Monetary policy implementation and uncovered interest parity: Empirical evidence from Oceania

Abstract

The close integration of Australian and New Zealand financial markets and the similarity of the monetary policy regimes provide the perfect backdrop for testing the empirical relevance of uncovered interest rate parity (UIP) in Oceania. We find that changes in the bilateral exchange rate have become more sensitive to the short-term interest differential over time. Most important, after the introduction of the Official Cash Rate regime in New Zealand, the responsiveness of the exchange rate has accelerated to such an extent that it is incompatible with UIP. Evidence on UIP over longer horizons is mixed with a 10-year horizon providing the strongest support for the theory since 1990.

Keywords:

• uncovered interest rate parity
• cash settlement balances system
• official cash rate system
• instrument rule

Acknowledgements

The views expressed in this paper are those of the authors alone and should not be interpreted as reflecting the official position of the Reserve Bank of New Zealand on policy matters. The authors take full responsibility for any errors. We thank our colleagues, participants of the 2009 NZESG workshop and seminar participants at the Swiss National Bank for helpful comments. Our sincere thanks go to the referees and the editor-in-chief for making detailed comments and suggestions. Michael Reddell provided valuable feedback on the timing of the introduction of the OCR operating procedure at RBNZ. The first author wishes to thank the Swiss National Bank for its hospitality and for providing excellent research support.

Notes

 1. For notational simplicity we drop the subscript k from both interest rates from here on. In addition, we ignore the subtle differences between the unbiasedness hypothesis and the UIP proposition.

 2. Hodrick and Srivastava (1984, 1986); Sarno and Taylor (2002).

 3. A descriptive analysis of the reforms and the circumstances that led to the adoption of a formal inflation target can be found in Reddell (1999), Sherwin (1999), and Guender and Oh (2006).

 4. Archer, Brookes, and Reddell (1999) describe the reason for the Reserve Bank's decision to switch from the CSB target to the OCR framework.

 5. For lack of a better term, we refer to this period as the cash settlement balances target period even though the Reserve Bank changed the target rather infrequently and often relied on ‘open-mouth operations' (Guthrie & Wright, 2000) instead. For some time during this period the Reserve Bank also attempted to communicate to the public the stance of monetary policy through a monetary conditions index.

 6. The dramatic surge in carry trades that began in 2001 may have contributed somewhat to this increase in volatility. Being high-interest rate countries, both countries attracted sizeable inflows of short-term foreign investment. As a result, foreign exchange turnover in Australian and New Zealand dollars increased by 98% and 152%, respectively, over the 2001–2004 period (Galati & Melvin, 2004). The Kiwi Dollar and the Australian Dollar remained carry trade target currencies over the 2004–2007 period according to the 2007 Bank for International Settlements (BIS) Triennial Survey of Foreign Exchange and Derivatives Market Activity.

 7. Figure A1 in the appendix superimposes the change in the quarterly exchange rate on the quarterly yield differential. It is apparent that the OCR period witnessed far smaller fluctuations in the yield differential than the other subperiods while the volatility of exchange rate changes increased ever so slightly during the OCR period relative to the pre-OCR period. The summary measures of quarterly holding period returns and changes in the exchange rate are roughly the same as those reported for raw data in Table 1.

 8. We wish to state expressly that we present the empirical results based on daily and weekly data only for completeness sake. As such, we do not wish to over-emphasize these results. Statistical inference based on daily and weekly data is hampered by the presence of ARCH and GARCH effects. The existence of overlapping observations also introduces a moving average error into the estimated regression equation. For daily and weekly observations this poses a serious problem as the standard errors are not consistent. Thus, for daily and weekly data, hypothesis tests of the statistical significance of regression coefficients are–strictly speaking–not valid. The problem is less severe for monthly observations (k = 3 for monthly data on 90-day bank bill rates; the resulting MA error is of order k–1 = 2). All results reported were computed using OLS. Standard errors of the regression coefficient are based on the Newey-West procedure in EVIEWS with 2 or 3 truncation lags. The application of GMM where lags of the country-specific interest rates served as instruments yielded very similar results.

 9. For the OCR period, averaging the daily data over the quarter reduces the size of the estimated coefficient on the interest rate differential. It also increases the standard errors, thereby producing lower t-values in Wald tests. The explanatory power of the regression based on averaged data is also lower compared with that of the regression based on last-observation-of-the-quarter data.

