(2.99)
1 − 1 0
0
0
0
0 0
0
0.99(0.65)
Q(7)=39.42[0.00]
Notes : Q denotes a likelihood ratio test for overidentifying restrictions as suggested by
Johansen and Juselius (1994) and is distributed as a with the corresponding degrees of
freedom given in parentheses. Numbers in brackets denote marginal significance levels.
Numbers in parentheses below the coefficient estimates report estimated asymptotic standard
errors which are the square roots of the computed Wald test statistics developed by Johansen
(1991).
Case
H1
( β 1 = 0, β 2 = 1,
β 4 = − β 5, β 6 = β 7 )
CP/USD
MP/DM
H2
H3
( β 1 = 0, β 4 = β 5 = 0,
β1 = 1
β 6 = −β 7)
eb is excluded
0.23
0.12
0.00
0.00
0.11
0.09
Notes: Numbers correspond to marginal significance levels of the H5 test statistic (Johansen and Juselius, 1992)
distributed as a χ
2
with five degrees of freedom,
number of vectors on which the restrictions
construction of the test are those given above.
( p − r ) xr1 , [p = number of variables, r = cointegration rank, r1 =
are imposed].
34
The coefficient estimates necessary for the
14
12
10
8
6
4
74
76
78
80
82
84
LOFFCH
86
88
90
92
LPARCH
Figure 1(a) : Official and black exchange rates
Chile-U.S. case
9
8
7
6
5
4
3
2
78
80
82
84
LOFFMX
86
88
90
92
LPARMX
Figure 1(b) : Official and black exchange rates
Mexico-U.S. case
35
1.4
1.2
1.0
0.8
0.6
0.4
0.2
0.0
-0.2
74
76
78
80
82
84
86
88
90
92
PRCH
Figure 2(a) : The black market premium
Chile-U.S. case
1.0
0.8
0.6
0.4
0.2
0.0
78
80
82
84
86
88
90
92
PRMX
Figure 2(b) : The black market premium
Mexico-U.S. case
36
1.6
1.2
0.8
0.4
0.0
83
84
85
86
87
TRACE1
TRACE2
TRACE3
88
89
90
91
TRACE4
TRACE5
TRACE6
92
93
TRACE7
CRITVAL
Figure 3(a) : The Trace Test
Chile-U.S. case
2.0
1.5
1.0
0.5
0.0
80
82
84
TRACE1
TRACE2
TRACE3
86
88
90
TRACE4
TRACE5
TRACE6
92
TRACE7
CRITVAL
Figure 3(b) : The Trace Test
Mexico-U.S. case
1 is the 5% significance level
37
1.0
0.8
0.6
0.4
0.2
0.0
80
82
84
86
88
BETA
90
92
CRITVAL
Figure 4(a) : The test for the constancy of beta
Chile-U.S.case
1.0
0.8
0.6
0.4
0.2
0.0
83
84
85
86
87
88
BETA
89
90
91
92
93
CRITVAL
Figure 4(b) : The test for the constancy of beta
Mexico-U.S.case
1 is the 5% significance level
38
1.0
1.0
0.8
0.8
0.6
0.6
0.4
0.4
0.2
0.2
80
82
84
86
88
LAMBDA1
BOUND1
90
0.0
92
80
82
BOUND2
CRITVAL
84
86
88
LAMBDA2
BOUND3
90
92
BOUND4
CRITVAL
1.0
0.8
0.6
0.4
0.2
0.0
80
82
84
86
88
LAMBDA3
BOUND5
90
92
BOUND6
CRITVAL
Figure 5 (a) : The eigenvalue test : Chile-U.S. case
1.0
1.0
0.8
0.8
0.6
0.6
0.4
0.4
0.2
0.0
0.2
83
84
85
86
87
88
89
LAMBDA1
BOUND1
90
91
92
93
83
BOUND2
CRITVAL
84
85
86
87
LAMBDA2
BOUND3
1.0
0.8
0.6
0.4
0.2
0.0
83
84
85
86
87
88
LAMBDA3
BOUND5
89
90
91
92
88
93
BOUND6
CRITVAL
Figure 5 (b) : The eigenvalue test : Mexico-U.S. case
1 is the 5% significance level
39
89
90
91
92
BOUND4
CRITVAL
93