**The Pricing Behavior of Korean Manufactured Goods During Trade
Liberalization **

**1. Introduction**

The idea that competition from imports would force oligopolistic firms in protected domestic markets to behave more competitively in setting their price is both old and widely accepted among economists and policy makers alike. There are, however, only a few empirical studies directly examining the response of domestic prices to changes in corresponding import prices. Some of these studies, conducted in the context of developed countries by Ceglowsky(1992), Levinsohn(1993), and Swagel(1996), are usually known as "competing goods effect" literature and they in general found that while the competing goods effect is significant, it is typically small. In the context of developing countries, however, we found only one empirical study on the topic, the 1989 paper by Corbo and McNelis. They conducted a comparative analysis of the response of domestic prices of manufactured goods to changes in external prices in Chile, Israel, and Korea during the period from the mid-1970s to mid-1980s. They found that domestic prices of manufactures in these countries responded substantially to changes in external prices, and that the coefficients of external prices and unit labor costs change systematically with the degree of openness of the economy.

However, while they intended a comparative analysis of pricing behavior for the three
countries, the lack of uniform data prevents it from being a serious and genuine
comparative analysis.^{1} At the same time, their findings of high external price
coefficients for these three countries appear to be overstated.^{2} Taken
literally, their high external price coefficients indicate a strong linkage between
domestic and external prices which then provides the basis for a substantial competitive
effect of import prices on domestic firms in setting the domestic prices of manufactured
goods. However, the existence of such large competitive pressures of foreign imports on
domestic prices of these industrializing countries, indicated by their results, is
doubtful. In fact, domestic prices of manufactured goods in one of these countries, namely
Korea, did not appear to respond to changes in external prices during a similar period, as
reported by Yang and Hwang(1994). We maintain that their external price coefficients are
highly overstated due largely to misspecification of their model (which will be discussed
later).

The size and speed of the response of domestic prices to changes in external prices are important for the purpose of stabilization policy. For instance, if the response of domestic price to changes in external prices is quick and substantial, then exchange rate stabilization is a more effective instrument for reducing inflation in manufacturing prices. But if domestic (manufacturing) prices do not follow corresponding external prices, then a macroeconomic policy of reducing excess demand and wage policy is more effective. At the same time, the coefficients of external prices also have important implications for competition policy. If the linkage between the two prices is strong, then open international trade through trade liberalization should be considered as a viable alternative, or at least a complementary policy, to promote domestic competition.

Given the presumed importance of the size and speed of the impact of external prices on
domestic prices in the formulation of these policies, it is necessary for economists and
policy makers to accurately measure of the size and speed of the impact of import prices
on domestic prices, particularly on prices of manufactured goods. The purpose of this
paper is to estimate the size and trajectory of the response of domestic prices of Korean
manufactured goods to changes in corresponding import prices and other determinants by
extending the work of Corbo and McNelis.^{3} We use monthly aggregate time series
data for the period from 1975.1 to 1996.6, which includes recent years from the mid-1980s
to mid-1990s when import liberalization was pursued aggressively in Korea. In addition, we
use both ordinary least squares and a Kalman-filter time-varying parameter regression,
which allows the capture of the time-varying price response of domestic manufactured goods
to changes in corresponding import prices during the period of import liberalization.

**2. A Brief Review of Import Restrictions and Liberalization in South Korea**^{4}

Import protection and liberalization in Korea proceeded with the stages
of economic growth. Import liberalization in the initial stage of economic growth from
1961 to 1967 was introduced mainly as a way of promoting exports. Since it was impossible
to pursue this export-promotion strategy while maintaining extreme forms of import
restriction, the Korean government freed export producers from the negative effects of
tariff protection and quantitative restraints by allowing tariff-exempt imports of raw
materials and investment goods. In the next period of economic growth during 1967-78,
however, little progress was made in import liberalization. During this period, the
average basic tariff rate was raised from about 17 percent in 1963-67 to about 26 percent
in 1968-72. Subsequently, the tariff rate was lowered to 20 percent in 1973-74 and 1976
and to 19 percent in 1978. According to Hong(1991), the nominal rates of protection for
manufacturing during the early stages of economic growth were 39 percent, 24 percent, and
25 percent, respectively for 1963, 1970, and 1978.^{5} Similarly, the effective
rates of protection for domestic sales for total manufacturing were 26 percent, 18
percent, and 13 percent for the same years.^{6}

Import restrictions in the second stage of economic growth, roughly from the mid-1970s to early 1980s, were mostly on consumption goods, as final-consumption goods dominated both Korea’s industrial production and its exports. According to Hong (Table 7.2, p.261), the rates of nominal protection during this period for durable consumption goods were 54.7 percent and 26.1 percent, and for manufacturing they were 19.1 percent and 19.4 percent for 1978 and 1982, respectively.

