Socialist and world market prices: An ingrowth?

JOURNAL OF COMPARATIVE

ECONOMICS II,2

1-39 (1987)

Socialist and World Market Prices: An Ingrowth?’ JOZEF M. VAN BRABANT

United Nations, New York, New York 10017 Received January 28, 1986; revised July 16, 1986 van Brabant, Jomf M.-Socialist

and World Market Prices: An Ingrowth?

This paper presents tests of the relationship between East-West trade prices and intra-Council for Mutual Economic Assistance (CMEA) prices for transferable ruble transactions for two sets of Hungarian data. A special test on the significance of the introduction of the Moscow modification of 1975 is executed where feasible. This is relevant for more than half of the commodities studied. Well-behaved relationships between world and CMEA prices are obtained for the majority of commodities, but frequently the elasticities diverge significantly from unity. This suggeststhat changes in averageEast-West trade prices are transmitted into the CMEA price formula only partly. For those.products for which the relationship is statistically weak, other determinants of CMEA prices should be explored. 0 1987 Academic mess IIK. Journal of Economic Literature Classification Numbers: 052,42 1,423.

There are few questionsabout the Council for Mutual Economic Assistance (CMEA) that have been debated for so long, and about which the answers have remained so elusive, as the determination of intragroup trade prices.* The theory is quite clear: contract prices for regular transactions denominated in transferable rubles (TRs) are negotiated bilaterally on the basis of reference prices derived from 5-year averagesof main world market prices (WMPS).~ For some goods,this referenceprice is augmentedwith “saved height charges.” ’The views expressedhere are my own and do not in any way reflect an expression of opinion on the part of the Department of International Economic and Social Affairs of the United Nations Secretariat in New York with which I am affiliated. I am most grateful to an anonymous referee and the editor of this journal for detailed comments and constructive suggestionson an earlier version of this paper. ’ I shall deal here only with the active European members of the CMEA: Bulgaria, Czechoslovakia, German Democratic Republic, Hungary, Poland, Rumania, and the Soviet Union. It is not clear to what extent these principles apply also to other socialist countries. 3 For a broad conceptual framework of the evolution of CMEA price-determination principles, seevan Brabant (1983; 1987, Chap. 4). The latter source discussesthe definition and comparative importance of base, reference, contract, and transaction prices. 21

0147-5961187 $3.00 Copyright Q 1987 by Academic Press,Inc. All rights of reproduction in any form reserved.

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JOZEF hi. VAN BRABANT

These are calculated by the same averaging method as followed for the reference ~rice.~ Note that both the average annual reference price and freight chargeare converted at the TR’s averageannual exchangerates for convertible currencies5 Becauseit is difficult and time consuming to identify WMPs, the question of how actual CMEA trade prices are in fact set remains fundamentally unanswered6 Given the complexity of arriving at proper prices and the importance that the centrally planned economies (CPEs) attach to their trade with market economies (MEs), this paper reports on tests of the hypothesis that in negotiating intra-CMEA prices they refer back to the prices obtained in their EastWest transactions, perhaps after suitably modifying them to reflect quality and other differences between the two goods traded. It is known that “prior experience” has been one of the valid documentations for the construction of referenceprices.’Given the growing diversification of the structure of output and trade, and the sharply increased importance of East-West trade in recent years as compared with when the pricing prescriptions were first endorsed, the CPEs have probably increasingly relied on East-West trade prices (EWPs) as an alternative to pure WMPs. This paper presents empirical findings about the above hypothesis. Section 1 discussesthe “ingrowth” concept, its possible importance in CMEA trade, and how this m ight lead to prices that could be seriously distorted from the point of view of the long-held CPE objective to simulate “competitive prices.” The specification of the models tested and a discussion of their lim itations follow. Section 3 clarifies the nature of the two complementary sets of price 4 The freight principle statesthat the net transportation gains, i.e., alter deducting actual freight charges, derived from trading with CMEA partners instead of with other economies, given the geographical proximity of the members, should be divided equally. This is particularly important for some raw materials. For details, see Koz’menko (1986), Shanina (1976, pp. 169-173), and Teichmanowa (1976). 5 The logic of the Bucharest principles suggestsconversion of W M R at averageannual exchange rates. Spa&k (1984, p. 439; 1985, p. 1) report such a practice, which was apparently recommended by the Standing Commission on Currency and Financial Questions possibly in late 1978 as reported by Konstantinov (1982, p. 128). In earlier experiments, which are discussedin van Brabant (1984, 1985a,b), I explored four different dollar exchange rates of the Soviet ruble, average annual and average December rates of each reference and of each negotiation year, as well as the average annual rates of the TR. On the whole, the conversion at average December ruble rates tended to yield superior results. Average December rates of the TR might be better stiIl, though this cannot be verified for lack of data. 6 These identification problems are elaborated in van Brabant (1983; 1987, Chap. 4). In van Brabant (1984, 1985a,b), I examined the relationship between CMEA trade prices and WMPs as derived from quotations in commodity exchanges.But the hypothesis could be tested for only a few products, owing mainly to the lack of information about the proper reference WMPs. This empirical problem is a symptom of much deeper-seatedissuespertaining to the identification of proper reference WMPs and agreeing on them for transaction purposes. ’ For a historical backdrop, see van Brabant ( 1987, Chap. 4).

