Interest Rates in Mexico The Role of Exchange Rate Expectations and International Creditworthiness

The link is explored between interest rates on domestic financial assets in Mexico and expectations of exchange rate changes and perceptions about default risks contained in Mexico’;s external debt. Interest rate differentials between peso- and U.S. dollar-denominated assets are shown to have reflected concerns about the exchange rate policy during the period considered. The evidence also suggests that the interest rate on a U.S. dollar-denominated Mexican domestic asset is linked to the yield implicit in the secondary market price for external debt issued by Mexico. [JEL C12, F31. F34, O54]

Abstract

The link is explored between interest rates on domestic financial assets in Mexico and expectations of exchange rate changes and perceptions about default risks contained in Mexico’;s external debt. Interest rate differentials between peso- and U.S. dollar-denominated assets are shown to have reflected concerns about the exchange rate policy during the period considered. The evidence also suggests that the interest rate on a U.S. dollar-denominated Mexican domestic asset is linked to the yield implicit in the secondary market price for external debt issued by Mexico. [JEL C12, F31. F34, O54]

This paper investigales the recent behavior of interest rales in Mexico by exploring their link to expectations of exchange rate changes and perceptions about the default risk associated with holding Mexican financial assets. A central theme is the extent to which domestic debt and external U.S. dollar-denominated debt issued by Mexico are linked: namely, based on default risk, how closely associated are the interest rate on domestic debt denominated in U.S. dollars and the implicit yield on external debt.

During the period January 1987 to July 1990,1 the Mexican authorities launched a major effort of macroeconomic adjustment and structural reforms, including the liberalization of financial markets. Responding to these efforts, inflation fell sharply from an average of 8 percent a month during 1987 to an average of 1½ percent a month, where it remained from the middle of 1988. However, domestic nominal interest rates fell less steeply; in real terms, ex post interest rates on domestic financial assets, which were negative during 1987, shifted to highly positive subsequently, despite the improvement in overall fiscal performance brought about by the reform efforts of the authorities.

It is in this context that we test the validity of the following hypothesis: the recent behavior of interest rates in Mexico has been closely linked both to expectations of exchange rate changes and to international perceptions of Mexico’;s creditworthiness, which is assumed to be represented by the yield implicit in the secondary market price of external debt issued by Mexico. It is argued that this hypothesis will hold if there are no perceived differences in the credit standing of Mexico’;s domestic and external debt, because, in that case, both the domestic and external debt would be subject to the same country-risk premium.

The methodology adopted to give empirical content to this hypothesis is presented in Section I. Because of some important differences in the characteristics of domestic versus external debt issued by Mexico, the hypothesis is split into two components. The first part concentrates on the role of expectations of exchange rate changes and states that some form of interest rate parity (covered and/or uncovered) should hold between two assets issued by the Mexican government, which are identical in all respects except for the currency of denomination. The second part deals with the country-risk component of Mexican interest rates and states that the interest rate on domestic debt denominated in U.S. dollars should be closely linked to the yield to maturity implicit in the secondary market price issued by Mexico.

Section II tests the first component of the hypothesis. Our investigation shows that when applied to specific financial assets, deviations from the covered interest parity condition (CIP) have been small and random occurrences most of the time. However, significant but brief deviations from CIP have occurred in periods when uncertainties in the economy were unusually large—for instance, between late 1987 and early 1988 when the authorities implemented their comprehensive adjustment program. The results from testing the uncovered interest parity (UIP) hypothesis suggest that the exchange rate policy suffered from a lack of complete credibility (also known as the “peso problem”) during the period under study. Under those circumstances, no strong conclusions can be derived about the validity of the UIP.

Section III uses the observed price in the secondary market for Mexican external debt to obtain the implicit yield associated with holding assets issued by Mexico abroad. This section then uses cointegration techniques to show that domestic interest rates of Mexican assets denominated in U.S. dollars are closely linked to the behavior of the implicit yield derived from the secondary market for Mexican debt. This result, combined with our findings on interest rate parity, leads to an important policy implication, namely, that a sustained decline in domestic interest rates is not only linked to the elimination of the wedge between public expectations of exchange rate devaluations and the preannounced rate of depreciation, but is also associated with an improvement in the underlying conditions in the Mexican economy, which affect international perceptions about the country’;s creditworthiness. Section IV presents some preliminary conclusions.

