-%p%~S r (d
POLICY RESEARCH WORKING PAPER 1 36
Coinzegration techniques are
Capital Flows
lSe;id to disentangle tne effect
and Long-Term Equilibrium o' apitalllmlsonthe
Real E h Rates ~equilibrium real exchange
Real Exchange Rates rate. Short-term flow; and
inl Chile zorrlofio inveunrent were
fsound to have no infuence,
but sustainable long-term
Ibrabim A. Elbadaawi inflows and foreign direct
Raimundo Soto
investment have an
apprecioiting e.fect. So, an
importan., part of the aual
apwrec,ation Co the Chilean
peso wou'd not require
Counterbalancing exchange
era. or macroecoinomic
The World Bank
Policy Research Departmnent
Macroeconomics and Growth Division
June 1994
Pouac RESEARCH WORKING PArER 1306
Summary findings
In the context of an empirical model, Elbadawi and Soto of the fundamentals) successfully reproduce the salient
examine rhe impact of capital flows, among other episodes in Chile's recent macroeconomic history.
fundamentals, on long-term exchange rates in Chile. Capital flows are disaggregated into four components:
The real exchange rate and its fundamentals were - Shorr-term capital flows.
found to be cointegrated during 1960-92. This L Long-term capital flows.
cointegmation allows a reinterpretation of uni-equational - Portfolio investment.
esimates of the equilibrium real exchange rate (ERER) * Foreign direct investment.
to be consistent with long-run forward-looking As expected from economic theory, short-tcrm capital
behavioral models. It also permits the cstimation of an flows and portfolio investment were found to have no
error-correction model capable of disentangling short- effect on the ERER (although they can affect the real
run from long-run shocks in observed movements of the exchange rate in the short run).
ERER. And the nonstationary nature of the fundamentals But long-term capital inflows and foreign direct
allows one to decompose innovations into pcrmanent investment have a significant appreciating effect on the
and transitory components - to get an empirical ERER.
measure of the susrainability of the fundamentals with To the extent that the recent inflow of capital to Chile
which the ERER is determined. is dominated by long-term capital flows tht are judged
In general, the estimate of the cointegration of the to be sustinable, an important part of the ensuing
ERER and its corresponding dynamic error-correction appreciation of the real exchange rare is consistent with
specification corroborates the theoretical model and cquilibrium behavior- reducing the need for
produces fairly consistent results- counterbalancing exchange rate or macroeconomic
The derived ERER index and the corresponding real policies.
exchange rate misalignment (for given sustainable values
This paper - a product of the Macroeconomics and Growth Division, Policy Research Department- is part of a larger effort
in the department to understand the links of foreign shodcs and macrocconomic policies. Copies of the paper are available free
from the World Bank, 1818 H Street NW, Washington, DC 20433. Please contact Rebecca Martin, room Ni 1-043, extension
39065 (34 pages). June 1994.
The Policy Resarch Working Paper Seies dissmi the flins of work n pro# to encoanage the chan of idea abows
deveopment wss Anobjectirv of theseies is toget thefinding out quicky. een if he prseaions are lss thn fidly polish Te
ppe carry thc names oftbe authorsadshould/ used andcitaccozigly. Thefings Mrhepremisand coduonmareshe
autho'?ew and sould mo he amibue to the Wodld am its E.ecative BaS ofd D;a: or ayof itsmbero
Produced by the Policy Research Dissemination Center
CAPITAL FLOWS AND LONG-TERM
EQUELIBRIUM REAL EXCHANGE RATES IN CHILE
Ibrahim A. Elbadawi*
The African Economic Research Consortium
Nairobi, Kenya, P.O.Box 62882
and
Raimundo Soto*
Macroeconomics and Growth Division
The Worid Bank, 1818 H Street, N.W.
Washington DC 20433, USA
* The usual disclaimer applies. Useful comments by J. Leon and P.L. Rodriguez are
acknowledged.
Table of Contents
1. Introduction .............................................. 1
2. An Empirical Model of the Equilibrium Real Exchange Rate .3
3. An Application to Chile ...................................... 9
3.1 The Long-Run Equiibrium ....9... 9
3.2 The Error Correction Model ... 13
4. Real Exchange and Equilibrium Real Exchange Rates Indices for Chile 15
5. Conclusions ............................................... 17
Tables and Figures
References
Appendix Tables and Figures
List of Tables
Table 1 Estimation Results of the Cointegration Equations
Table 2 Estmation Results of the Error-Correction Models
Table 4 Equilibum Real Exchange Rate and Misalignment
List of Figures
Figure 1 Real Exchange Rate and Capital Iflows, 1960-1992.
Figure 2 Real Exchange Rate: Actual and Equilibrium Values
Figure 3 Real Exchange Rate Misalignment
List of Appendix Tables
Table A.l Basic Data
Table A.2 Order of Integration Tests
Table A.3 Granger Causality Tests
Table A.4 Correlation Matrix for Capital Inflows Components
Table A.5 Estimated ARIMA Models for fundamentals
List of Appendix Figures
Figures A.1-A.7 Stability Tests
1. INTRODUCTION
During the last years Latin American countries, and notably Chile, have experienced an
important recovery in their ability to attract foreign capital, particularly from pivate lenders.
From an historical low level of US$ 9.5 billion per year in the 1985-88 period, capital inflows
to the region increased to US$ 25.3 billion per year in the 1989-92 period (Calvo et al, 1993).
This bonanza, however, has raised also some concerns on the part of the authorties which find
it increasingly difficult to pursue monetary and exchange rate policies. In the case of Chile, the
role of capital flows in disrupting macroeconomic management, the extent to which these inflows
are sustainable, and whether or not their influence on the real exchange rate is consistent with
equilibrium behavior, have been subjects of controversy in recent macroeconomic debates (see
Arrau et al, 1992).
From a policy point of view this debate is of importance. When dealing with massive
capital flows the authority faces a trade-off between allowing the exchange rate to appreciate,
thus having a negative impact on the competitiveness of exports, and sterilizing these effects, thus
inducing losses to the Central Bank. In the Chilean case this dilemma has a bitter precedent: the
appreciation of the exchange rate in the 1979-82 period has been repeatedly linked to the inflow
of more than US$ 11 billions and the subsequent debt crisis; the current inflow (US$ 9 billions
in the 1989-92 period) has raised justifiable concerns on the current policy. However, the
direction of causality between the RER and capital inflows has remained a controversial issue
in the Chilean economic literature. Edwards (1988) and Morande (1988) found evidence that
massive capital inflows during the late 1970s induced the appreciation of the RER and the loss
of competitiveness of Chilean exports. On the other hand, Corbo (1985), among others, presents
evidence of the reverse causality; i.e, that the mismanagement of the nominal exchange rate
(fixed for almost three years, despite large differentials between domestic and foreign inflation)
caused an increased overvaluation of the peso, inducing a substantial premium for deposits in
local currency and attrating foreign capital.