In the first half of 1978, import liberalization was again launched, albeit reluctantly; but as the balance of payments on current account deteriorated seriously in 1979, the import liberalization policy was halted in 1980. Import liberalization was resumed again in 1981 despite the enormous current account deficit, but was halted again in 1982. When the current account balance of payment improved substantially in 1983, import liberalization was again launched, and only since then has import liberalization been pursued as a long-term policy goal and has proceeded more or less on schedule to the present.

Actual data on the degree of import liberalization in South Korea are
available from four sources: An earlier study by Kim(1990) documents import protection and
liberalization for the early period from 1955 to 1985; Hong presents a comprehensive
discussion of import restrictions and liberalization experience for the period from 1961
through mid 1980s. More recently, the study by Yoo et al. (1993) provides a detailed
survey based on a cross-section study of import protection in 1990, and, finally, there is
the most recent study by Kim(1996) for the period from 1983 to 1996. While the general
pattern of import protection and liberalization in Korea from these studies is similar,
they are not directly comparable.^{7} Therefore, the necessary consistent
time-series data for import liberalization, required for serious quantitative analysis,
unfortunately are not available.

**3. The Model**

The specification of the price equation in Corbo and McNelis starts by
introducing two polar models of price determination, one for a closed economy and the
other for an open economy. When the economy is *closed* to trade in manufactured
goods, then manufactured goods prices are specified by a markup equation:

*P ^{c}_{t}* = a

where *P ^{c}* represents the rate of change in domestic
manufactured goods at time t in a closed economy; and

When all manufactured goods are homogeneous tradables and the economy
is completely *open* with no import restrictions, then domestic prices of
manufactured goods follow the law of one price, as specified in Aizenman (1986)

*P ^{o}_{t}*=

where *P ^{o}_{t}* represents the rate of change in
domestic manufactured goods in sector i at time t in an open economy;

*external prices* of manufactured goods in the local currency at time
t.

When the country is *semi-open*, it is reasonable to consider
pricing behavior as a linear combination of the two pricing rules as below:

*P ^{so}_{it}* = d

where d measures the extent that an open-economy pricing rule is in force and (1-d) measures the extent that the closed-economy pricing rule is in force. Putting (1) and (2) into (3) yields the price equation for a semi-open economy:

*P ^{so}_{it}* =d

Corbo and McNelis at this point assume that the rates of change in *UMC*_{t}
and *PE _{t} *are equal in deriving their final estimating equation:

*P ^{so}_{it} *= (1-d)a

While their assumption is made on statistical grounds, it is difficult to justify why
the rates of change in external prices of final goods will be the same as that of unit
material costs of imported raw materials. Once this assumption is made, however, the two
variables (*PE* and *UMC*) are now collapsed into one variable. Consequently,
the separate roles of external price and unit material cost of imported materials in the
determination of prices for manufactured goods cannot be ascertained. Not being able
ascertain separately the role of imported raw and intermediate materials in the
determination of domestic price is an important omission, particularly for a country like
Korea, which relies heavily on imported raw and intermediate materials.

In the present paper, we drop this restrictive assumption. By expanding Equation (4), we obtain the price equation for a semi-open economy:

*P ^{so}_{t}* = a

**4. Empirical Analysis**

*Data: *

The data required for the estimation of domestic price equations are
the two price indexes, two costs of production, and a measure for excess demand for total
manufacturing. The data for domestic prices for manufactured goods are ‘producer
price indexes for total manufacturing’ and the data for external prices (*PE*)
are ‘import price indexes on won basis’. Monthly price data are obtained
primarily from the Bank of Korea’s *Price Statistics Summary 1993,* which
provides a detailed and consistent data series for the period from 1970.1 to 1993.12. In
updating this primary source of price data by *Monthly Prices* (also published by the
Bank of Korea) for the 1994-1996 period, extreme care was exercised to make sure that the
categories in the two data series are essentially the same.