I Methodology of the Study

The main hypothesis of this study can be expressed as

(1+it)StEt(St+1)=[1+g(itsm)],(1)

where it is the domestic nominal interest at time t on peso-denominated treasury bills; itsm is the implicit yield at time t from the secondary market of Mexico’;s external debt; St is the spot exchange rate at time t, defined as the price of 1 U.S. dollar expressed in Mexican pesos; and Et(St+1) is the expected value at time t of the t + 1 spot exchange rate conditional on information available in time t. The expected rate of depreciation (or appreciation) of the Mexican peso is denoted by Et(St+1)/St.

Equation (1) states that domestic interest rates in Mexico, adjusted for expectations of exchange rate changes, are linked (through a “g” function) to the implicit yield from the secondary market of Mexico’;s external debt. In other words, domestic interest rates in Mexico are closely associated with the international perception of Mexico’;s creditworthiness, which is captured in the behavior of itsm 2 The implicit hypothesis behind this postulate is that there are no perceived differences in the credit standing of domestic and external debt of Mexico, so that both kinds of debt share the same country-risk premium.

Equation (1) would have been a form of UIP between it, and itsm, if g were to take a unitary value. This is very unlikely, however, since the assets involved in equation (1) have very different maturities. Indeed, while the representative Mexican Treasury bills have a maturity of 28 days, commercial bank claims on Mexico, which are traded in the secondary market, have a long-term maturity that runs between 20 and 30 years. Moreover, in contrast to the market for Mexican Treasury bills, the secondary market for Mexico’;s external debt is subject to barriers to entry, because each transaction requires high entry costs and complex documentation. In view of these differences, one cannot expect that domestic interest rates in Mexico, once expectations of exchange rate changes are taken into account, would always equal the implicit yield on Mexico’;s external debt. Instead, it is postulated that such variables “move together,” at least in the long run; that is, that they are cointegrated, where g is the parameter of cointegration.

To test this hypothesis, the following procedure was used. Tests were run to determine, first, whether UIP held between two Mexican assets with identical characteristics except for the currency of denomination: and, second, whether a Mexican asset denominated in U.S. dollars was cointegrated with the implicit yield for Mexico’;s external debt. The rest of this section explains the rationale and the details of this procedure.

Since August 1986, the Mexican authorities have been issuing PAGAFES,3 an instrument denominated in U.S. dollars but payable in pesos at the prevailing controlled exchange rate.4 This asset is identical to the CETES, 5 a peso-denominated treasury bill, except for the currency of denomination. Since both assets are issued by the Mexican Government, they contain the same country-risk premium. The primary interest rates on both PAGAFES and CETES are determined during weekly auctions. Although the Bank of Mexico has intervened at times in order to tighten or ease monetary policy, by and large, interest rates have been freely determined. Furthermore, both assets are freely traded in the secondary markets, although the market for PAGAFES is much thinner. Since the only difference between CETES and PAGAFES is the currency of denomination, if UIP holds, the interest rate differentials between the two assets should reflect only the expected depreciation (or appreciation) of the exchange rate in the controlled market; that is, if UIP holds

(1+it)=(1+it*)Et(St+1)St,(2)

where, from now on, it will refer specifically to the 30-day nominal yield on 28-day nonindexed domestic treasury bill (CETES) at time t; and it* is the 30-day nominal yield on 28-day domestic treasury bill (PAGAFES) at time t.6

Equation (2) is tested in Section II. The test enables one to determine the role of expectations of exchange rate changes on the behavior of the interest rates on CETES. Notice also that the hypothesis contained in equation (1) can be framed in terms of CIP:

(1+it)StFt=[1+g(itsm)],(3)

where Ft is the 30-day forward exchange rate at time t. A CIP version of equation (2) is

(1+it)=(1+it*)FtSt.(4)

Section 11 will explore the validity of both the covered and uncovered versions of interest rate parity; namely, it will test whether equations (2) and (4) held in Mexico during the period under study.

Next, we examine the country-risk component of domestic interest rates. Notice that equations (1) and (2) (or equations (3) and (4)) imply

it*=g(itsm).(5)

Therefore, equation (5) states that the interest rate on PAGAFES is some function of the yield to maturity of Mexico’;s external debt. This equation holds true for either the covered or uncovered version of interest parity.

As mentioned before, the maturities (and some other market characteristics) of Mexico’;s external debt and PAGAFES are very different. However, if the main hypothesis of this paper is true (namely, that after taking expectations of exchange rate changes into account, domestic interest rates are linked to the implicit yield on Mexico’;s external debt), one would expect the two interest rates to “move together,” at least in the long run; that is, we postulate that it* and itsm are cointegrated. Section III presents tests for that hypothesis.