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23

data at my disposal. The estimates are analyzed in the following two sections. The final section states several caveats to properly interpret the results. 1. RATIONALE OF AN INGROWTH MODEL According to principles established for their regular intragroup exchanges denominated in TRs, the CPEs construct reference prices as averagesof past WMPs. Contract and actual transaction prices are agreed on in bilateral negotiations. Obtaining prices that reflect “socially necessarylabor expenditures” is a cherished objective of the CPEs becausethis allegedlyleads to “equivalent exchange.”There are many problems with this ambition that cannot be examined in detail here. The most important ones are that point prices are available for only a comparatively small set of basic commodities traded on international commodity exchanges.For other products, suitable comparators have to be constructed. There is also a class of goods for which the W M P concept is simply not an operational one. This is especially crucial for the regional exchange of manufactured goods, whose importance has expanded rapidly over the years. In some cases, domestic prices or price lists of M E firms that are important in world trade have been utilized as gauges.But a growing component of trade has been evaluated according to criteria that have remained unclear. This paper explores the possibility that the CPEs for some goods refer back to averaged EWPs. In and of itself, this does not contradict the Bucharest principles, and hence the question mark in the title. The interpretation of the results based on a comparison of TR prices and EWPs depends essentially on whether the latter, as reported by Hungary, are part and parcel of a proper sample of WMPs or whether they reflect the peculiar position of CPEs in East-West trade.8 In the first case, the price observations reported here may usefully amplify the range of documentable WMPs and could thus provide further evidence to buttress the results of previous empirical inquiries based on W M P S proper (as in van Brabant, 1985a). In that case the title needs to be rephrased. If prevailing WMPs are actually not the real alternatives at which these countries trade in world markets, the observations on EWPs are logically the real pointers for the construction of reference prices from which CMEA contract and transaction prices are derived. Unfortunately, it is impossible to ascertain unambiguously whether WMPs and EWPs coincide. In view of the peculiar position of CPEs in global trade, there is considerable logical evidence as well as some empirical support for * Given its size and unique position among the CPEs, can Hungary’s trade experience be rep resentative for other CMEA members? The exercises were based on the contention that EastWest trade in important goods is on the whole competitive and must therefore result in fairly homogeneous prices. Furthermore+ the official CMEA principles include the stipulation of TR price uniformity, which may prevail on average,as argued in van B&ant (1983; 1987, Chap. 4) for the type of goods subjected to the empirical tests discussed in van Brabant (1985a,b).

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JOZEF M. VAN BRABANT

the hypothesis of a major divergence between EWPs and WMPS.~Some of these divergences may be completely random. But others, and probably the most important ones, are systematic in nature, although they certainly need not invariably stem from discrimination.” Systematic deviations between EWPs and WMPs m ight also result precisely from the “peculiar costs”incurred by MEs when they trade with CPEs. WMPs would therefore still be relevant to the discussion of the origin of CMEA trade prices, but only after some additional transformations to reflect the special circumstances and conditions typical of East-West commercial relations. One m ight group these systematic divergences into the effects of political animosities, trade policy measures, and commercial peculiarities. Clearly, a flareup of East-West tensions exacerbates the hindrances to smooth trade with MEs. Embargoesin one form or another can be circumvented but only at a cost. Many CPEs do not enjoy most-favored-nation status. Also, their agricultural exports are frequently subjected to discrimination, especially by the European Communities with which the CPEs conduct the bulk of their East-West commerce. In addition, their manufactures form the object of several types of nontariff barriers and informal arrangements of “managed trade.” Furthermore, the CPEs generally do not benefit from preferential arrangements such as those granted to developing MEs and have come to face formidable competition from the NICs in MEs (see Poznanski, 1985, 1986, among others). Finally, it is known that CPE exportables suffer from quality and service problems that are reflected in price discounts in MEs. Likewise, it is often a very costly operation for an M E firm to gain an export contract to a CPE. This is possibly reflected in a price markup. If CMEA trade prices for a number of important goods are derived on a regular basis from EWPs that diverge consistently from prevailing WMPs for political, systemic, or commercial reasons,the principles that are allegedly so cherished by socialist countries cannot possibly be attained. It m ight imply that the CPEs are importing into their own preferential trading system the results of all kinds of discriminatory trade policies that they have rejected as reprehensible in international relations.” That is why I call the EWP specification an “in-growth.” 2. THE SPECIFICATION OF THE MODEL The hypothesistested is that there exists a close relationship betweenaverage EWPs during a given period of time converted at the averageannual exchange 9 For anecdotal and other evidence, see the useful summaries in Ausch (1972, p. 84), Marer (1972, pp. 37-56), and Marrese and Vafious (1983, pp. 3-34). lo K&es ( 1985, pp. 2 18ff.) argues that any divergence stems from discrimination, but that is a non sequitur. ” The CPEs have articulated this stance in particular at UNCTAD fora; for one comprehensive declaration, see UNCTAD (1976, pp. 145-172).

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25

PRICES

rate of the TR and the corresponding TR price. Proper averagingis understood to proceed as per the rules enunciated in Bucharest with the Moscow modifications. In theory the following exact relationship should hold between the TR price and the average past W M P for any product:

(1)

PC, = PR, + (& - T,)/2

where PC is the actual contract price, PR is the reference price, F is the averaged freight charges that would have been incurred had the product in question been imported from an ME, T is the actual freight charge incurred for flows through third countries, and t is time. PR in this equation should be equal to PR~=~f~~’ WMPi

fort> 1975

and

k=3

fort=1975

k=5

fort> 1975 (2a)

r-t-1

=i ,z

WMPi

for 1965?< t< 1975

(2b)

r-t-1

where f is the first year under the original Bucharest price regime in which the new price basewas implemented, t is the current year, and k is the number of years over which the averaging is to be computed.” Likewise, Ft is the average of M E freight rates applicable in the reference years. Under the ingrowth model, the right-hand side of Eq. (2) would have EWPs instead of WMPs as base. There are several questions to be settled before embarking on estimation and interpretation. The first is whether CPEs really try periodically to construct intragroup trade prices on the basis of prices prevailing in East-West trade for all goods as called for by the principles. Even if they did so, they m ight allow discounts and markups for valid reasons.Consequently, since the CPEs set contract and transaction prices bilaterally, in contrast to baseand reference prices that are supposedly uniform for regular TR-denominated transactions, the divergence from the proper reference price m ight be systematic because the products traded in the two different relations are not really identical. It m ight also be random simply becausebilateralism itself allows for all kinds of ad hoc adjustments. This includes the application of the freight charge, which has been utilized as a particularly flexible instrument for generating TR prices that deviate from referenceprices (Ausch and Bartha, 1969, p. 112). As a result, the relationship between CMEA prices and EWPs should be stoI2 The averaging over a fixed 5- or 6-year period before 1975 pertained roughly to the 10 years between 1965 or 1966 and 1974. Prior to the mid-1960s, a l- or 2-year reference base was more common.