This paper contributes to this debate by estimating the long-run equilibrium path between
the RER and capital flows, among other fundamentals. The cointegrated equilibrium is obtained
from the basic model of the real exchange rate that characterizes the equilibrium as "the relative
price of non-tradables to tradables goods which, for given sustainable values of other relevant
-2 -
variables such as taxes, intemational terms of trade, commercial policy, capital and aid flows and
technology, results in the simultaneous attainment of internal and external equilibrium", (Edwards
(1989), pp. 16). We extend the standard models of Rodriguez (1989) and Elbadawi (1993) by
allowing capital flows to be disaggregated into four components: short-term capital flows, long-
term capital flows, portfolio investment and direct foreign investmenL
Unlike under other definitions (eg, the PPP theory), the equilibrium real exchange rate
(ERER) in this case experiences movements in response to exogenous and policy-induced shifts
in its real fundamentals. Furthermore, this notion of equilibrum is essentially intertemporal as
the path of the ERER is affected not only by the current values of the fundamentals, but also by
anticipations regarding the future evolution of these variables. Regarding capital flows,
consistency with arbitrage theories suggests that short-term flows and portfolio investment should
have no influence in the ERER; on the contrary, long-term flows and foreign direct investment
are expected to have permanent effects on iL In addition to such equilibrium movements, the
observed RER is also influenced -in the short to medium run by transitory shocks to the
funamentals and by macroeconomic and exchange rate policies, which are not part of the
fundamentals. RER misalignments can occur (as in the standard PPP theory) when those policies
are inconsistent with the fundamentals. For example, in a system of pegged nominal exchange
rates, expansionary fiscal and monetary policy can be a cause of persistent real overvaluation;
Edwards (1989) and Elbadawi (1989) provide strong empirical evidence on this. In this context,
we test the controversial issue of whether public saving (or contractionary fiscal policy) is an
efficient tool for sustaining real exchange rate targets, as it has been suggested in the Chilean
debate-
It is important to stress that given the intertemporal nature of the ERER definition
employed in the paper, an empirically consistent modelling of the RER is not trivial. Elbadawi
(1993) has shown however that a - cointegration-error -correction approach is adequate in this
framework because it accounts for the following desirable properties: (i) it is consistent with a
behavioral model specifying the ERER as a forward-looking function of the fundamentals; (ii)
'Edwards (1986) and chapter 2 of Edwards (1989) fonnalize this concept in the context of an intertemporal
optimizing model; see also Lizondo (1989).
-3-
it allows for flexible dynamic adjustments of the RER toward the ERER; (iii) it allows for the
influence of short to medium run macroeconomic and exchange rate policy on the RER; and (iv)
stochastic non-stationarity suggests a time series-based decomposition of the fundamentals into
permanent (sustainable) and transitory components. In the following section we state the basic
traded-nontraded model which gives the ERER that solves the equilibrium condition in the home
goods market under static expectations and assuming a given level of capital flows. Under unit-
root non-stationarity and cointegration, this model is equivalent to a model that solves the ERER
as a forward-looling function of the fundamentals. The endogenization of domestic absorption
as a function of anticipated future RER depreciation permits this forward-looking solution and
an appropriate re-interpretation of the standard model as a cointegrated relationship.
Despite the fact that the basic model allows the estimaton of the long-run link between
capital flows and the RER, the cointegration equilibrium does not offer any guidance on the
debate on the direction of causation. We address this issue by testing for the existence of
feedback effects from the RER to capital flows in the context of an error-correction model
(ECM), the dynamic counterpart of the cointegration equiibrium (Engle and Granger 1987,
Phillips and Loretan, 1991).
In section 3 the model is applied to the Chilean case to estimate a long-run cointegration
specification for the ERER, as well as the corresponding short-nm error-correction model. In
section 4 the estimated long-run relationships are used to derive the ERER and the corresponding
HER misalignment Conclusions and some policy implications are collected in section 5.
2. AN EMPIRICAL MODEL OF THE EQUILIBRIUM REAL EXCHANGE RATE
We extend the standard RER models of Roddiguez (1989) and Edwards (1987) to analyze
the effects of financial flows on the equilibrium real exchange rate in a cointegration-error
correction framework. Consider a small economy with three sectors (importables, exportables and
non-tradable goods) for which the international price of traded goods is assumed to be exogenous.
The domestic price of tradables, then, is determined by the level of tariffs and the nominal
4 -
exchange rate (E). Let P,and Pm be the dollar-denominated international pAces of exportables
and importables and t, and t,, the net export and import tax rates, respectively. The (domestic)
price index of tradable goods is defined as:
Pr = E(1 -t)P;J'.f(I +t ,)JPl (1)
On the other hand, the price of non-tradables is endogenously determined as the result of
the interaction of supply and demand. The latter is disaggregated into private and public
components (Em and EGN, respectively); we assume that the proportion of private expenditure
allocated to non-tradable goods depends on the prices of exports, imports and non-traded goods
(P1, P., and P., respectively), and that government expenditures in non-tadables is a fraction (g,)
of total government expenditure. Hence, the total demand for non-traded goods is expressed as:
EN EEPN+ EG = d,jP P,PfA-g. Y +gNg. (2)
wrhere d(.) is the proportion of private expenditure (absorption less total government expenditure)
in non-traded goods, A is absorption, Y is income, and g is the ratio of government expenditures
to income.
The supply of nontraded goods, which is also specified as a fraction of income, depends
on the prices of tradable and non-tradable goods:
SNV SnM(P- Piw PJ Y 3
Equation (4) sets the equilibrium condition in the non-traded goods market (SN =EN),
which in turn determines P.:
s*(PX, Pl, P) = dQt, P P.){4 -g] + gj g (4)
Defining the real exchange rate, e, as the relative price of non-traded to tradable goods we have:
-5-
e_ _ - X3 (5)
EPx PM- Et Pm {l-t)5(1tM)I- a5;
Equations (4) and (5) can be solved for the level of the RER that ensures instantaneous
equilibrium in the nontraded goods market, for given levels of the exogenous and policy
"fundamentals".
A
e = e(-y, TOT, t,, ,, X )(
(#) (2) (+) ft) (+) (2)
where TOT represents the terms of trade (Px*IP.*). Equation (6) implies that higher levels of
absorption, trade taxes, and public expenditures on nontradables are consistent with a more
appreciated RER The effects of TOT and total government expenditures cannot be deternined
a priori; the empirical evidence, however, shows that improved TOT and higher govermment
expenditure tend to lead to RER appreciation3' The former arises because the income effect of
an improvement in the TOT usually domiinates its substitution effect, while the latter is due to
the tendency of governments to spend more on non-traded goods than the private sector-
Following Elbadawi (1993) we extend the basic model of equation (6) by endogenizing
private absorption as a function of net capital inflows and the expected real exchange rate
depreciation:
A = (NIC [t -eJ) (7)
y y
(t) (J)
where NEa are net capital inflows and te, is the expected real exchange rate. As shown below,
this extension yields a forward-looking expression for the EREFR as a function of the expected
path of its fundamentals.