II. Covered and Uncovered Interest Rate Parity

CIP and UIP are alternative hypotheses about the nominal interest rate differentials between financial assets that are identical in all respects except for the currency of denomination. CIP relates this interest rate differential to the forward premium (or discount) on foreign exchange, while UIP relates it to the expected change in the spot exchange rate between the currencies of the two countries over the holding period.7

Covered Interest Parity

Equation (4) represents a CIP relationship between CETES and PAGAFES. Figure 1 shows the evolution of (1 + it) and (1 + it*)(Ft/St) over the period January 1987 to July 1990. The figure shows that deviations from covered interest arbitrage were small for most of the observations, with notable exceptions occurring between November 1987 and March 1988. That period, however, coincided with the beginning of the current stabilization program in Mexico, when a series of structural reforms were announced, a major devaluation took place, and the authorities implemented a price-wage pact between the Government, labor, and business, with the aim of controlling prices. These developments undoubtedly increased uncertainties about the future course of the economy, which, combined with some official intervention in domestic financial markets, might have prevented covered arbitrage from holding.

The deviations from CIP are presented in Table 1. As the table shows, 71 percent of the deviations are less than 0.5 percentage point, and 86 percent of the deviations are less than 1 percentage point. By comparison, the average bid-ask spreads of the exchange rate in the controlled and the free markets during the period were 0.95 percent and 2.1 percent, respectively. Besides being small, the deviations from interest parity have a mean very close to zero and are uncorrected at all lags. Indeed, the results of a Q-test for serial correlation indicate that the null hypothesis that the series is white noise cannot be rejected at the 1 percent significance level.

Figure 1.
Figure 1.

CETES vs. PAGAFES (Adjusted)

Citation: IMF Staff Papers 1991, 003; 10.5089/9781451930801.024.A007

Note: CETES (Certificados de la Tesorería); PAGAFES (Pagares de la Tesorería de la Federación). The unit on the y-axis is 1 + i. where i is the interest rate per month.aAdjusted for the premium in the forward market.

These results indicate that, with the exception of a short period, when uncertainties in the economy were unusually large, deviations from interest rate parity have been small and random occurrences. This evidence seems to provide strong support for CIP.8

Table 1.

Analysis of the Deviations from Interest Rate Parity

(Series: (1 + it) − (1 +it)(Ft/St)

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The null hypothesis of no serial correlation is rejected at the 1 percent level only if the value of x2 is greater than the critical value of 21.7.

Uncovered Interest Parity

The uncovered version of interest parity between CETES and PAGAFES is represented by equation (2).

The empirical literature on UIP has broadly tested equation (2) (see, for example, Cumby and Obstfeld (1980) and Lizondo (1983a)). However, since Et(St+1) is an unobservable variable, the empirical tests have been joint tests of equation (2) and two hypotheses of expectations behavior: (1) the rational expectations hypothesis that states that Et(St+1) is a mathematical conditional expectation, based on the true probability distribution underlying the behavior of the exchange rate; and (2) the hypothesis that the foreign exchange market is “weakly” efficient, in that expectations of the future exchange rate incorporate all information contained in past forecast errors of the exchange rate. These joint hypotheses imply that the forecast error

t=St(1+it)(1+it*)St+1,(6)

should have zero mean and be serially uncorrelated.

The joint hypothesis containing UIP and the hypothesis of weak market efficiency was tested for Mexico using nonoverlapping monthly observations covering the period January 1987–July 1990. Before these results are discussed, it should be pointed out that over this period, the exchange rate was not freely floating, but was managed by the authorities and followed three different regimes: (1) from January 1987 to December 1987 the exchange rate was depreciated by an unspecified amount each day; (2) from January 1988 to December 1988 the exchange rate was fixed except for a small change in February; and (3) from January 1989 to May 1990 the exchange rate was depreciated by an announced 1 peso per U.S. dollar a day.9 Recent studies have analyzed the problems involved in testing the U1P hypothesis in the presence of intervention in exchange rate markets. In particular. Krasker (1980) and Lizondo (1983b) have shown that in the presence of a small and positive probability of a devaluation, an efficient exchange rate market will imply that the expected value of the future spot rate will reflect the probability of that event. However, as long as the devaluation does not take place, the expectation of the future spot rate will consistently overestimate the realized future spot rate. As a result, the forecast error in the exchange market will show a positive bias,10 but this will not be sufficient to reject the joint hypothesis that UIP holds and that the market is weakly efficient.