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JOZEF M. VAN BRABANT

chastic rather than exact. Becausethe systematic factors are unlikely to change annually provided the products are properly identified (see Section 3), and becausenothing at all is known about the degreeto which the freight charges are imposed on individual goods or the preciselevel of thesecharges,a doublelogarithmic stochastic respecihcation of the relationship between PC and PR is appropriate: logPC,=a+blogPR,+u, (3) where ut is a multiplicative error term whose distribution is presumed to exhibit mean zero and constant variance.r3 This model posits that changes in EWPs are transmitted into the CMEA price formula, although the level of EWPs and CMEA prices may be different.14The chief reasonswhy the CPEs are more likely to aim at the transmission of price changesthan at absolute levels are examined by van Brabant (1985a, p. 240). At first glance, the hypothesis implies that price negotiators scrutinize every product price each year so as to ensurethat the changesin EWPs are accurately reflected in transaction prices. Taken strictly, this is implausible. For the period antedating the Moscow modification in 1975, the above rule requires only that EWPs be ascertained at intervals. But there is ample evidence that negotiators never did reexamine prices comprehensively.‘5 Matters changed dramatically when the sliding price formula was introduced in 1975. Not only should the prices in principle be recomputed every year, but the CPEs should have recomputed them also becauseprice fluctuations in the 1970swere bigger than before. Yet if negotiators could not reset prices comprehensively at great intervals, would they have done so more recently on an annual basis? Price negotiations tend to be protracted and so hlled with irreconcilable and tendentious documentation that it is impossible to redo every price on an annual basis. Nevertheless, the hypothesis can be interpreted in a slightly different format. The model tested here does not require that prices for all products traded in the CMEA be restudied every year in a comprehensive way. Nor does it depend uniquely on the unity test. All it does is test whether, in case of really important changesin EWPs, a repercussion can be detected in CMEA prices ” If the proper relationship were as depicted in Eq. ( 1) with some random factor added, estimation would have to proceed without duly factoring (F, - T,) into the equation. This would lead to an error in the independent variable, and hence might entail possibly serious econometric problems if F, were not random but systematically related to PR,. I am assuming here that F( is applied randomly for each commodity tested, which enablesme to use OLS since the covariance between the independent variable and the error term is zero. F, might even be a constant proportion of PR, in which caseOLS on the double-logarithmic specikation would still yield consistent elasticiw estimates. I4 For a much more exhaustive aggregativeverification of the impact of changes in WMPs on average intra-CMEA prices, see Hewett ( 1974). I5 Ausch (I 972, p. 94) for Hungary, and Rutkowski (1977, p. 4) for Poland.

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and the degreeto which the transmission of EWP changesinto CMEA prices takes place. Price negotiators must have a feel for whether a reexamination is in order, especially in the case of the most important primary goods that comprise the bulk of the products dealt with in this investigation. It should be remembered that all CPEs maintain at least one research institute that closely monitors international prices of a wide variety of important traded goods and services. Precisely becausereference prices for some products included in this study fluctuated widely and because the guidelines made it possibleto reset prices, one would expect negotiators to attempt to alter prices for important goods if changesin WMPs could be easily ascertained. 3. THE NATURE OF THE DATA AND STATISTICAL TESTS For a number of years, Hungary has been reporting price indices for ruble and dollar trade, and in some cases absolute prices in convertible currency transactions.‘6 It is important to emphasizethat these data are not unit values as conventionally defined but are averagesof real prices for well-identified products traded within the calendar year, possibly with more than one partner.” Regarding absolute values, which are reported in the compendium of foreign trade price statistics (in K&t), the information is actually confined to nonsocialist exports and imports. Owing to the highly restricted distribution of KAA, the longest series of data at my disposal covers the years 1958 to 1977.18Since ruble and dollar price indices are compiled in the standard foreign trade yearbook (KSE), a much longer seriescan be tested and the test can be administered to many more products than what is available from K&l. There are three conditions to be fulfilled to test this supplementary or alternative information: (i) a given product must be traded in both currency relations, (ii) the dollar price series must be sufficiently long to permit the calculation of average EWPs and the comparison with TR prices in CMEA trade, and (iii) the product in question must have been traded nearly continuously with the two currency zones. Regarding the first condition, the trade I6 Remember that dollardenominated commerce is not restricted to MEs. ” In all honesty I must admit that this is a logical conjecture rather than a solid confirmation based on firm meihodological explanations. The products for which price indices and absolute values are available should be identical up to a constant (the base year). Unfortunately this is not always the case, but I have not yet been able to come up with a convincing explanation for this state of affairs except perhaps that one series refers to the average price in nonsocialist trade and the other to the averageprice in convertible currency trade. On several occasions since late 1984, I have sought clarification of the nature of these data from the Hungarian statistical office and knowledgeable Hungarian researchers.So far my efforts have regrettably been in vain. ‘* More recent editions of KAA than the one available to me are occasionally referred to in the specialized Hungarian literature. These are apparently internal documents circulated only within the Ministry of Foreign Trade. Several inquiries addressed to the Hungarian central statistical office regarding this and other matters have thus far been completely ignored.