2 See, for example, Edwards (1989).
3Other potential determinants of the RER. such as productivity changes, can be included by an appropriate re-
specification of s, (.).
-6 -
An empirically convenient version of the model described by equations (6) and (7) is:
Publiclnv
loge, - ,loge,41=a0+a1 log WT-21og OPEN AEt3I099JE4Lg( IDP4 )
GDPt
(8)
L-ongcap, Shoncap, PonfolioInvt F.DirectInv(
GDPr GDP, 7 GDP GDP,
= Ft
where F, represents the a vector of fundamentals (TOT, Openness, etc) and 5 is a vector of
coefficients. Note that NKI has been decomposed into long-term capital inflows (Longcap), short-
term capital inflows (Shortcap), portfolio investment (Portfolio Inv) and foreign direct investment
(F. Direct hv). In addition, public investment as ratio of GDP is included as a proxy of (l-gN),
given the difficulties to obtain reliable data on EGN.
The variable OPEN is defined as the sum of exports and imports as ratio to the GDP. ts
use as a proxy for commercial policy (t, tJ is justified because of the difficulty of obtaining
good time-series data on t1 and tE and also because it may account not only for explicit
commercial policy:but also for implicit, though very important, factors such as quotas and
exchange controls. Note that the empirical regularities regarding the signs of TOT and
govermment expenditres are assumed.4 Since the equation is relevant for the determination of
the long run RER, the short-run capital flow and portfolio investment component of capital flows
are expected to have non-significant effects.
The model in equation (8) can be solved recursively to yield:
- log et a f tpj t-(9)
Note that the equilibrium exchange rate (a) is determined by the expected long-run path of the
fundamentals. Hence, in order to have an empirical measure of the RER it is necessary to
estimate the sustainable level-of the fundamentals. Williamson (1993) recommends an ex-ante
d Equation (8) appears in several foims in the empirical tradition of the RER literature: e.g. Edwards (1986).
Elbadawi (1989 and 1993), Mwidlak et al (1987) and Valdes et al (1990).
-7-
approach, the so-called "fundamental equilibrium exchange rate" (FEER), which calls for
specifying (or assuming) behavioral specifications for the fundanentals and using the real
exchange rate equations in the context of a bigger model to derive the trajectory of the
equilibrium real exchange rate given the assumed paths of the fundamentals. The approach used
in this paper exploits the time-series properties of the variables to get the long-run trajectory of
the RER and its fundamentals and, consequently, corresponds to the ex-post version of the FEER
concept.
Stochastic non-stationary, cointegration and the ERER
When fundamentals are characterized by unit-root-nonstationary processes, the model in
equation (9) is consistent with the following long-run cointegrated equilibrium (Kaminsky,
1988)5:
logei _ , t ^ 31Fe t .nt -(10)
_-a;
where 1I(l-X)8' is the cointegrating vector and q9 is an uncorrelated random disturbance.
This is an important advantage of cointegration, as it allows the derivation of a simple
empircal framework from a much more complicated theoretical model. Nevertheless, to
determine the ERER it is necessary to find a practical approximation to the concept of
"sustainability" on the part of the fundamentals. Here again stochastic non-stationarity proves to
be a useful property. The permanent (or sustainable) components of the fundamentals can be
obtained by using a suitable time-series decomposition technique (eg, Beveridge and Nelson
(1981) or Campbell and Manldw (1987)).
s The idea of cointegration states that even though individual series may be.non-stationary, there may exist a
linear combination of them which is stationary. More formally, let the n-vector y, be composed of n-non stationary
variables (y, .- y.), then y, is said to be cointegrated if there exists at least one n-element vector [ such that P'y,
is trend stationary. This is a mild definition of cointegration (Campbell and Perron, 1991), which is more suited to
the empirical analysis of economic date since it allows the inclusion of deterministic components (such as trends and
structural break dummies) in the cointegration model. -.
-8-
This specification is also consistent with a dynamic error-correction model6, which
describes the short-run movements of the RER as arising from the presence of transitory shocks
to fundamentals and non-fundamental variables (such as exchange rate and monetary policies),
as well as a result of the self-correcting mechanism that adjusts previous period disequilibria:
A loge,, =b4-/j6Ft-loge,)+bi'AFt.i .4 AlogZ,,1 e, (11)
where Z, is a vector of stationary variables (including the rate of change in domestic credit to
GDP, short term capital inflows and the rate of nominal exchange rate devaluation), and the
disturbance ,+1 is a stationary random variable composed of the one-step-ahead forecast error in
the RER (i.e. Alog e,, - 4log et+1, ). The error-correction term 6'F1 - loge) in equation (11)
clearly incorporates the forward-looking sources of RER dynamics. Suppose, for example, that
we start from an initial condition of real overvaluation (i.e. the error-correction term is negative);
then, the self-correcting mechanism immediately calls for a future depreciation in the actual RER.
This effect is captured by the negative error-correction term and its positive coefficient in the
Alog e,+, specification. The speed at which this automatic adjustment operates depends on
parameter bo, which falls in the interval 10,1). A value of bo equal to one indicates prompt
adjustment in just one period; the smaller the value of bo, the slower the adjustment is.
In addition to the long-run (equilibrium) impact of the fundamentals on the RER, which
is captured by the cointegration vector, temporary changes in the fundamentals may also have
short-run effectsv which are captured by the vector b1. The effects of short-run shocks in exchange
rate and macroeconomic policies are given by the coefficients in b2. For example, as pointed out
by Edwards (1989), a nominal devaluation will help the adjustment process only to the extent that
the initial situation is one of overvaluation, and only if the nominai eichange rate adjustment is
accompanied by supporting macroeconomic policies; i.e, in terns of our equation the erTor-
correction term is negative and other policy variables included in vector Z (eg, the rate of
6Engle and Granger (1987).
-9 -
domestic credit expansion net of real GDP growth) do not offset the effects of the nominal
devaluation.
3. AN APPLICATION TO TEE CHILEAN CASE
In this section the model presented above is estimated for the period 1960-1992, using
annual observations of the corrected real exchange rata (see Figure 1), calculated by CIEPLAN.
The series is an altemative measure to the official RER calculated by the Central Bank, that takes
into account measurement problems with the official CPI during the 1972-74 period (when most
transactions were undertaken at black-market prices) and 1976-78 (when the official price index
presents methodological miscalculations).
All variables were tested to verify whether they can be represented more appropriately
as difference or trend stationary processes. Table A.2 in the Appendix collects the results of
applying unit-root tests to the data. It is apparent that with the only exception of the short-term
capital inflows all fundamentals present evidence of non-stationarity. Rejection of the unit-root
hypothesis for the first difference of the vanables ensures that we are dealing with integrated
processes of first order, I(l). It has been frequently argued that standard unit-root tests are
sensitive to the presence of stuctural breaks; a stationary variable affected by breaks can easily
"mimic" non-stationary patens and the ADF test has been proved quite sensitive to this problem
(Perron, 1989). In the Chilean case this may-be an -important issue because of the economic
reform package applied from 1975 onwards. To overcome this limitation we use a procedure
suggested by Perron (1989), which modifies the conventional ADF test by introducing a set of
dummy variables which control for the presence of breaks. The results, presented in the last two
columns of Table A.2, confiLrm that none of the fundamentals are stationary processes.