Table 2 presents an analysis of the forecast error, ∈t. The most important result is that although the Q-test indicates that at the 1 percent significance level we cannot reject the hypothesis that the errors are uncorrelated, the mean of the forecast error is positive.11 These results imply that St[(1 + it)/(l + it*)] overstates the future spot rate. Based on our previous discussion, however, the existence of a positive mean in the forecast errors does not allow us to accept or reject the hypothesis that UIP holds under conditions of weak market efficiency, since the tests for UIP are not appropriate in the context of a peso problem. However, the lack of autocorrelation in the forecast errors seems to indicate that the interest rate differentials between CETES and PAGAFES incorporated all the information available to generate predictions of the future exchange rate in conditions where the peso problem prevailed, namely in a situation where the exchange rate was not allowed to float freely and there was always a small probability of a devaluation.12 Indeed, the differential between the spread in the interest rates on CETES and PAGAFES and the preannounced depreciation of the exchange rate was greatest in the early part of 1988 following the transition to a fixed exchange rate regime, which can be taken as an indication of a lack of full credibility in the exchange rate policy (Figure 2). However, as the authorities persisted in their policies and financial conditions improved, the differential has tended to decline.

Table 2.

Analysis of the Forecast Error

(ϵt=St[(1+it)/(1+it*)]St+1)

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The critical value for the test with 9 degrees of freedom at the 1 percent level is 21.7.

III. Domestic Interest Rates and the Risk of Default in the Secondary Market

In the previous section we showed that the interest rate differentials between domestic nonindexed assets (CETES) and dollar-denominated domestic assets (PAGAFES) have, in general, satisfied the CIP condition. The evidence also indicated the presence of a peso problem, suggesting that the interest rate differentials may be attributable to expectations of large exchange rate changes (which did not. in fact, take place). These results, however, do not explain the persistence of high interest rates on PAGAFES, which ranged between 14 percent and 44 percent a year in the period under study. In the context of a relatively open economy with few restrictions on financial flows, a plausible explanation is that all Mexican assets (indexed or not) contain a risk premium that reflects the market perception of the country’;s credit standing. Therefore, it can be argued that the interest rate differential between PAGAFES and a risk-free asset reflects primarily the country-risk premium.13 Figure 3 shows the recent evolution of the interest rate on PAGAFES versus LIBOR (London interbank offered rate), which can be considered a relatively risk-free interest rate.

Figure 2.
Figure 2.

Interest Rate Differentials vs. Exchange Raie Depreciation

(In percent per month)

Citation: IMF Staff Papers 1991, 003; 10.5089/9781451930801.024.A007

In this section, the hypothesis, as represented in equation (5), is tested that the interest rate on PAGAFES is linked to the yield to maturity implicit in the secondary market price for external debt issued by Mexico. The rationale behind this hypothesis is that, from the point of view of creditworthiness, there should be no distinction between the domestic and external components of the debt.14 Hence, the country-risk premium implicit in the secondary market price for Mexican debt issued abroad should be equal to the country-risk premium contained in domestic debt with identical characteristics.

Figure 3.
Figure 3.

Interest Rates on PAGAFES vs. LIBOR

(In percent per year)

Citation: IMF Staff Papers 1991, 003; 10.5089/9781451930801.024.A007

Note: PAGAFES (Pagares de la Tesorería de la Federación); LIBOR (London interbank offered rate).

As mentioned in Section I, in testing the above hypothesis, one encounters the problem that the characteristics of the assets—that is, external claims on Mexico and PAGAFES—are very different in terms of maturities and market access,15 However, if our hypothesis is correct, one would expect the two interest rates to move together, at least in the long run.

Traditional econometric theory cannot help in testing the hypothesis. A basic assumption underlying most econometric analyses is that the data processes involved are stationary and ergodic. As will be shown below, however, the series for both PAGAFES and the implicit yield for Mexico’;s external debt are integrated processes of order one, I(1), and become stationary only in their first difference. Since the original series are nonstationary, the means and variances of the series are not constants, and the usual statistical properties of convergence to the population mean and variance do not apply. As a result, traditional regression analysis relating the behavior of these variables might just reflect “spurious correlations.” Recent developments in cointegration analysis, however, provided a tool for analyzing whether there is a meaningful close relationship in the long run for variables that are I(1) processes. (See Granger and Weiss (1983) and Engle and Granger (1987).)

Two variables, x and y, following an I(1) process, are said to be cointegrated if there exists a constant. A, such that

zt=xtAyt(7)

is a stationary process; that is, the series zt is integrated of order zero, I(0). An important result of cointegration analysis is that if two variables are I(1) and cointegrated, there must be Granger causality in at least one direction, since one variable can help forecast the other (see Granger (1986)). This corollary will be used in interpreting the results.