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JOZEF M. VAN BRABANT

statistics do not ensure that a product is identical in dollar and ruble relations. Rather, I assumethat any divergence has remained constant over the sample period. There are, unfortunately, not that many goods for which the above reinterpreted conditions hold, even if one is prepared to compromise on the length of the time series. Several products have no reported price or price index in the m iddle of the series. This is not as serious for TR as it is for dollar relations. The latter data must be available for a continuous period of 5 years, except in 1975, so that a proper averagecan be prepared. Of course, one needs to be able to do this for a sufficient number of years to permit regression analysis at all. In cases where only one or two observations in a series are m issing, I have endeavored to estimate a link on the basis of actual trade returns or price variations in other CMEA or East-West relations; in some cases, I have simply inserted a plausible “guesstimate” so as to make econometric analysis feasible at all. I9 Accordingly, two different setsof products can be tested. The first one takes advantageof the lim ited information about absoluteprice levels for Hungarian exports to and imports from nonsocialist countries for selected goods. The commodities for which continuous data are available up to 1977 and comparable TR data can be compiled comprise 13 export and 20 import products. The second set of data makes use of dollar price indices, which can be constructed in most cases up to 1984, and comparable TR price indices. The number of goods for which more or less continuous and comparable time series can be compiled extends to 36 export and 26 import items. In principle, there should be no distinction in price behavior for exports and imports for an identical product, and one could therefore pool all available commodities into a single sample. Unfortunately, this hypothesis cannot be tested directly, owing to the exceedingly narrow range of goods for which both export and import price series for dollar and ruble trade are available. In the case of the set of data based on dollar prices, only one product, diesel fuel, is common; for the data basedon the price indices, possibly three common products are availablebut two have weak statistical fits. Although the common items were very few indeed, it may neverthelessbe instructive to keep exports and imports separate basically because the former can be presumed to be more homogeneousthan imports in provenance, quality, and price. On the basis of the above data, regressions of the form of Eq. (3) were run on the data available in absolute values. A slightly modified form was explored for the price indices, for reasons explained in Section 5.20 I9 A note on the data, including the sources and methods used to obtain the missing links, is available from the author. M Au equations were run in linear as well as in log-linear form for data converted at all alternative exchangerates available to me. But only the results of the log-linear specification of data converted at annual TR of observation years are reported becausethey are marginally superior and slightly easier to interpret. Other results are available upon request.

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One final caveat before presenting the results: when should one reject a coefficient as being not significant or as diverging statistically from an a priori value? Since the test is for a positive relationship between the TR prices and EWPs, I perform a standard one-tailed t test on the elasticity coefficient. Furthermore, because it is presumed that the elasticity parameters should not differ significantly from unity, a two-tailed test is appropriate since there is no a priori reason for excluding random effects that bias the prices downward or upward as compared with EWPS.~’For the reason given, I have chosen a 95% confidence interval around the coefficient, but other values could have been explored. 4. ESTIMATES BASED ON ABSOLUTE PRICE DATA The most important parameters for the equations involving absolute prices are listed in Table 1. Of the 13 export products, four equations, those of televisions, fresh eggs, calcinated alumina, and light bulbs, have negligible, and one other has small, explanatory power, in addition to the former having the wrong sign. These features indicate that there must have been significant other influences, aside from average WMPs, that had a key role in the formation of TR prices. Similarly, on the import side, polyethylene, polystyrene, and newspaperhave negligible, and potassium and nitrogenous fertilizer have small, explanatory power, and the former has the wrong sign. Though an excellent equation is obtained for phosphate fertilizer, the elasticity parameter has the wrong sign. Needlessto say, the elasticity values of these commodities were usually very far from unity with coefficients that were statistically significant only in rare cases. Though most of the other equations listed in Table 1 have very high t and R2 statistics, in the majority of casesthe elasticity parameter diverges significantly from unity, the value that should have been attained if strict adherence to the Bucharest principles existed. In fact, there are only four export and import products each for which the elasticity estimated did not differ from unity, but only two import equations, those for timber and raw cotton, have significant explanatory power in contrast to the export equations, all of which exhibit high explanatory power. There are more cases in which elasticity is lessthan unity than where it exceedsthat value. This feature was noted earlier in van Brabant (1985a,b) and attributed in part to the treatment of the phantom transportation charges explained in Eq. (1) that are imposed on some goods but not on others. Part of the reason here may also be that Hungarian import prices are probably c.i.f.; thus EWPs should be higher than CMEA *I I could have tested directly for the significance of a coefficient differing from unity, and this would have required a two-tailed test as well.

30

JOZEF M. VAN BRABANT TABLE 1

RELATIONSHIP BETWEENTR PRICE INDICES AND AB~~LLJTEPRICESIN NONSOCIALISTTRADE, 197 1- 1977 a (BASED ON HUNGARIAN DATA)

b coefficient *

t=

R2

SE

Exports Gasoline Dies& fuel Calcinated alumina Aluminum Autobus Televisions Light bulbs Slaughter hogs Pork meat Salami Dried sausage Poultry Fresh eggs

0.812 1.063 -0.045** 0.494** 0.592** -1.305** 0.326** 2.495** 1.296 0.854 0.718** O&8** -0.034**

8.82 30.59 -0.64$§ 5.49 5.55 -2.02*j 2.49* 10.45 6.73 6.84 8.80 6.86 -0.88$j

0.940 0.995 0.242 0.860 0.861 0.452 0.561 0.956 0.901 0.904 0.939 0.905 0.27 1

0.129 0.036 0.015 0.020 0.063 0.064 0.03 1 0.043 0.062 0.050 0.022 0.015

Imports Crude oil Diesel fuel Pine timber Timber Cellulose Raw cotton Raw phosphates Copper Zinc Caustic soda Synthetic rubber Polyethylene Polystyrene Cellulose-basedfiber Potassium fertilizer Nitrogenous fertilizer Phosphate fertilizer Newspaper Writing paper Wrapping paper