'See Corit;ar and Marshall (1980).
Figure 1
REAL EXCHANGE RATE AND NET CAPITAL INFLOWS
IN CHILE: 19601992
Net Capital Inflows
-- ~~~~~~10
\140 T
00 4
120 - 7- 0
1 00
Real Exchange Rate a'
60-1
40 14 1(. 1i~I I.* .2[1t'1.t- ~IIII20
1960 1963 1966 1969. 192 1975198 1981 1964 1987' 1990
- 11 -
_.t il-
3.1 The Long-Run Cointegrated Equilibrium.
Once confirmed that the variables behave as integrated processes, tests for cointegration
can be undertaken. We use the two-step procedure for estimating cointegration-error correction
models suggested by Engle and Granger (1987). In the first step the cointegrating regression is
estimated by ordinary least squares; its errors are used in the second step to estimate the error-
correction mechanism and the short-term dynamic model. Despite evidence that this procedure
may be non-optimal because of the presence of nuisance parameters (Campbell and Perron,
1991), we rely on it because of two facts: (a) as discussed below, in this particular case the
estimation is likely to be free of nuisance parameters and, (b) the sample, though spanning a
long-run horizon, is of low frequency (annual data) so that an alternative non-linear estmation
may yield inconsistent results (Phillips, 1983).
Two conditions ensure that the OLS estimation of the cointegration regression is
asymptotically optimal: errors should be non-corelated and right-hand side variables should not
be Granger-caused 'y left-hand side variables (Phillips and Loretan, 1991). The results in Table
I show that none of the cointegating equations present evidence of serial correlation of any
order, and also that errors are staionary. The results in Table A.3 in the appendix also show th,
excepting government expenditure, the RER does not Granger-cause any of the fundamentals?
With regards to the causality between the RER and capital flows, the tests suggest that the long-
run causality among these variables would be as that suggested by Edwards (1988). We
acknowledge, however, that causality can change as a result of policy shocks and- other breaks;
the small number of observations available, however, preclude us from making a formal testing.9
The above considerations allow us to estimate directly the cointegration regression. The
results, presented in Table 1, strongly corroborate the theoretical model outlined in section 2, thus
permitting the interpretation of equation (7) as the long-run equilibrium relationship. As a first
'Note that tough causality can appear as a spurious result, absence of causality is never a spurious result
(Granger and Newbold. 1974); finding absence of feed-back effects supports the notion that the first step is free of
nuisance parameters.
'Estimating Granger test in partitions of the samplc feg, 1960-1974 and 1975-1992) did not alter the basic results
of the test, but the reduced number of observations in each sub-sample limits the confidenoe on the inference.
- 12 -
TABLE 1
Estmation Results of the Caintegration Equatins
1960.1992
Exiended Rnal
Model Model
Constant 4.92 4.B1
(18.1) (18.4)
Log of Terms of Trade -0.12 -0.11
(-2.53) (-2.42)
Openness -1.09 -1.11
(-12.4) (-1.8)
Log Government Expendures 0.31 0.31
(% of GDP) (2.79) (2.74)
Log Public hwestment -0.15 -0.13
(% of GOP) (-2.38) (-2.37)
Long Term Capital Inflows 0.89 0.97
(% of GDP) (2.50) (2.79)
Foreign Direct Investment -1.67
(% d GDP) (-1.11)
Porfolio Investnent -1-03
(% of GDP) (-0.75)
Dummy -026 -023
1971-1973 = 1 (-3.67) (-3.63)
R2 0.96 0.96
Durbin-Watson stat 1.52 1.46
Box-Pierce 0 test 12.1 12.9
ADF test on residuals -4.19 -4.03
Note: () Long-term capital flows indudes foreign direct nvestnent in te final modal.
Critical values for the ADF test on the residuals are 4.15 and -4.90 at 5 and 1%, respectiel.
- 13 -
approach to modelling the data we separate capital inflows among portfolio investment, foreign
direct investment and long-term capital inflows. Results are conclusive that the first two
components convey no information whatsoever for the estimation of the RER, because when
tested separated or jointly yields statistically non-significant parameters. Table AA in the
appendix shows that these results are not due to colinearity among these variables, since their
contemporaneous correlation do not reach 0.35 for any pair of them. Moreover, since there is a
presumption that the breakdown between the two latter may be to some extent inaccurate
(because of financial funds fungibility) we deemed reasonable to use an aggregated measure in
the final model.
One of the most interesting fmdings in Table 1 is that of the importance of the volume
of trade (degree of openness) in determining the level of the RER. The negative and significant
sign supports the notion that reforms aimed at reducing tariffs and eliminating trade restrictions
are consistent. with a more depreciated RER. In the case of the Chilean reforms, tariffs were
reduced from a high 80% average during the 1960-1974 period to a low level of 20% in the
1975-92 perod; subsequently the volume of trade increased from 25% to 55% of the GDP. With
an elasticity of the RER to openness which clusters around 1, three-quarters of the 45%
depreciation of the RER can be linked to the increase in trade volume. This result is consistent
with previous research and in particular with ongoing parallel research by Quiroz and Chumacero
(1993), which by means of a simulated real-business cycle model conclude that the decline on
tariff, at a minimurm, depreciated the RER in the order of 40%.
The results for the ratio of government expenditures to GDP show a positive elasticity,
implying that fiscal spending tends to concentrate more on non-traded goods compared to the
private sector and that, consequently, unsustainable government deficits lead to exchange rate
overvaluation. The small magnitude of the effects points to the fact that substantial public saving
is required to sustain a high RER in the presence of capital infl6ws. The last years witnessed a
bitter discussion among Chilean economists on this issue (see Arrau et al 1992) as increasing
capital inflows called for an appreciation of the Chilean peso at the cost of reducing export
competitiveness. Our result adds to mounting evidence that, in order to provide a sustainable high
RER in an efficient way, measures outside the fiscal ara should be used. Moreover, the
-14-
composition of govermnent expenditures also matters.'0 The significant coefficient of the ratio
of public investment to GDP suggests, as expected, that government capital expenditures
concentrate in traded goods.