In the remainder of this section, we will derive the yield to maturity implicit in the secondary market price of Mexico’;s external debt; we will then test the hypothesis that the interest rate on PAGAFES can be explained by the yield to maturity in the secondary market by determining whether both variables are cointegrated and, if so, what is the direction of the Granger causality. Finally, we present an error-correction model of the short-run dynamics of the adjustment process.

Implicit Yield in the Secondary Market for Mexico’;s External Debt

The implicit yield to maturity for Mexico’;s external debt was obtained from the observed secondary market price, Pt (from Solomon Brothers data),16 and the application of the following present value formula:

Pt=k=1nCt(1+itsm)k+FV(1+itsm)n,(8)

where itsm represents the implicit annual yield to maturity evident in the secondary market price for Mexico’;s external debt in period t. The face value, FV, is set at 100, since the discounts quoted in the secondary markets apply to $100 worth of contractual debt; the contractual coupon payment, Ct, is the interest rate on six-month LIBOR plus the interest rate spread paid by Mexico (assumed to be 13/16 percent over the entire period); and the average maturity, n, was assumed to be equal to 20 years. A monthly series for itsm was constructed covering the period August 1986–July 1990.

In modeling the market value of claims of debtor countries, Dooley (1988) has argued that the secondary market price of debt should equal the creditor’;s expected present value of total debt-service payments. This argument, which requires the assumption that creditors are risk neutral, can be formalized as

Pt=Et[k=1nCt(1+itf)k+FV(1+itf)n],(9)

where Et is the expectations operator during period t, and itf stands for the risk-free market rate (which in this case is taken as corresponding to the rate on six-month LIBOR prevailing at period t).

A comparison between equations (8) and (9) reveals that the implicit yield obtained from the price in the secondary market depends on the creditor’;s expectations about receiving full payments on the contractual debt. The derivation of a specific formula for the probability of default on Mexico’;s external debt is subject to specific assumptions about the distributional properties of debt-service payments.17 In what follows, no attempt is made to derive a measure of the probability of default. Instead, by testing the hypothesis that the long-run behavior of the interest rate on PAGAFES can be explained by the behavior of the implicit yield for Mexico’;s external debt, one is also implicitly testing the hypothesis that the probability of default on Mexico’;s external debt also applies to Mexico’;s domestic debt.18

Test for Cointegration

The first step is to test whether the series on the interest rate on PAGAFES (it*) and on the implicit yield from the secondary market (itsm) are I(0); that is, whether the series are stationary. This is done by using the Dickey-Fuller (DF) and the augmented Dickey-Fuller (ADF) tests. In both tests, the null hypothesis is that the series have a unit root, and the alternative hypothesis is that the series are I(0). The ADF test consists in running the following regressions using ordinary least squares:

Δit*=β1it1*+j=1pγ1.jΔitj*+ω1.t(10)
Δitsm=β2it1sm+j=1qγ2,jΔitjsm+ω2,t,(11)

where the number of lags in each equation (p or q) is selected, such that ω1,t and ω2t are white noise. The difference between the DF test and the ADF test is that in the former, γ1,j = γ2,j = 0.

In both tests, the test statistic is the ratio of each βi (i = 1,2) to its corresponding standard error. The null hypothesis is rejected if the βi is negative and significantly different from zero.19

The results from the tests, which covered the period August 1986–July 1990, are presented in Table 3. In the case of the ADF test, two lags were needed for ω1,t to be white noise (that is, p = 1 in equation (10)), whereas only one lag was enough for ω2,t to be white noise (that is, q = 2 inequation (11)). As shown in the table, the null hypothesis of a unit root cannot be rejected at the 5 percent significance level, indicating that the series on interest rates on PAGAFES and on the implicit yield for Mexico’;s debt are nonstationary processes (Figure 4).

Tabie 3.

Unit Root Test for

it* and itsm

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Note: DF is the Dickey-Fuller test; ADF is the augmented Dickey-Fuller test. The numbers in parentheses indicate the number of lags sufficient for ωi, t (i = 1,2) to be white noise.

Since the test statistics for it* are negative, it can be concluded that if follows an I(l) process. However, since the test statistics for itsm are positive, it is necessary to test for the stationarity of the first difference; that is, Δitsm. Indeed, the test statistics for Δitsm are negative and significant, implying that the original series is I(l). The result that it* and itsm are I(1) is in line with the hypothesis of rational expectations and market efficiency.