0.633** 0.77721 0.707** 0.938 0.67 I** 0.916 0.855** 0.022** 1.419** 0.495** 0.395** -0.238** 0.219** 0.502** 1.103 1.996 -2.323** 0.029** 0.500** 0.634**

12.91 36.59 9.95 8.23 10.88 4.05 17.46 0.28$ 9.01 7.94 5.56 -0.98$§ 1.36$ 7.96 2.28* 2.54* -6.194 1.79 6.64 8.13

0.97 I 0.996 0.952 0.931 0.960 0.766 0.984 0.788 0.942 0.927 0.862 0.168 0.275 0.927 0.510 0.564 0.885 0.390 0.898 0.930

0.107 0.029 0.070 0.07 1 0.046 0.143 0.057 0.004 0.126 0.068 0.040 0.080 0.085 0.046 0.146 0.207 0.169 0.212 0.048 0.056

0.123

’ No Durbin-Watson statistics are reported becausewith only seven observations the series are exceedingly short. * Based on a 95% confidence interval around 6, two asterisks (**) indicate that the coefficient differs significantly from unity. ’The significance, with five degreesof freedom, is better than at 99% for those with no marker; those with * are significant at 95%, ;t at the 90% level, and those with $ at less than 90%. A section mark (0) indicates that the coefficient has the wrong sign, although it may deviate from zero in a statistically significant way, as marked.

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prices and the full increase in EWPs should not be passed on to CMEA productsz2 It is interesting to note in this connection that for some commodities for which I could not obtain reasonable estimates when use was made of prices from Western commodity exchange quotations as reported in van Brabant (1985a,b), I have now generated rather good statistical relationships. These include raw phosphates, raw cotton, and zinc. For some of the final products including aluminum, crude oil, gasoline, and imported diesel fuel, however, the fits are now weaker. Though the coefficient for copper improves markedly, the equation remains basically very weak. Finally, there are a number of products that could be included in the earlier sample that can no longer be tested here because Hungary apparently does not trade those products with MEs. In the two earlier studies, I ordered products according to the degree to which a good fit was obtained and the proximity to unity of the elasticity parameter. The best relations were for fuels; fairly good ones were obtained for some metals and industrial raw materials, but the results for foodstuffs and agricultural raw materials were rather poor. This ordering does not hold for the products tested here, although it should be noted that the product m ix of the two samples differs substantially. Whereas, for the export products, good fits are obtained for fuels and foodstuffs though the coefficients for some of the latter diverge from unity, the relationships for raw materials and manufactured products are rather poor. On the import side, however, the equations for agricultural and some industrial raw materials are rather good. A tight fit is obtained for the fuels in the sample, but the elasticities diverge significantly from unity. Regarding the one common product, diesel fuel, the behavior on the export side is as anticipated, with the coefficient not statistically diverging from unity. On the import side, however, there is a strong presumption that the proportional changein the TR import price is consistently well below the proportional change in the averageEWP as experienced by Hungary. Imported quantities of diesel fuel are probably small amounts acquired on special terms in dollar relations, and this kind of small random transaction may explain why the fit is not very good. 5. THE RELATIONSHIPS FOR PRICE INDICES There should be no difference in the formation of CMEA prices over time, provided the correct reference prices have been identified and averaged according to the official principles. I have neverthelesschosen to test explicitly for a shift in pricing behavior following the transition from a fixed to a sliding ” But there is simply no methodological clarity about this feature.

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JOZEF M. VAN BRABANT

reference period in 1975 at a time of very sharp changes in absolute and relative EWPS.~~These large shifts may have motivated the CPEs to scrutinize more closely the application of the formal price-formation principles, particularly for well-defined goods for which WMPs are publicly quoted in international commodity exchanges.Using the symbols defined above, the model for such a structural break in the price-formation practice that I test is log PC, = a + ((b + c)D,)log PR, + ut

(4)

where DI is a dummy variable set equal to zero for observations prior to 1975 and to 1 otherwise. The coefficient c^then determines whether there is a break in the series.24The following relationships should hold: ifbO

so that (6 + c^)approaches 1 from below

(54

ifb>l+c^
so that (b + I?)approaches 1 from above.

(5b)

Table 2 lists the estimated 36 export and 26 import equations for the price indices. With six exceptions-pharmaceutical raw materials, diesel motors, and raw vegetableoils and fats among the exports and copper, sulphur, and PVC powder among the imports-all exhibit at least a fairly tight fit with more than adequate t statistics. However, not all t’s for the break coefficients are statistically significant. In addition to the above exceptions, the peculiar behavior of zinc needsto be stressed.But of 18 export and 16 import equations for which the test of a structural shift could be performed, 4 of each have statistically insignificant break coefficients, and several others are significant at only the 90 and 95% levels. Also, 2 each of the insignificant and significant coefficients of the export and 2 significant coefficients of the import equations have the wrong sign as per the criterion of Eq. (5). Nevertheless, use of this method of ascertaining the effect of the Moscow modification, in some contrast to what I reported before (for example, van Brabant, 1985a), does yield a plausible case. In spite of the fact that most equations are highly significant, many yield elasticity values that differ from unity, sometimes very markedly so. Unlike *’ On the strength of the argument set forth above, I have earlier presented regression results for two separatetime spans. This was done initially on an experimental basis because,somewhat surprisingly, the results obtained for many commodities did not show a tight statistical fit and, even when a good relationship was obtained, the estimated price elasticity parameters diverged significantly from unity. 24The superiority of the statistical relationships obtained in this way, when the time-shift parameter is significant, over the corresponding estimates based on Eq. (3) with separatetime spans is apparent immediately. The determination coefficients are generally higher, the Durbin-Watson statistics converge closer to the value two although measurable differences evidently remain, and smaller standard errors are obtained. In a number of cases,the elasticity parameters move closer to unity and become statistically more significant. The alternative computations are available upon request.

