The transactions demand for money in the presence of currency substitution: evidence from Vietnam

Abstract

Currency substitution – the use of foreign money to finance transactions between domestic residents – is widespread in low income and transition economies. Traditionally, however, empirical models of the demand for money tend to concentrate on the portfolio motive for holding foreign currency, while maintaining the assumption that the income elasticity of demand for domestic money is invariant to the degree of currency substitution. A simple re-specification of the demand for money is offered which more accurately reflects the process of currency substitution by allowing for a variable income elasticity of demand for domestic money. This specification is estimated for Vietnam in the 1990s. Using a standard cointegration framework evidence is found for currency substitution only in the long-run but well-defined wealth effects operating in the short-run.

Notes

Calvo and Végh (1992) use the term currency substitution to describe the use of a foreign currency as a means of exchange and the term dollarization denotes the use of a foreign currency in any of its three functions: unit of account; means of exchange; or store of value. This distinction is maintained here.

In Matsuyama et al. (1993), for example, these costs arise as a result of the random matching of agents. Alternative mechanisms might include menu costs or other costs associated with maintaining parallel payment technologies.

For example Bahmani-Oskooee and Malixi (1991), Adam (1992, 1999), Arize (1992, 1994), Perera (1993), Chowdhury (1997), Dekle and Pradhan (1997), Choudhry (1998), Tan (1997), Weliwita (1998), Buch (2001), and the survey paper by Sriram (2001).

This is consistent with the anecdotal evidence suggesting that private sellers of goods or services rarely refuse to conduct trade in US dollars instead of Vietnamese dong, and that curb markets in foreign currencies are widespread in all the cities. In both cases, these activities are officially forbidden but very widely tolerated.

In 1990, the structure of GDP was agriculture 38.7%, industry 22.7% and services 38.6% and in 1999 agriculture 25.4%, industry 34.5% and services 40.1%. Source GSO (2000), Table 17 p. 28. The sample correlation between industrial output and total constant price GDP, based on quarterly data from CIEM (2000), is 0.983.

Other specifications of the expected depreciation do not reject the model, but preliminary results have indicated that this type of specification leads to the best goodness-of-fit of the model.

The literature on ‘non-linear cointegration’ is relatively new and incomplete, see Park and Phillips (2001). Empirical work to date tends to focus on non-linearities in the adjustment process. See for example Enders and Granger (1998).

The initial parameter value δ = 2 was chosen using a simple grid-search method using the maximum of the first eigenvalue within the Johansen cointegration analysis as the criterion. Having chosen this initial value the logistic function was scaled by a factor of 2. This does not alter the model but allows a direct interpretation of the coefficients of the model.

Given the relatively under-developed financial system in Vietnam, it is found that the demand for money can be fully described in terms of the income and the rate of exchange rate depreciation; the vector Z has therefore been dropped from the empirical specification.

Preliminary results suggest that the deterministic components included in the dynamics should be a constant restricted to the long-run relationships, unrestricted (centred) seasonal dummies but no deterministic trend.

Full details of all the results are available from the corresponding author.

The non-linear least squares estimation is performed using PcGive 10.0 ; see Hendry and Doornik (2001) vol. 1, chapter 17, pp. 247–8.

The evidence from Hansen's instability tests is supported by recursive estimation results, which are available on request from the corresponding author.