As discussed above, given the characteristics of PAGAFES and commercial bank claims on Mexico traded in the secondary market, it is reasonable to expect divergence in the behavior of the variables in the short run; however, we would expect the variables to move together in the long run. To test if the two variables are cointegrated, we first need to form the cointegration equation:

i^t*=Ai^tsm+zt,(12)

where, as mentioned before, zt should be an I(0) process if it* and itsm are cointegrated. In this sense, zt measures the extent to which the system deviates from the long-run relationship between it* and itsm. If zt is stationary, such relationship will hold in the long run.

An estimate of/I. the long-run coefficient relating it* and itsm, is derived from the following vector autoregression (VAR) model:

it*=1.089(8.050)it1*0.325(2.353)it2*+1.184(3.509)it3sm0.978(2.841)it5smR2=0.693.(13)

The numbers in parenthesis are t-values.

Figure 4.
Figure 4.

Interest Rate on PAGAFES vs. Yield to Maturity

(In percent per year)

Citation: IMF Staff Papers 1991, 003; 10.5089/9781451930801.024.A007

Note: PAGAFES (Pagares de la Tesorería de la Federación).

Using equation (13) to obtain the long-run relationship between it* and itsm, the estimated form of equation (11) is

i^t*=0.872itsm,(14)

where it* is the estimated value of it*, and zt = it* − it*.

To test whether zt is an I(0) process, we again used the DF and ADF tests. In addition, we also analyzed the Durbin-Watson of the cointegration equation (CRDW). In the context of the present exercise, this test consists in obtaining the DW statistic by running zt against a constant. The null hypothesis that zt has a unit root will be rejected if the CRDW is significantly above zero.

The results from the tests, which are presented in Table 4, are somewhat mixed. The CRDW test rejects the hypothesis of a unit root in the residuals of the cointegration equation at the 5 percent significance level and therefore supports the hypothesis that it* and itsm are cointegrated. However, neither the DF nor the ADF supports the hypothesis of cointegration at the 5 percent significance level. The DF statistics are significant at the 10 percent level, but the ADF statistics are on the borderline.

Since our results are not conclusive, we further investigate whether it* and itsm are cointegrated by testing for the existence of a generating mechanism having what is called an error-correcting form:20

Table 4.

Tests for Cointegration Between

it* and itsm

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Note: DF is the Dickey-Fuller test; ADF is the augmented Dickey-Fuller test; and CRDW is the Durbin-Watson of the cointegration equation.

The literature reports values ranging from -2.89 (Schwert (1988)) to -3.17 (Hall and Henry (1988)).

Hall and Henry (1988) report a value of -2.84, which probably corresponds to the upper end of the range.

Δit*=ρ1zt1+lagged(Δit*,Δitsm)+d(B)vt,(15)

where dB is a finite polynomial in the lag operator B, and vt is white noise.

Equation (14) is, therefore, a standard error-correction model, which includes the lagged residuals of the cointegration regression. The variables it* and itsm will be cointegrated if ρ1, the coefficient of the lagged error term, is significantly negative. This is so because equation (15) indicates that the amount and direction of a change in it* will take into account the size and sign of the previous deviation from equilibrium zt-1 If the variables are cointegrated, zt is stationary and is therefore inclined to move toward its mean (which is zero), and hence, the equilibrium relationship (14) tends to be restored.

Table 5 presents the results from estimating a simple error-correction model that includes the lagged error term, zt-1. Since ρ1 is significant at the 5 percent level, this implies that there exists an error-correlation mechanism that tends to restore the long-run equilibrium relationship between it* and itsm. Therefore, this test supports the hypothesis that it* and itsm are cointegrated.

Finally, as mentioned above, an important result of cointegration analysis is that if two variables are I(1) and cointegrated, there must be Granger causality in at least one direction. Our conjecture is that, Mexico being a small open economy, the implicit yield for Mexico’;s external debt causes (in Granger terms) the domestic interest rates on PAGAFES. Since the Granger is an F-test, it is applicable only to stationary variables; therefore, the test is applied to the first differences of it* and itsm. Table 6 reports the results from this test. As is shown, when Δit* is the dependent variable, the F-value is significant at the 5 percent level, supporting the hypothesis that itsm causes (in Granger terms) it*. Moreover, the test rejects the hypothesis that it* causes (in Granger terms) itsm, since the F-vaiue was found to be not significant at the 5 percent level when Δitsm was treated as the dependent variable.

Table 5.

Error-Correction Model

(Dependent variable: Δit*)

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Note: Only the lagged variables that were found to be significant are included; R2 is the coefficient of determination; and DW is the D urbi n-Watson statistic,

On balance, the evidence seems to support the hypothesis that it* and itsm are cointegrated. Nevertheless, the tests are not very robust and further investigation would be useful as more data become available.