SOCIALIST

AND

WORLD

MARKET

PRICES

33

what one might expect a priori, for the series with the shift coefficient, the number of elasticities that differ from unity is higher than in the case of the short series; 12 as against 5 of 18 export equations each. For imports the proportions are reversed; 7 each of the 16 long equations and the 10 short equations. From earlier experiments based on comparing results obtained from time series with fewer observations, it is evident that the parameters in a number of cases are very sensitive to the time period of measurement, and additional observations would clearly be desirable to obtain more robust statistical inferences. On balance, there are slightly more commodities with an elasticity parameter less than unity. It is worth pointing out that several equations have very high values of the elasticity parameters even when the fit is very tight. These include those of bauxite, light bulbs, vegetables,pickles, paprika powder, and sausageamong the exports and cotton and aluminum products among the imports. These all are presumably rather “hard” commodities in CMEA trade and may therefore warrant some markup over the regular TR reference price possibly to compensate for the overvaluation of the TR relative to alternative dollar earnings. The opposite extreme is obtained for steel pipes and medicaments among the exports and coal, bar and profile steel, pharmaceutical raw materials, cement, and PVC powder among the imports. All have elasticity parameters well below 0.6 and some approximate zero even though a tight fit is obtained. The elasticity parameters of many of the equations approximate unity. At any rate, they come sufficiently close to warrant drawing a slightly stronger conclusion than I dared to advance in van Brabant (1985a,b). Though the Durbin-Watson statistics cannot really be properly interpreted owing to the small sample size, it is nonethelessnoteworthy that none of the equations has a value that even comes close to the upper tail of the test value at 15 observations of 5% probability, whereasthere are 10 export and 6 import equations for which the statistic is well within the lower tail and several would be in the zone for which the test is indecisive if sufficient observations were available. This suggests the potential presence of positive serial correlation. For the equations with good fits, if this tentative result is contirmed when data through 1985 become available, it will be necessary to test a slightly different model to circumvent the serial correlation problem. 6. AN INTERPRETATION AND DIRECTIONS FOR FURTHER WORK While the results obtained from the “ingrowth” model for Hungary exhibit few surprises, several elements can now be added to the preliminary conclusions presented in van Brabant (1985a,b). In the first place this concerns the range of products for which there appearsto be a strong relationship between world and CMEA prices, as per the CMEA price-formation principles. Also,

Exports Gasoline Diesel fuel Fodder seed Bauxite Calcinated alumina Metalhrrgical aluminum Hot-rolled steel pipes Cold-rolled steel sheets Aluminum products Autobus Finemetal mechanical lathe Metal drilling machine Pharmaceutical raw materials Men’s shoes Women’s shoes Leather and other gloves Televisions Refrigerators Medicaments Light bulbs vegetables Fruit Canned fruit Canned vegetables

4 2 2 2

1

2 2

1

2 2 2

1

2 2 2 2 2

1 1 1

3

1 1 1

Period”

1.075 1.059 0.892

1.540**

0.980 0.709** -0.939 0.874 0.474** 1.601

1.161 1.214 0.712**

0.949 0.671** 1.204

0.911

0.27 1

1.028

2.897 0.812

1.723 1.295

0.87 1

6b

6.89 7.13 0.883 17.55 14.38 20.96 -6.198 6.75 7.75 18.48 7.60 6.27 6.20 6.82

11.70 11.75

18.66 10.33 4.54 3.65 8.15 19.77 2.28* 5.33

tbc

0.003**

0.063**

-0.090

0.110”

0.003** -0.175** -0.079 0.014** -0.034** -0.050

Eb

0.57*g

13.36

-0.70$

0.14$ -3.35 -2.59* 0.42SQ -2.74**§ -5.34 5.86

tce

0.997 0.982 0.841 0.958 0.895 0.986 0.984 0.781 0.945 0.945 0.856 0.864 0.192 0.975 0.963 0.982 0.948 0.851 0.883 0.982 0.906 0.83 1 0.828 0.854

R2

0.738 0.682 1.428 1.058 1.552 1.199 1.341 0.717 1.742 1.349 1.634 0.257 0.980 1.208 1.587 1.498 1.510 1.275 2.369 2.133 1.145 1.294 1.740

1.113

DWd

THE RELATIONSHIP BETWEENTR AND DOLLAR PRICES, 197 1-I 984 (BASEDON HUNGARIAN EXEDRT AND IMEDRT PRICE INDICES)

TABLE 2

0.045 0.089 0.076 0.049 0.116 0.105 0.130 0.035 0.043 0.033 0.026 0.049 0.030 0.029 0.095 0.084 0.103 0.082

0.041

0.050 0.118 0.067 0.076 0.063

SE’

Imports Coal and anthracite Coniferous sawn wood Cellulose Raw cotton Sawn logs, etc. Pulpwood Iron ore Ferro-alloys Copper Zinc Synthetic rubber Polystyrene Hot-rolled bar and profile steel Foundry tinplate Aluminum products Pharmaceutical raw materials Cellulose-basedfiber

Canned tomatoes Pickles Red paprika powder Wine Pork for slaughter Salami Dried sausage Poultry Fresh eggs Canned meat Diesel motors Raw vegetable oils and fats

1

2

1 2

2 2

5 1

2 2 2

1 1

2

0.527** 0.805 1.087 2.397 0.663** 0.973 1.225 1.179 0.178 -0.941 1.266 1.286 0.366** 0.904 1.572** 0.393 1.089