Since the data for unit labor costs in each sector (*ULC*) are not
available, they are constructed following a method originally used by Hooper and
Mann(1989), utilizing data on earnings, man days, employment, and an industrial production
index in each sector.^{8} The data on unit material cost (*UMC*) is the
‘raw materials and intermediate goods’ component of import price indexes on won
basis, also available from *Price Statistics Summary 1993* and updated using various
issues of *Monthly Prices*. Finally, excess demand (*ED*) is measured as the
rate of change of new orders less the rate of change of inventories in each manufacturing
sector, the same measure as used in Corbo and McNelis. Data on producer’s shipment
indexes and producer’s inventory indexes in each sector are obtained from *Economics
Statistics Years Book* and *Monthly Statistical Bulletin*, both published by the
Bank of Korea.

*OLS Estimation Results:*

Table 1 reports the results of applying ordinary least squares to Equations (5) and (6) for total manufacturing together with all relevant regression statistics. We use the maximum likelihood method of Beach and MacKinnon(1978) to deal with possible autocorrelation. One of our tasks is to show, using the Korean case, how misspecifcation in the Corbo and McNelis’study overstates the magnitude of the impact of changes in the prices of imported manufactured goods on corresponding domestic prices.

The first row in Table 1 presents regression results based on the Corbo and McNelis
specification of combining the influences of changes in external price and unit material
costs. We note that the estimated coefficient for *PE*, 0.376, is fairly high and
significant. Since the two price variables are measured as rates of change, this estimated
coefficient is the external price elasticity and it indicates a high degree of openness of
the economy. Estimated coefficients for the unit labor cost and excess demand are found to
be very small and insignificant.

The second row in Table 1 presents regression results based on Equation (6) by allowing
for the separate roles of external price and unit material cost in the determination of
domestic price. We note that the estimated coefficients for external price and unit
material cost are 0.221 and 0.200, respectively, and are both substantial and highly
significant. When the second variable *UMC* is combined into one variable, as in the
first row based on the Corbo and McNelis specification, then the estimated coefficient for
the first variable *PE* (0.376) measures the gross effect of *PE* on *PD*.
This is approximately equal to the sum of the direct effect of *PE* on *PD*
(0.221) and the indirect effect of *UMC* on *PD* (0.200 *0.860), where 0.860 is
the slope coefficient in the regression of *UMC* on* PE*. This is a classic case
of specification bias due to the omission of a variable.^{9} Reviewing the
regression results in the first two rows in Table 1, it is clear that the estimated
coefficient for external price of 0.376 captures the sum of direct and indirect effects on
domestic price. The extent of the upward bias in the estimated external price coefficient
in the Corbo and McNelis study is equal to the size of indirect effect of *UMC* on *PE.*

While we are successful in separating the role of changes in external price and imported material cost in the determination of domestic price, the estimated external price coefficient in the first two rows may still not give a true measure of the impact of import prices on domestic price during trade liberalization. Indeed, we argue that it is substantially affected by the high rate of inflation during the period of 1980-81. There are two reasons why we believe this is the case. First, we note that the estimated external price coefficients in the Corbo and McNelis study rose sharply in the two periods of rapid inflation during 1974-75 and late 1979-1981 for both Israel and Korea. However, no serious import liberalization took place either in Israel or Korea during these periods. Second, examination of the actual data for the rates of changes in the dependent variable and external prices for Korea reveal sharp increases for the two variables during the 1980-81 period.

The effect of high inflation during the 1980-81 period on the estimated coefficient for the external price is first tested in two ways. First, we re-estimate the final price equation with data for the two sub-periods: 1975.1-1982.12 and 1983.1-1996.6. The estimated external price coefficient obtained from the first sub-period which includes the high inflation period of 1980-81 turns out to be 0.393 (in the third row), substantially higher than 0.221 in the second row. In contrast, the estimated external price coefficient from the second sub-period is very small and insignificant, as shown in the fourth row of Table 1. Second, the fifth row in Table 1 presents regression results with the final price equation after introducing the slope dummy for external price. The slope dummy variable takes the values of "1" for the 1980-1981 and "0" for the remaining years. Once we introduce the slope dummy variable for external price, the magnitude of the estimated coefficient for the external price becomes very small and insignificant, results similar to those displayed in the fourth row. In contrast, the coefficient for the interaction term for the external price, DV*PE, is 0.697 and highly significant. Examining this regression, it is interesting to note that the estimated coefficient for the imported unit material cost, UMC, is 0.108 and highly significant. In addition, the effect of inflation is also tested by omitting the years 1980 and 1981 from the regression and the results are presented in the last row of Table 1. As expected, the magnitude and the sign of estimated coefficients are very similar to those with the slope dummy variable in the above regression results in the fifth row.