IV. Conclusions

This paper explored whether domestic interest rates in Mexico can be linked to expectations of exchange rate changes and to perceptions of the default risk contained in Mexico’;s external debt.

Tests were conducted on the CIP and UIP hypotheses using two assets that contain the same country-risk premium but differ in the currency of denomination. It was shown that, with the exception of a short period when uncertainties in the economy were unusually large, deviations from CIP were small and random occurrences. In addition, the evidence suggests that the “peso problem” prevailed during the period under study; that is, expectations of the future spot exchange rate consistently overestimated the actual future rate, in a regime where the exchange rate was not allowed to float freely. This result is consistent with rational expectations and efficiency in the exchange rate market and does not allow one to reject the validity of the UIP hypothesis.

Table 6.

Tests for Granger Causality (Dependent variable:Δi*t)

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Significant at the 5 percent level.

A central issue in the paper was to investigate the extent to which both the domestic debt and external U.S. dollar-denominated debt issued by Mexico behave in the same fashion—that is, the extent to which they are affected by the same considerations about prospects of the Mexican economy as reflected in the risk of default. In this context, it was hypothesized that the interest rate on a U.S. dollar-denominated domestic asset is linked to the yield implicit in the secondary market price for external debt issued by Mexico. This implies that, from the point of view of creditworthiness, there is no distinction between the domestic and external components of Mexican debt. The empirical tests were not totally conclusive; however, on balance, the evidence suggests that once covered for exchange rate changes, domestic interest rates in Mexico and the implicit yield derived from the secondary market for Mexican external debt are cointegrated; that is, they move together in the long run. Moreover, it was shown that while the latter variable causes the former (in Granger terms), the inverse relationship did not hold.

An important policy implication derived from these results is that in order to achieve a permanent decline in domestic interest rates, policies undertaken by the Mexican authorities need to be successful not only in reducing the wedge between public expectations of exchange rate devaluations and the preannounced rate of depreciation, but also in improving the underlying conditions of the economy that affect international perceptions about the country’;s creditworthiness.

Important steps have already been taken in that direction. As the Mexican authorities continued in their adjustment efforts, credibility in the announced rate of depreciation of the peso was further enhanced during 1990–91. Moreover, the completion of a comprehensive debt-reduction package with foreign commercial banks in early 1990, coupled with a deepening of structural reforms and the progress achieved in the negotiations toward a North American free trade area, has improved international perceptions of Mexico’;s creditworthiness and helped increase Mexico’;s access to international capital markets.

Reflecting these developments, the implicit yield derived from the secondary market for Mexico’;s external debt declined from 25 percent in early 1990 to about 16 percent at mid-1991, and the interest rate on a U.S. dollar-denominated asset declined from 15 percent to 10 percent over the same period.

REFERENCES

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*

Hoe E, Khor is a Senior Economist in the Asian Department, He holds a Ph.D. from Princeton University.

Liliana Rojas-Suarz, an Economist in the Financial Studies Division of the Research Department, holds a Ph.D. from the University of Western Ontario

The authors wish to thank Charles Adams. Sterie T. Beza. Guillermo Calvo. Mohamed El-Erian, Robert Flood. Eliot Kalter, Claudio Loser. J. Saul Lizondo. and Sweder van Wijnbergen for their comments and suggestions. and Agustin Carsten and other members of the Banco de Mexico for the data used in this study.

1

For some of the tests conducted in this paper, we were able to extend the sample period somewhat.

2

Since 1982, Mexico’;s access to international capital markets has been severely restricted, and virtually all new lendings to the country have taken the form of concerted facilities in the context of debt restructuring arrangements. The few voluntary bond issues that have taken place in recent years have been subject to high—albeit declining—coupon rates, providing further evidence of the lower credit rating of Mexico relative to industrial countries.

3

Pagares de la Tesorería de lit Federación.

4

The interest rate on PAGAFES is indexed to the exchange rate in the controlled market, so that there is still a risk of an exchange loss in the event of a divergence in the spread between the exchange rates in the controlled and free markets. Since the investor would ordinarily have to remit his or her gains through the free market, an unanticipated increase in the spread between the controlled and free rates would lead to an exchange loss. During the period under study, the spreads between the two rates were less than 2 percent, except for a short episode in November–December 1987 when the spread widened to about 25 percent. However, the risk of a widening spread between the exchange rates in the controlled and free markets should affect equally the interest rates on both CETES and PAGAFES.

5

Certificados de la Tesorería.