-1.671 2.944 3.847 1.240 1.428 - 13.208 1.562 1.121 1.328 1.121 1.736 0.705 7.40 8.67 17.69 5.95 10.15 11.94 4.01 5.33 2.27 -1.12*9 6.56 11.30 10.89 4.55 10.37 4.23 17.97

-2.02?# 13.73 4.31 16.32 5.61 -4.228 8.59 13.19 7.54 5.03 1.67t 0.90$

1.46$ -6.57

0.794

0.020 0.009** -0.083

- 1.42$ 2.60**§ -1.82t 2.76* -3.10 -5.16

0.76$ -2.55* -2.52+

-2.37’ 3.658 -1.8lt 3.33 -3.57 -3.77 -3.92

3.48

-0.05 1 0.046 -0.017** 0.390** -0.106 -0.079

0.014** -0.027 -0.163**

-0.217** 0.026;. -0.045 1.406** -0.096** -0.035 -0.066

0.400**

0.887 0.986 0.993 0.930 0.928 0.947 0.819 0.882 0.415 0.914 0.899 0.958 0.938 0.923 0.931 0.837 0.991

0.843 0.959 0.913 0.987 0.910 0.646 0.966 0.974 0.885 0.761 0.260 0.093 0.724 2.379 1.678 2.039 1.205 1.873 0.608 1.511 1.178 1.022 0.956 1.438 2.337 1.338 0.637 1.068 1.475

1.021 1.240 2.382 2.321 1.266 1.223 0.779 0.915 1.407 0.610 0.247 0.264 0.078 0.050 0.03 1 0.151 0.047 0.052 0.109 0.107 0.022 0.172 0.104 0.054 0.018 0.09 1 0.083 0.036 0.034

0.182 0.033 0.135 0.037 0.073 1.238 0.862 0.037 0.065 0.09 1 0.169 0.171

1 1 7 2 2

Wrapping paper Beer Powdered milk Sulphur PVC powder 0.996 1.242 -0.07 1** 1.007 0.601**

-0.130 1.088 0.246** 0.959

66

12.66 9.30 -2.95**$ 3.47 3.32

-2.07$§ 5.49 5.30 15.26

fbc

-0.035 0.043**

-0.020

0.133**

tb

-2.43* 5.09$

-2.01 t

14.19

t=c

0.986 0.946 0.721 0.601 0.581

0.989 0.791 0.779 0.987

R2

1.798 1.895 2.328 0.512 2.406

2.137 1.599 1.367 1.946

DWd

0.042 0.059 0.011 0.243 0.072

0.029 0.070 0.078 0.035

SE’

n 1: 1971-1984; 2: 1975-1984; 3: 1971-1979; 4: 1975-1982; 5: 1976-1984; 6: 1971-1982; and 7: 1975-1980. b See Table 1. Two asterisks (**) indicate that the single or combined coefficient is significantly different from unity at 95%. c SeeTable 1. The section mark (5) indicates that the single or combined coefficients has the wrong sign. d Durbin-Watson statistics are listed for illustrative purposes only. The maximum number of observations (14) falls short of the minimum required to utilize the standard tables. e Standard error.

6 2 2 1

Nitrogenous fertilizer Phosphorous fertilizer Cement Writing paper

Period”

TABLE S-Continued

3

F $

s 2

g

5 #

SOCIALIST AND WORLD MARKET

PRICES

37

some of the earlier tentative results can now be strengthened by looking at three questions for the experiments performed thus far: (i) Is there sufficient evidence to prefer EWPs over WMPs? (ii) Can one generalize apparent TR price-formation practices? (iii) What further tests could strengthen the conclusions that can be drawn? Regarding the first question, it is necessary to recall that any answer is strictly contingent on having access to a substantial degree of commodity homogeneity between the samples studied earlier in van Brabant (1985a,b) and those reported here. Unfortunately, there are only three exports, gasoline, diesel fuel, and possibly metallurgical aluminum, and five imports, coal, cotton, iron ore, copper, and zinc, that can be compared directly. Recall that the earlier relationships for nonfuel m inerals and metals were not very robust in contrast to those for liquid fuels which were excellent. For the two liquid fuel products common to the two estimations, the statistical indicators in the ingrowth model are inferior, and sometimes quite markedly. For coal, I now obtain a much better relationship, though the coefficient falls well short of unity. The results for the metals and nonfuel m inerals remain m ixed. For aluminum all indicators deteriorate slightly, and the same is true for iron ore even though a higher coefficient of determination emerges,but some of the indicators for copper and zinc are much better than what I obtained earlier. The picture for cotton is mostly worse than the earlier estimates.These m ixed results are somewhat disconcerting. Whether one can generalizeon the basisof the evidencefound so far depends critically on the fraction of total trade covered and on more reliable information as to the factors that influence actual price setting in the CMEA. There is scattered information for a few individual products (Brabant, 1987, Chap. 4), but it is too disparate to permit the derivation of trends. Regarding coverage of the commodities that could be studied, very laborious computations are required for technical reasons.25Since such key products as oil and gas in imports and key machinery products in exports could not be tested in the ingrowth model because no comparable data for CMEA and thirdmarket trade are available, the size of the sample studied here is not likely to exceed 10%of CMEA trade and only a fraction of the hundreds of commodities traded. The representativenessof the smaller number of commodities discussed earlier (as in Brabant, 1985a) is higher, though much more skewed, since it includes key imports such as oil and natural gas. Of course, the sample of absolute prices studied in Section 4 is even narrower than that for the commodity price indices. Conclusions, therefore, must of necessity remain tentative at this stage. *’Becauseof the lack of detailed information and different nomenclatures used in the commodity price statistics and the trade returns, I have been unable to establish the precise share of trade covered by the commodities studied.