10. Lothian and Wu (2005) propose an alternative way of testing the UIP hypothesis. It consists of pairing the yield on the foreign asset with the change in the exchange rate and then comparing the foreign return to the return on the domestic asset. Essentially, this approach requires, first, forming the sum of and s_{t  + 1} − s_t and, second, regressing this variable on i_t . The results remain largely the same with one exception. The null hypothesis that the coefficient equals unity cannot be rejected even though the estimated coefficient proves to be rather large at 3.133. Estimating the restricted version of the UIP test equation produces a much poorer fit as the adjusted R ² drops markedly (0.032). It is debatable whether this approach improves upon the standard way of testing the UIP hypothesis. After all, it imposes the restriction that the coefficient on the foreign interest rate equals unity.

11. Using survey data on expected exchange rates, King (1998) also finds support for the UIP hypothesis in Oceania over the 1987:Q2–1995:Q4 period.

12. A predictive test for coefficient stability as suggested by Chow (1960) yields mixed results. The log likelihood ratio test rejects the null hypothesis of a stable coefficient on the interest rate differential (break point 1999:2) while the F-test does not. Adding to the regression equation a dummy variable, which allows the coefficient to be different in size during the OCR period, leads to a positive but statistically insignificant coefficient on the interaction term. Other tests for structural stability (CUSUM, Chow breakpoint) produce no discernible evidence for instability.

13. There is no serial correlation present in the residuals of the regression equations estimated for the OCR period.

14. RBNZ formally announced its intention to switch to the OCR system on 8 February 1999. Financial markets had not been given advance notice of the impending change in the operating procedure. However, it was no secret that RBNZ had contemplated making changes to the operating procedure as early as December 1996. These deliberations were put on hold when RBNZ introduced the monetary conditions index (MCI) as a communications device with the help of which RBNZ attempted to signal its intentions about the future course of monetary policy to financial markets and the public at large. Dissatisfaction with the performance of the MCI led to its eventual abandonment. For further details about the change in the operating procedure at RBNZ, see Archer et al (1999).

15. The published long-term interest rates are only imperfect measures of the true yields on long-term government bonds. Ideally, one would use zero-coupon constant maturity interest rate series. Alexius (2001) proposes a method to remove the effects of coupon payments on bond prices.

16. Mussa (1979), Alexius (2001), Razzak (2002), Chinn and Meredith (2004), and Lothian and Wu (2005) are proponents of this view. As pointed out in the introduction, risk is likely to be a minor factor in the current context.

17. Notice the surge of the interest rate differential during the Asian Currency Crisis (shaded area) when RBNZ initially tightened while the Reserve Bank of Australia eased the stance of monetary policy. At the time, a monetary conditions index figured prominently in the monetary policy deliberations within RBNZ.

18. The current change in the exchange rate could also be regressed on the lagged interest rate differential. Estimating this specification of the regression equation over the whole sample period yields results identical to those reported in Tables 6 and 7. Slight differences in the coefficient estimates emerge if the two specifications of the regression equation are estimated over sub-sample periods. However, these differences are not pronounced enough to warrant separate reporting. These results are available from the authors upon request.

19. Estimating the same regression over the 1986:Q3–1999:Q1 period, i.e. the pre-OCR period, yields an adjusted R ² of 0.241 and an estimated coefficient of 1.297, which is statistically significant at the 1% level.

20. Alternatively, we could have specified the regression equation as: . The sample period extends from 1996:3 to 2008:4 and includes the OCR period. The coefficient estimates, standard errors, test statistics and goodness of fit statistics are exactly the same for the whole sample period as reported in Table 7. For the OCR period (1999:2–2008:4) the regression results based on the alternative specification differ minutely from those reported in Table 7. For instance, the slope coefficient is now estimated to be 0.713 with standard error of 0.199.

21. In Table 7, the constant in the regression is statistically significant over the whole sample period and the ‘Independence’ period. This is indicative of the existence of a risk premium. Since the early 1990s, the sovereign credit risk of both Australia (July, 1992) and New Zealand (April, 1993) has been rated AAA by Standard and Poor's for long-term bonds issued in local currency. Fitch Ratings are identical, although New Zealand's sovereign rating history with this agency extends back only to 2002. There are slight differences in the ratings of long-term debt issued in foreign currency, with Australia enjoying AAA status since 2003. New Zealand has been rated AA + since 1996. From 1996 to May 1999, New Zealand's credit rating for debt issued in foreign currency was slightly better than Australia's.

22. We have used a simple Phillips Curve, i.e. one without forward-looking inflationary expectations. Likewise, the expected output gap next period has been dropped from the IS relation. These forward-looking expectations are not crucial to the results. Forward-looking expectations of the rate of domestic inflation do appear in equation (11) however, because in the IS relation the output gap depends inversely on the expected real rate of interest. The additive disturbance in equation (11) is a composite term consisting of random disturbances and exogenous variables that appear in the IS relation, the UIP relation, and the Phillips curve. For further details on the derivation of equation (11), see Guender (2010).