The sign of capital inflows is, as expected, positive and significant implying that an
increase in foreign exchange appreciates the real exchange rate. The magnitude of the estimated
elasticities 'in the range of 1- suggests that the effects are quite strong and, again, raises doubts
on the ability of the authority to sustain a real exchange rate above the long-term equiibrium by
altering its policy mix. The data in Table A.1 show that longterm capital inflows increased from
-10% of GDP in the 1983-87 period to 4% in the 1989-92 period. Other things constant, this
change in the capital account of the balance of payments accounts for an appreciation of the RER
of about 15%_
The effects of shocks to the terms of trade, as remarked in Section 2, are theoretically
ambiguous. The negative sign obtained suggests, contrary to conventional results, the dominance
of substitution over income effects Two non-exclusive explanations can be suggested for this
phenomenon: (a) it is likely that in this regressions TOT captures only substitution effects in the
demand for traded goods, because income effects are channeled through the expansion of trade
volumes -directly captured in the degree of openness- andlor in the increase of sustainable long-
run capital inflows; (b) a more circumvoluted explanation suggests that if wages are indexed
backwards -as it was the practice for most of the 1960-1992 period- and foreign demand expands.
the increase in exports and aggregate demand would induce a rise in prices (or inflation) which,
in tunm, implies a reduction in current real wages. The cut in real salaries allows the supply of
non-tradables to increase, thus reducing the RER. Schmidt-Hebbel and Serven (1994) found a
similar behavior when simulating an intertemporal rational expectations model for the Chilean
economy. Note that the negative sign has also been found in other two studies. Valdas et al
(1990) found a negative effect for TOT in a model which controlled indirecty for income effects.
Repetto (1992) also found a negative coefficient when estimating, a RER equation which
included, among other variables, capital inflows in the specification. The size of the coefficient
in the latter was, however, twice as large as in our case (see Appendix Table A5).
m However. in the Chilean case Elbadawi (1993) found these effects to be negligible.
- 15-
Finally, a dummy variable was introduced to capture the severe disarray in the economy
during the 1971-1973 period, in which an excessive expansion of domestic credit was
accompanied by drastic pnce and currency controls and an increasingly distorted foreign trade
structure.
This estimated cointegration equation is used below to estimate the short-term dynamic
models. However, prior to discussing the results on the ECMs it is important to note that the
cointegaion estimation of the corrected RER is remarkably stable along the 30-year period.
Standard stability tests (Cusum and Cusum of squares) as well as the recursive estimation of
residuals and parameters show little evidence of instability (see Figures Al to A7 in the
appendix). The former suggest no evidence of structural breaks, not even in 1975, which points
to the fact that the cointegrating vector accounts for the break in the series unveiled by unit-root
tests.
3.2 The Error-Correction Model.
To perform the estmation of the short-run model of the RER we follow the methodology
suggested by Phillips and Loretan (1991). Its main diffemnce with standard ECM specification
is that it includes leads of the right-hand side variables to capture the presence of potential feed-
back effects from the RER to the fundamentals. Drawing from our previous results on causality
we test those variables in which there was some evidence of two-way causality, i.e, nominal
exchange rate movements and government expenditure. Capital inflows are also included as a
double check on our previous results on causality test.
The results reveal a wealth of dynamic effects that were missing in staic studies and that
help sharpening our theoretical predictions. Fist, note that leads of the fundamntals are not
significant, consistent with prior evidence on causality which suggested the absence of feed-back
effects.'-" In addition, short-term capital inflows -which proved to have no effect in the long-run-
are quite important in the short-mn. Second, note that the size of the coefficient of openness in
"Non-significant lags and leads wce sequentially deleted but the results are not affected by the ordering of
deletion because colinearity among variables is small.
S
- 16 -
the short-run model does not differ markedly from that of the long-run model, implying that the
markets internalize the effects of increased openness rather quickly (within one year). On the
contrary, only half of the change in government expenditure has direct effects on the RER,
creating a dynamic pattem of adjustment toward the equilibrium RER. Note that while static
models can capture the former relationship, they miss the implicit dynamics of shocks which,
even when having small direct effects, tend to build an important long-run effect.
The most interesting result concerns the effects of nominal devaluations on the RER- As
anticipated by causality tests, feedback effects between the two variables were likely to exisL The
estimated parameter for anticipated devaluations is positive, consistent with rational expectations
models of the current account balance (see Obstfeld, 1985). On the other hand, the
contemporaneous effect is negative, which is consistent with previous empirical literature
(Edwards, 1985). The aggregated effect, nevertheless, recovers the superneutrality of monetary
models, i.e., that monetaxy shocks do not have effects on the rate of change of real variables (like
the RER) because the latter effect offset the previous negative effectL'
A crucial parameter in the estimation of ECM is, naually, that associated with the error-
correction term. As mentioned, it measures the degree of adjustment of the actual RER with
regards to its equilibnum leveL While the estimates of the speed of adjustment in Table 2 are
smaller than the 0.78 estimated by Elbadawi (1993) for Chile using a similar framework, our
estimates are much larger than the 0.19 obtained by Edwards (1989) for a group of developing
countres using a partial adjustment model. Note that Edwards estimates suggest that very tittle
adjustment actually takes place and, furthermore, that the adjustment may take an extremely long
period to complete. The comparison also shows how different results can be when a dynamic
specification is proposed and tested, instead of assuming (ex-antej a partial adjustment model.'3
The error-corection coefficients can be manipulated, in the context of the error-correction
specification, to derive the corresponding adjustment speed in terms of the number of years
12Ajoint Wald test cannot reject the null hypothesis that the sum of both coefficients is zero at 95% confidence.
3 An adjustment parameter of 0.19 implies that a shock dissipates in about 30 years.
- 17 -
TABLE 2
Estimation Results of the Error-Correcton Models
1960-1992
General Model Final Model
Error Correction Term 0.50 0.56
(speed of adjustment) (2.72) (3.43)
A Opemess -0.88 -0.99
(-6.77) (-9.15)
A Log Gayv Expenditures 0.21 0.17
of GDP) (2.49) (2.80)
A Lead Log Gov. Expenditures 0.17
(% of GDP) (1.77)
A Long Term Capital Inflows 0.60 0.51
(% of GDP) (2.31) (1.82)
Lead A Long Term Capitl Inflofis 0.22
(% of GDP) (0.77)
AShort Term Capital lnflows 0.54 0.63
(% of GDP) (2.60) (3.69)
Nominal Devaluation -0.10 -0.14
(-1.14) (- 903
Lead Nominal Devaluation 0.09 0.13
(1.27) (2.02)
A Log Pubrfc InvestMent -0.11 -0.12
(% of GDP) (-1.48) (-2.00)
A Dummy 1971-73 0.20 0.22
(4.81) (6.44)
Dummy 1979 0.10 -021
(2.58) (-2.84)
Adjusted R2 0.85 0.85
Durbin-Watson stat 200 2.23
0(7) 13 6.3
Note: Dummy 1979 lkes value 1 in at year and 0 otherwise.
- 18 -
required to eliminate a given exogenous shock. According to our calculations it would takes
around 1 year to eliminate 50% of the shock and 5 years to clear 99.9% of it."