38

JOZEF

M. VAN

BRABANT

CMEA prices appear to be derived in part on the basis of actual EWPs either because they offer a handy reference to WMPs or because they are realistic indicators of alternative trade opportunities in world markets. There is undoubtedly more than ample evidence to reject the proposition that EWPs or WMPs do not influence TR prices at all. But the transmission is more direct and faster for some goods than for others, including rather surprisingly a number of industrial raw materials. For others, especially the very hard goods that are not easily documented through quotations of commodity exchanges, the evidence suggeststhat TR prices contain an element of compensation for the degree of hardness. That this is not the case for the “super hard” goods, such as crude oil, petroleum products, and natural gas, which constitute such a large component of trade stems perhaps more from the nature of the CMEA than from sheer adherence to the Bucharest pricing principles on the ground of any doctrinal or systemic commitment. There is very little doubt that for many products, elasticities diverge from unity for systematic reasons that are unrelated to averageEWPs or WMPs per se. Similarly, though there is evidence of better adherenceto the formula since 1975, it is admittedly not very persuasive in many cases. Except for the passing of time, which hopefully will steadily augment the observations for Hungary and possibly the surfacing of some recent volumes of K4.4, any further refinement and extension of the tests reported here could usefully envisage comparisons of CMEA prices with East-West unit values or of unit values in both CMEA and East-West relations. The first case can be studied for Hungary. The latter could be applied to several other CPEs, notably Poland and the USSR. This would permit longer and more diversified time series also for larger CMEA participants than Hungary. Such an exercise would not, of course, give rise to major inherent or conceptual obstacles once the vexing problem of unit values, the perhaps even more involved task of ensuring comparability of identified goods in Western and Eastern sources, and the rather intractable problem of the stability of differences between CMEA and world commodities have been resolved. Although these are knotty issues, it might be worth the effort to explore how much headway could be made with these data sources. REFERENCES Ausch, !Gndor, Theory and Practiceof CMEA Cooperation.Budapest: Akadhmiai Kiad6, 1972. Ausch, Shdor, and Bartha, Ferenc, “Theo!ztkal problems of CMEA In&trade Prices.” In Tam& FGldi and Tibor Kiss, Eds.,Socialist World Market Prices,pp. 101-125. Leyden: Sijtho$ Budapest: Akadhmiai Kiad6, 1969. Hewett, EdwardA., Foreign TradePrices in the Councilfor Mutual EconomicAssistance.Cambridge: Cambridge Univ. Press, 1974. K4A, Klilkereskedelmidrstatisztikaiadatok,1957-1977.Budapest: Khponti Statktikai Hivatal, 1978.

SOCIALIST AND WORLD MARKET

PRICES

39

Konstantinov, Yuriy A., Mezhdunarodnaya valyutnaya sistema stran-chlenov SEV. MOSCOW: Finansy i statist&a, 1982. K&es, Andr& The CMEA Countries in the World Economy: Turning inwards or Turning outwards. Budapest: Akademiai Kiad6, 1985. Koz’menko, Vasiliy, “Transpottnyy faktor i tsenoobrazovanie na rynke strati SEV.” Vneshnyaya torgovlya 1:25-28, 1986. KSE, Ktilkereskedeimi statisztikai t%&tyv, . Budapest: Kozponti Stat&ikai Hivatal, annual. Manx, Paul, Postwar Pricing and Price Patterns in Socialist Foreign Trade (1946-1971). Bloomington, IN: IDRC, Indiana University, 1972. Marrese, Michael, and Vadous, Jan, Soviet Subsidization of Trade with Eastern Europe: A Soviet Perspective. Berkeley, CA: Institute of International Studies, 1983. Poznanski, Kazimietz, “Competitiveness of Polish Industry.” Ms. Washington, DC, Oct. 1985. Poznanski, Kazimietz, “Competition between Eastern Europe and Developing Countries in the Western Market for Manufactured Goods.” In Joint Economic Committee, Congress of the United States, East European Economies: Slow Growth in the 1980s. Vol. 2, pp. 6290. Washington, DC: U.S. Gov. Printing Office, 1986. Rutkowski, Jemy, “Stabilnosc i elastyczno& ten w obrotach wzajemnych kraj6w RWPG.” Handel Zagraniczny 22, 1:3-8, Jan. 1977. Shanina, Valentina A., Transportno-ekonomicheskie svyazi stran-chlenov SEV. Moscow: Nauka, 1976. Spa&k, Petr, “MBHS: i%kladni &trek platebniho a uveroveho mechanismu Elensljch stat8 RVHP.” Finance a’Ufer 34,7:433-440, July 1984. Spa&k, Petr, “Posilit Uohu kolektivni m&y.” Svet Hospod&stvi 26, 17:1, 1985. Teichmanowa, Eufemia, “Koszt tmnsportu w obrotach towarowych miedzy krajami RWPG.” Handel Zagraniczny 21, 11:37-39, Nov. 1976. UNCTAD, Proceedings of the United Nations Conference on Trade and Development, Fourth Session, Vol. I, Reports andAnnexes. New York: United Nations Publ. no. E.76.II.D.10, 1976. van Brabant, Jozef M., “On the Determination of Trade Prices in Eastern Europe.” Ms. New York, Sept. 1983. van Brabant, Jozef M., “The USSR and Socialist Economic Integration: A Comment.” Soviet Stud. 36, 1:127-138, 1984. van Brabant, Jozef M., “The Relationship between World and Socialist Trade Prices: Some Empirical Evidence.” .I. Comp. Econ. 9, 3:233-25 1, 1985a. van Brabant, Jozef M., “World Prices and Price Formation in Intra-CMEA Trade: SelectedEmpirical Evidence.” Osteuropa- Wirtschaji 30,3:163-180, 1985b. van Brabant, Jozef M., Adjustment, Structural Change, and Economic Eficiency: Aspects of&ionetary Cooperation in Eastern Europe. New York: Cambridge Univ. Press, in press.