Therefore, once we account for the effect of high inflation during the 1980-81 period,
changes in external prices do not seem to affect the pricing decisions of manufactured
goods in Korea. Does this mean that domestic prices do not respond to changes in external
price at all? Not necessarily. It simply says that the rate of change in domestic prices
does not respond *on the average* to the rate of change in import prices. As is well
known, the OLS estimates are based on a fixed-coefficient model and thus provide only
averages for the variable coefficients for the whole sample period. In fact, this evidence
should caution us about the limitations of OLS estimates, particularly when the period of
estimation includes a unusual subperiod (like 1980-81), which may be structurally
different from the rest of the period. It is therefore desirable to examine how the
estimated coefficients vary over time as the economy gradually opens up over the course of
the period. We now turn to a time-varying parameter regression model.

Time-Varying Parameter Regression:

The price equation in (6) can be written now with time varying as follows:

P^{so} = a_{0}(1-d_{t}) + d_{t}PE_{t}
+ a_{1}(1-d_{t})ULC_{t}
+ a_{2}(1-d_{t})UMC_{t}
+ a_{3}(1-d_{t})ED_{t}
+e_{t}

= **X**_{t}^{’}**b _{t}**
+ e

The coefficient vector **b _{t} **is assumed to
follow AR(1),

**b _{t} = c** +

where **c**, **b _{o}**,

(8)

where . Thus the implied restrictions imposed on parameters in (7), are

, and **h _{t}
=** n

If we relax the assumption of 's being constant, we would then have a different
stochastic process for bt. In estimating the time varying equation (6'), we estimated both
the model characterized by (9) and the model without restrictions on c, b0 and Q.10

Figure 1 presents the time path of external price coefficient, dt , in (6') and (7)
without restrictions. Examining the trajectory of dt in Figure 1, we note two main
features: First, ignoring the sharp rise in the path during the first few years, it starts
from a low level of 0.1 and rises sharply during the 1980-81 period and falls afterward
even more sharply through 1983 and stays more or less at the same level through 1984. How
should we interpret the sharp rise in the path of dt during the 1980-81 period when we
know that trade liberalization did not take place in Korea during this period? It is
reasonable to argue that the sharp rise in the path of dt during the 1980-81 period may be
the result of the high inflation during that period. There are two pieces of evidence for
our argument. First, the estimated coefficient for the external price became very small
and insignificant, as presented in Table 1, when we account for the effect of inflation
both by introducing a slope dummy variable for the 1980-81 period and by omitting the
years 1980-81 in the OLS estimation. Second, when we re-estimate the price equation with
the Kalman filter with the interaction term for the external price, DV*PE, the trajectory
of dt shows no such sharp increase during the 1980-81 period as in Figure 1.11 Based on
these results, we can forcefully argue that the large increase in the time path of dt
during the 1980-81 period is largely due to inflation.

Second, starting in about 1985, the time path of external price coefficient, dt, shows a
small but clearly increasing trend. This particular pattern is consistent with our
understanding of trade liberalization undertaken in Korea. As discussed earlier, while the
import liberalization movement was on and off in 1980-82, a renewed movement toward import
liberalization was followed since then as the current account situation substantially
improved in 1983. Then, as Hong(1991) points out, "this time, however, it (meaning
trade liberalization) was pursued as a long-term policy goal with a definite annual
liberalization timetable." Focusing on the time path of dt for the period between
1985 and 1996, we observe that firm's price-setting behavior appears to be increasingly
influenced by import liberalization pursued during the period. However, it is important to
note that the magnitude of the domestic price response to external price changes in Figure
1 is much smaller

5. Conclusions

In this paper we re-examine the impact of changes in external prices on pricing decisions of domestic manufactured goods for Korea, utilizing the monthly data from 1975.01 to 1996.12. The main results of this paper show that domestic prices respond to changes in external prices, but the magnitude of the elasticity of price with respect to external price is 0.221, which is much smaller than what Corbo and McNelis(1989) reported for Korea. Our results clearly show that their unusually high price elasticity (0.48 for Korea) reflects a combined effect of changes in external prices and in imported material costs due to an unusual assumption in their model specification. Further, our results show that their unusually high price elasticity also reflects the effect of inflation during the high inflationary periods of 1974-75 and 1980-81. Finally, our results also show that Korean firm's price-setting behavior appears to be increasingly more influenced by import liberalization since mid-1980s. The timing coincides with the pursuit of import liberalization starting in 1983 as a long-term policy goal in Korea.