4. RER AND ERER INDICES FOR CHILE.
The estimated cointegration equation can be used also to compute the equilibrium real
exchange rate, ERER, which is determined by the "sustainable" or "permanent" values of the
fundamentals. The computation is not straightforward, however, because fundamentals are
integrated processes, i.e., their fluctuations correspond to a combination of permanent and
transitory shocks, of which only the former are of interest when computing the ERER. To
disentangle permanent and transitory shocks we use the Beveridge and Nelson (1981)
decomposition method, which generates a measure of the permanent component as the gain
function of the innovations of an ARIMA model (Table A.5 in the appendix presents details of
the time-series estimation). Of the fimdamentals of the RER, the TOT and public investment can
be characterized as a random-walk process, for which all innovations are permanent. In the rest
of the cases, the gain function is less than 1, implying that only a fraction of each shock remains
in the long-run.
It should be noted that this. decomposition, which yield a unique dy-namic path for each
fundamental, does not provide a unique solution for the ERER in terms of the intercept.'5 In order
to emphasize the importance of the extemal balance for this analysis, we normalize the ERER
index according to a resource-balance criterion. Hence, we scale the ERER index so that its
average is equal to the average of the actual RER over the years in which the resource balance
is 'close' to its equilibrium level.'6 Table 4 and Figures 2 and 3 present the estimated equilibrium
'4Adjustment periods were calculated as: (I+a)' = (1+4), where t is thc&numler of periods, 8o is he error-
correction coefficient and a = 05 and 0.99.
* ' This is because the mtional expectations solution for the ERER is not unique. If we assume the unknown
ERER function to be given by g*(x) and the corresponding rational expectations solution to be given by a general
Taylor approximation g(x/O) (which is assumed to approximate g*(x) fairly closely); then, using the regression on
the observed RER: y = g(x/O) +F e to estimate g(x/0) by IO'/O) = g(x/8) does not guarantee that (x&/) and g(xIB)
are equal for each point x in the space of the fundamentals (see Elbadawi (1983) on the validity of the Taylor series
intepretation of the regression estimators).
16"The resource balance is dubbed 'close' to the equilibrium if it is positive.
- 19 -
RER computed with the corrected RER cointegration equation and the estimated permanent
component of the fundamentals; the corresponding RER misalignment is calculated as:
RER Mia1ig,,,,en = RER - ERER
ERER
Our estimates agree with those of Edwards (1987) and Elbadawi (1993) in that the ERER
show some variability. It follows that at least part of the observed RER variability is related to
equilibrium behavior, and that analyses of real exchange rate misalignment based on historical
comparisons of observed RER levels (i.e. the PPP approach) may lead to erroneous conclusions.
The figures show a remarkable success on the part of the computed index in reproducing
well known overvaluation (and undervaluation) episodes of the recent macroeconomic history of
Chile. In particular, note that misalignment is first low but stable in the 1964-1970 periodL During
the administration of Dr. Allende, which expanded markedly both fiscal expenditures and
domestic credit (see Table A.1), misalignment increased to a high level of 28% and remained
quite high even after the coup d'etat of 1973. The reform process started in 1975 not only
brought the RER to its equilibrium leveeL but amounted also to a real depreciation of 25% for
the period 1975-1978, compared to the previous 15 years- The fixing of the nominal exchange
rate in 1979 and the massive flow of foreign borrowing induced a wave of wide misalignment
in the RER (which peaked at 21% in 1980-81 period). Despite important nominal devaluations
in 1982, the year of the debt crisis, macroeconomic mismanagement did not bring much relief
to the burden of the RER in the foUowing years, until the Buchi administration came up with a
high-real-exchange-rate macroeconomic proposal in 1986. The high real exchange rate policy
reversed the chronic tendency towards appreciated RER, but as foreign capital retumed to flow
to Chile it became increasingly costly for the Cental Bank to sustain it. Despite measures to
allow outflows of capital, the large volume of capital inflows continued to suggest a more
appreciated equilibrium RER. In this context. the nominal revaluation of 5% in 1992, though a
signal in the conrect direction, seems to be insufficient to align the RER during 1992.
20-
Figure 2
REAL EXCHANGE RATE
ACTUAL AND EQUILIBRIUM VALUES
180 -
160
140 -
120i "-
80 - -
60 '
40 i
1964 1961 i970 1973 1976 1979 198;: 1985198B I1991
| - - EQUILIBRiUM -ACTUAL
Figure 3
REAL EXCHANGE RATE
MISALIGNMENT
40%-
o0-Govenmntj e C
30%- - surategy
20%tj;
-20% /0- Import SubstitUion _ Trade Fixed Nominal
Strategy Liberarza*on Exchange Rate
-30%1 1 17 1 , , , , , ,
1' 64 i961 1970, 1973 1976 1 i982 t II986 I i98910
- 21 -
TABLE 3
Equilibrium Real Exchange Rate and Misalignment
1965-1992
Corrected Real Exchange Rate Misaligrnent
Actual Equilibium (percet)
1965 140.0 129.3 8.3
1966 133.6 115.3 15.9
1967 127.1 1182 7.5
1968 118.0 1112 62
1969 112.0 98.9 13.3
1970 111.4 108.8 2A
1971 1212 122.2 -0.8
1972 162.0 131.8 229
1973 132.8 103.9 27.9
1974 853 70.1 21.7
1975 62.1 612 1.6
1976 77.1 74.5 4.0
1977 92.7 83.5 11.1
1978 84.5 89.8 -5.9
1979 85A 80.5 6.1
1980 100.0 825 21.2
1981 121.5 102.8 182
1982 109.9 103.7 6.0
1983 89.5 83.7 6.9
1984 87.9 79.0 11.3
1985 70.9 66.0 7.3
1986 59.9 57.6 3.9
1987 55.6 54.9 12
1988 52.0 54.6 -4.7
1989 532 54.3 -2.1
1990 51.7 59.0 -12.4
1991 53.3 60.3 -11.6
1992 56.3 61.7 -c.7
Ncte: Tie- misaligrnent is calculated as (RFR-ERER)JERER.-
- 22 -
5. CONCLUSIONS
The real exchange rate has been at the heart of the openness-oriented reforms in Latin
America. It has been argued that highly competitive real exchange rates have driven trade reforns
and made Latin American products attractive in world markets (World Bank, 1993). It is not
surprising, therefore, that the recent massive capital inflows into Latin America and the
subsequent real appreciation of their currencies have generated considerable consternation for
policy makers and political leaders, as well as concern among economists and experts."'
In Chile, the debate on the role of capital flows started in the early 1980s, at the onset
of the debt crisis. Issues related to the role of capital flows in disrupting macroeconomic
management -including the direction of causation between real exchange rates and capital flows-,
the extent to which capital flows are sustainable and hence whether or not their influence on the
real exchange rate is consistent with equilibrium behavior, have been the subject of much
controversy.