6

Twenty-eight-day PAGAFES were issued beginning in January 1988. Hence, data for PAGAFES in 1987 were proxied by using interest rates on PAGAFES of 91-and 182-day maturity. Data for interest rates and exchange rates are closing bid rates corresponding to the last Wednesday of every month.

7

Testsfor developed countries have usually validated the CIP but do not support UIP. See, for example, Cumby and Obstfeld (1980).

8

Some of the empirical tests have analyzed the extent to which deviations from covered arhitrage can be explained by transaction costs. Many of these tests have followed the methodology suggested by Frenkel and Levich (1975), by which four transaction costs are identified: the cost of transactions in domestic and foreign securities and in spot and forward exchange rates. This methodology is not applicable here, however, because transactions in both CETES and PAGAFES are done in Mexican pesos: hence, the transaction costs of moving from one currency to another are not present. However, some transaction costs involved in the sale and purchase of the two assets still remain, which can account for the small deviations from covered arbitrage.

9

At the end of May 1990, the authorities reduced the depreciation of the exchange rate to 0.80 peso per U.S. dollar a day, and in mid-November, the depreciation was further reduced to 0.40 peso per U.S. dollar a day.

10
This is the well-known peso problem. Formally
Et(St+1)=S¯(1+αQt),
where S¯ is the fixed exchange rate, Qt is the probability of a devaluation, anda is the amount of a devaluation in percent. Hence, the forecast error
t=Et(St+1)St+1=S¯αQt>0,
as long as the devaluation does not take place (that is. as long as St+1 = S¯). It has also been argued that if the probability of a devaluation depends on economic variables that tend to show autocorrelation (such as the level of international reserves or the credit expansion to the public sector), the forecast error in the exchange rate market will also show autocorrelation.
11

Only five of the forecast errors were found to be negative. Again, in line with predictions from the model of the peso problem, a large and negative forecast error was observed in November 1987 when the peso was actually depreciated by 18 percent.

12

In a previous study of UIP in Mexico, Lizondo (1983a) found that during the period from May 1977 to December 1980, the forecast errors also had a positive mean but showed a small positive autocorrelation at the first lag. A possible explanation for the discrepancy between his results and the ones presented in this study is that, over the period covered in Lizondo’;s study, the interest rate was regulated by the Mexican authorities. This is in contrast with the most recent period covered here, when, with the exception of brief subperiods, the interest rate was allowed to float.

13

As noted in the previous section, the interest rate on PAGAFES should also incorporate a premium reflecting the risk of a divergence between the free and the controlled exchange rates.

14

On the appropriateness of treating domestic and external debt in a similar way, see Dooley (1987) and Guidotti and Kumar (1991).

15

Some have argued that the effective maturity of Mexico’;s external debt is infinite since it is subject to repeated rescheduling. However, this argument is no longer valid, since most loans have been converted into 30-year bonds whose principals are collateralized.

16

The series used correspond to the average price on restructured obligations for which data are available since 1986. Data on trade credits are available only since July 1988.

17
Assuming that the probability distribution governing debt-service payments is binomial, with π = probability of full payment, and 1 — π = probability of default, equation (9) can be rewritten as
Pt=πk=1nCt(1+itf)k+πFV(1+itf)n.
Therefore, it is straightforward that the probability of full payment equals
π=Ptk=1nCt(1+itf)k+FV(1+itf)n.
Since the denominator on the right-hand side of the above equation equals the present value of full debt service, the probability of full payment is the ratio of the observed price in the secondary market to the present value of full debt service.
18
We have also conducted the cointegration tests using an estimated yield from holding Mexico’;s external debt for one month ith—that is, the interest rate over a holding period identical to the maturity period of the domestic assets. Such interest rate was defined as
ith=CtPt+EPt+1PtPt,
where Ct and Pt have been defined above, and (EPt+1Pt)/Pt represents the expected capital gains from holding the asset during one month. The series for EPt+1, the one-month-ahead forecast of the price in the secondary market for Mexico’;s external debt, was constructed using an autoregression process. The series ith was found to be I(1), and the cointegration test results from using ith were not very different from using itsm. However, since in the process of forecasting EPt+1, one year of observations is lost, in the following discussions only the tests using itsm will be reported.
19

It should be noted that under the null hypothesis, all the test statistics have nonstandard distributions, and the critical values are taken from tabulations compiled by Dickey, Fuller, and other investigators.

20

Granger (1983) has shown that if two variables are bothI(l) without trends in mean and are cointegrated, such a data-generating mechanism exists.

IMF Staff papers: Volume 38 No. 4
Author: International Monetary Fund. Research Dept.