This paper contributes to the debate by estimating the cointegrating long-run equilibrium
path between the RER and capital flows, among other fundamentals. The cointegration model
allows a re-interpretation of static estimates of the equilibrum real exchange rate (ERER) model
to be consistent with long-run forward-looking behavior and flexible short-run dynamics
(Elbadawi, 1993). Furthermore, stochastic non-stationarity provides an empirical measure to the
concept of "sustainability" of fundamentals. The estimation of the long-run cointegration
equilibrium equation of the ERER and the corresponding dynamiic error-correction specification,
strongly corroborates the theoretical model and improves the results of previous studies.
Among the components of the capital account, our results suggest that only long-term
capital flows and direct foreign investment are cointegrated with long-term ERER, with an
elasticity clustering around 1. Short-run capital flows, on the other hand, were found to have
influence on the RER in the short-run only. This findings agree with the notion that if capital
flows are regarded to be genuinely long-term, their effect on the real exchange rate is a true
equilibrium phenomenon, and in this case no policy action will be required. The rather
'7 See Calvo, Leiderman and Reinhart (1993) for an exposition of the debate.
- 23 -
appreciable effect estimated for capital inflows is in sharp contrast with the positive but small
estimated effect for the ratio of govemment expenditure to GDP. The latter implies that fiscal
spending tends to concentrate on non-traded goods compared to the private sector and that,
consequently, unsustainable government deficits lead to exchange rate overvaluation. However,
the comparison of the two effects suggests that, sterilizing the appreciating effects of capital
inflows would require significant and sustained fiscal retrenchment.
In terms of the long-run effects of other fundamentals, the estimated elasticity of the
volume of trade (degree of openness) is the most interesting. The result supports the notion that
trade liberalization requires a more depreciated ERER It also corroborates the view that without
a significant real depreciation, Chile's trade liberalization could have been difficult to sustain.
With an estimated elasticity of the RER to openness around 1, we calculate that three quarters
of the 45% depreciation of the RER can be linked to the increase in trade volume. This finding
is consistent with parallel research by Quiroz and Chumacero (1993) which, using an entirely
different methodology, estimate that the decline in tariffs accounts for a depreciation of the ERER
of 40%.
The results of the dynamic error-correction model reveal a wealth of information that was
missing in previous studies and that help sharpen our theoretical predictions. Consistent with the
empirical literature, we obtained a negative contemporaneous effect of nominal exchange rate
devaluations on the RER (Edwards, 1989). On the other hand, the estimated positive parameter
for anticipated devaluations corroborates the predictions of the rational expectations models of
the current account balance (Obstfeld, 1983). In addition, the aggregated nuU effect recovers the
supemeutrality of monetary models.
In addition to providing estimates of the order of magnitudes of influence of capital flows
and other fundamentals, our approach also allows computing indices for the ERER and RER
misalignment Using proxies for the "sustainable" path of tde fiunamentals -suggested by their
underlying data generating processes- and subject to a sensible normalization rule, the estimated
long-run equation was used to derive indices of the ERER. The estimated RER index and the
corresponding rate of RER misaligmnent are successful in reproducing the salient episodes and
characteristics of the recent macroeconomic history of Chile.
- 24-
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(1987), "Tariffs, Terns of Trade and Real Exchange Rate in an Intertemporal Model of
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(1988), "El Monetarismo en Chile, 1973-1983: Algunos dilemas economicos", in F.
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September.
TABLE All
SELECTED MACROECONOMIC INDICATORS: 1960-1992
Rowl Exchange Rala Government Domestic Gross Fixed Openes Nominal Long-tarm Foreign Direcl Ponrtolio shon-term
OfficCal Corr,eced Expendilures Credil Investment ExchangeR a Capital lows Investment Investment Capital nows
(lUIOmtOO) (IlIOulOC) (% of GDP) (% ofGOP) I% ofGDP) (% ofGCP) (ISUS$11) (%ofIGOP) (h ofGOP) (% of GOP) (%of GOP)
1960 306.0 133.3 22,9 23.4 20.7 30,6 0.00110 0.6 0.7 -0.2 -0.6
1961 287.6 144.4 23.1 19.9 20.0 28.4 0.00114 2.5 1.2 -0.2 -0.9
1962 289,5 160,0 24.7 26.7 21.4 25.4 0.00162 3.0 0.8 -0.1 0.5
1963 302.9 132.1 23.1 15.0 23.1 28.2 0.00213 36 -0.6 -0.1 -2.6
1964 302,9 147.1 22.1 20.9 21.4 26.6 0.00278 3.1 -0.2 -0.1 -2.2
1965 294.3 140.0 25.4 21.7 19.9 27.3 0,00374 2.7 .0.6 -0.1 -0.5
1966 276.4 133.6 25.2 20.3 18.5 29.9 0,00437 1.9 -0.6 -0.1 0.4
1967 261.7 127.1 23.2 21.5 18,3 28.6 0.00579 1.1 0.0 -0.2 0,4
1968 244.0 118.0 20.2 19.5 19.3 29,2 0.00767 1,7 2.2 -0.1 0.8
1969 234.1 112.0 21.3 16.7 19.6 32,8 0.00998 2.8 1.2 .0.1 0.3
1970 223 2 111-4 22.9 17.2 20.4 31.3 0.01223 3.0 -1.0 -0. 1 -0.3
1971 242.7 121.2 27.6 31.1 18.3 22.7 0.01580 -0.4 -0.7 -0.1 0,6
1972 264.2 162,0 29.7 42.9 14.8 22.0 0.02500 0.9 -0.0 -0.0 0.3
1973 398.2 132.8 42.3 63.6 14.7 29,6 0.36000 -0.5 -0.0 -0.1 0.7
1974 136.1 85.3 25.0 43.6 17.4 40.2 1.8720 5.3 .5.0 -0.1 .0.5
1975 103.7 62.1 21.0 58.0 15.4 52.9 8,5000 -1.7 0,7 -0.1 1,1
1978 99,0 77.1 18.7 43.6 12.7 45.9 17.420 0.5 -0,0 -0.1 1.5
1977 106.4 92.7 19.8 43.8 13.3 43.0 27.960 0.3 0.1 -0.1 3.8
1978 88.1 84.5 19.4 41.0 14.5 44.5 33.950 8.7 1.1 0.0 2.8
1979 87.4 85.4 19.6 41.0 15,6 49.4 39,000 8.8 1.1 0.2 2.3
1980 100.0 100.0 20.2 44.6 17R8 49,8 39,000 7,5 0.8 -0.2 3.3
1981 118.1 121.5 23.0 51.0 19.5 43.2 39.000 9.9 1.2 -0.1 3.4
1982 109.5 109.9 28,1 87.9 15.0 40.6 73.430 5.3 1.8 -0.1 -2.7
1983 87.7 89.5 25,7 87.4 12.9 48.4 87.530 -7.1 0.7 -0.0 -10.0
1984 87.0 87.9 28.6 108.8 13.2 49.6 128.24 -4.2 0.4 -0.1 3.0
1985 70.4 70.9 30.4 118.9 14.8 55.4 183.88 -11.3 0.7 0.2 2.5
1988 63.3 59.9 27.9 114.4 15.0 57.4 204.73 -17.3 0.7 1.2 4.5
1987 60.2 55.8 28.1 107.1 16.5 62.9 238.14 -9.7 1.2 3.7 -0.7
1988 64.7 52.0 26.3 92.5 17.0 67.5 247.20 -1.0 0, 3.9 -0.3
1989 62.8 53.2 21.6 82.2 17.9 72.1 297.37 -3,7 0.7 5.5 2.4
1990 51.4 51.7 22.8 79.1 18.8 70.2 337.09 3.8 0.9 2.8 4,3
1991 53 0 533 21 5 70.0 17.5 68.8 374.51 1.4 1.8 0.2 *0.5
1992 58.1 563 20 5 668 18.6 64.7 380.22 2.4 0.9 0,9 2.8
Sources: Col. (1), (3) to (6) Central BSank of Chile,
Col. (7) to (12) IMF (IFS)
Col. (2) CIEPLAN and IMF (IFS)
- 28 -
Table A.2
Order of fntegraton Tests
1960-1992
Vadable A.D.F. Test AD.F. Test Perron Test Perrn Test
Level First Difference Level First Difference
Real Exchange Rate -322 (4) -5.36 [1) -2.02 (2) -5.76 (0)
Terms of Trade -3.16 (4) -5.20 (0) - -
Openness" -0.69 (1) -520 (2) -93 (2) -5.71 (2)
Govemment Expenddiures (% of GDP) -3.19 (2) -5.60 (0) -
Long-Term Capital Inflows (% of GDP) -2.41 (0) -627 (1)
Portforio Inves:rnent (% of GDP) -3.16 (1) 4.30 (0) - -
Foreign Direct lImesltent (% of GOP) -2.64 (4) -6.44 (1) - -
Short-Term Capital Inflows (% of GDP) 4.38 (0) - -
Public Investment (% of GDP) -D.79 (1) -5.04 (1) - -
Nominal Devaluaions (%) -262 (1) -6.54 (0) -5.66 (0) -
Crftical Values - at 5% -3.56 -2.97 -3.60 -3.60
- at 10% -323 -2.63 -3.35 -3.35
otes 'Numbers of lags in parenthesis, n Tests for the presence of a strucirta brek in 1974-75.
Table A.3
Granger Causarly Tests
1960-1992
Null Hypothesis: Null Hypothesis:
RER does not Granger- RER is not
cause: Granger-caused by:
Openness 0-43 2.48k
Govremment Expenditure (% GiDP) 3.01 17.8"
Termns of Trade 1.15 2.63*
Long-Tern Capital Inflows (% of GDP) 0.86 2.66'
Public lnvestrnent (% of GDP) 0.05 329"
Short-Term Capital Inflows (% of GDP) 1.75 0.67
Portorio Investment r% of GDP) 1.40 0.45
Note: ) Signficaant at 10%, (") sinficant at 5%.
-29 -
Table A.4
Correlation Matrix oa Capital Inflow Components
1960-1992
Capital Inflows wet Irvestment LnTa aia nlw
Shod-t-ermn Capita Inflows -0.05 0Q12 0.16
Porfoibo Inmestment - -0.10 -0.33
Foreign Direct Investment -- 0.16
Long-Term Capital Infiows -
TABLE AS
Previous Es4imates of RER equaions.
Effects an fthe lon-n RER of an increase of 1% in:
Govemment Terms of Capital Period
Expenditures Trade Inflows
Corbo (1 985. quartely data) 0.21 - - 1977-1983
Valdes, Muchnic and Hurtado (1990) 0.23 -029' - 1960-1982
Marshall and Schmidt-Hebbel (1991) 1 .09 014 - 1960-1988
Arrau et al (1992 quartery data) 0.8 -1.1 - - 1977-1991
Repetto (1992) 0.30- -0.30 0.016 1960-1990
Elbadawi (1993) 0.85 0.29 - 1965-1990
Quiroz and Chumacero (1993) 0.41 0.25 - 1960-1988
Edwards (1989, quartely data) - - 0.15 1977-1981
Edwards (1989) 0.30 0.04 - 1962-1984
(panel for 12 Latinamerican countries)
Note: C) An offsetting parameter of 0.27 was also found in this estimate.
Parameter non-signiftcant at 5%.
- 30 -
TABLE A.6
Estimed ARIMA models for fundamentas
1960.1992
l~~~~~~~~~~~~~~~~~~~~R Qp7)
(1 -0.45L+O.48LALog[Openj=0.02+4
-2-70 [(245) 05 55
I - O.45L+ 0.48L) A Log [Gov. ExpJ = (1+1.78L-0.812) + E
(3.40) (5.01) - (320) (-1.62) 0.7 7.42
(I - 0.14L0 + 078L5 A Log[ gcapj = (1 - O.35L)
(478) (4.07) (1.75) 0.40 4i58
-31-
Figures A.1 and A.2
Stciity Tests Cusumrn
015-
50-
- ~ ~ ~ ~ ~ uu 5~ --~
1.2
0.7
0.D -----
0.25 -- -~~~~~~~~. --
0.D0
72 74 76 78 80 82 84868890 92
- CUS- U .s - SX 9Sificnce
Stdity Tes-ts: Cusun of Swwxes
0.75 --
0.25- . . .- . I . S I * . * . * , * . .
72 4 76 78 80E 828 68 0 92
-CuSuI of squoes --5 s9nfirconc
-32-
Figures A.3 and A.4
Recusive Estimation of the Coefficient of Opens
0.00-
-0.25
-0.50-
-0.75 -
'LOU -----
-1 25 -
-t50
1976 1978 180 1982 1984 186 S9W8 1990 1992
Recusive Estiation -+- 2 SE.
Recursnve Estimation of the Coefficient of Goverrnent Expditires
0.7-
0.6.
06 , ~~~~~~~~~~~~~---------
05
0.4
0.3
0.2 - ---
0.1 - ---- ---------
007681978 1980 19B2 184 186 8B 1M90 1992
Recursive Estirotion +- 2 SE
-33-
Figures A.5 and A.6
Recu-sive Estimation of the Coefficient of Capital mnfmlw
0.05-
0.04 .
0.024
0.01 ', -- ---- -------- .
0.00- ,
1976 1978 980 1982 1984 1986 1988 1990 19
_ Rec.rve Estimation 2 SE.
Rearsive Esition of te Coefficent of hiestrnet
0.02-.
*sS1 , 8 *--- --------,
O0.0 . --------
-0.01 --
zm ~~~--- -------~ ~~ ----~--.--
--.04 N, ,- - - ;.> .
1976 1978 1980 1982 1984 1986 1968 1990 1992
-Recusive Estimotion +- 2 SE.
-34-
Figure A.7
Recusive Estiotion of the Coefficent of Trrns of Trade
0.05
0.00 --
. ,,,. ~ ~~-------------.--
-0.05
-0.20 -.7. -- -
-025!
176 1978 19019884 90 1992
=Reasive Estirno n 2 SEt
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