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Domestic concentration, foreign trade and economic performance
International Journal of Industrial CYganization 3 (1985) 1-19. North-Holland
DOMESTIC CONCENTRATION, FOREIGN TR4DE ECONOMIC PERF6RMANCE
AJPiD
Manfred NEUMANN, Ingo BABEL and Alfred HAID* Universityof Erlangen-Nhberg,
O-8500 Nfirnberg I, FRG
Final version received September 1984 In a competitive fringe model the impact of concentration on price-cost margins can be shown to be ambiguous. Therefore, looking at the relationship “between rates of return and concentration ratios is not sufficient for perceiving collusive elements. In this paper a model is set up that permits discerning collusion more clearly. Additional evidence concerning collusion can be gained by analyzing the influence of exports and imports on pric+cost margins. Applying this approach to West Germany yields the result that the collusiveness prevailing in domestic markets has not visibly decreased in spite of a substantial increase in foreign trade.
1. Introduction It seems to be widely believed that the more recent d.evelopment of foreign trade has contributed to intensifying competition in most industrialized countries. Two pieces of evidence are usually adduced for drawing this conclusion. First, in most structure-performance studies where foreign trade variables have been included the import ratio d:splayed an adverse effect on price-cost margins [see, for example, Esposito and Esposito (197 l), Neumann, Bobel and Haid (1979), Jacquemin, de Ghellinck and Huvencers (198C), Ravenscraft (1983), Martin (1983)]. Second, over the last decades the share of imports in most industrialized countries has substantially increased. From these observations it seems to follow that anticompetitive effects residing in domestic concentration have been counteracted by competition of foreign supply. Moreover, since domestic concentration has increased only at a glacial pace and has apparently been checked in many countries by anti-trust policies the conclusion seems to be warranted that the overall intensit:r of competition has increased [for the U.S., see Shepherd (1982)]. This inference, however, appears to rest on shaky ground. Looking more closely at the evidence will reveal that the relationships involved are a good deal more *We are grateful to Bernard Gahlen, Stephen Martin, Dennis C. Mueller, Roger Sugden,
and to the anonymous referees of this Journal for helpftil comments and suggestions. We- are also indebted to participants in workshops of the International Institute of Management, Berlin, and the OECD. This usual disclaimer applies. 0167-7187/85/$3.30 0 1985, Elsevier Science Pd~lishcrs
R.V. (North-Hollamc’)
M. Neu-
2
et al., Concentrationand foreigts m& impact
complex. To make the claim of an increased intensity of competition convincing other pieces of evidence must fit into the picture consistently. It will be shown in this paper that such is not the case. The additional evidence we wish to call upon is concerned with the explanatory power of alternative measures of concentration and the interpretation of the statistical impact of foreign trade variables on primost margins. It is well known that in a price leadership or competitive fringe model the g firm group are favourably affected by the price-cost margins of the lea leading firms, i.e., the concentration ratio, cumulative market share of provided the leading firm group acts as a. cartel [Saving (1970)]. It has furthermore been shown that in? a Coumot model it is the Herfindahl index which, instead of the concentrai ion ratio, has a favourable impact on pricecost margins [Cowling and W erson ( 1976)]. Looking at these polar cases suggests that the predictive ower of the concentration ratio and the IIerfindahl index should be ;:smpared. It could further be investigated whether collusitin or competitive behaviour prevails to a larger extent. In the next step it should be examined whether the prevalence of either mode of behaviour has changed more recently. The remainder of the paper is organized as follows. In section 2 the basic model is set up along the lines suggested by Encaoua and Jacquemin (1980), Clarke and Davies (! 982) and Neumann (1983). The implic&ians regarding concentration, imports and exports, and demand shifts are discussed in sections 3 through 5. In the subsequent section 6 empirical estimates are given for West German industries. They may, first, serve as an example for the testing procedure outlined above. ‘I&y are, however, also tieresting on their own because they throw some light on the recent development of conipctition in Germany.
2 The basic model
We shall start from the presu=mption of a market structure characterized by a group of dominant firms and a competitive fringe where both kinds of firms are assumed to behave differently. This point of vantage has been chosen because the dominant firm model constitutes a general case from which cases of homogeneous firms with the same type of behaviour can be derived as special cases. Let total demand Q =zf= 1 qi +x be served by k large firms and a cokmpetitive fringe. Firms belonging to the fringe act as price takers. Profits of’any one of the leading ohms are given by ni =P(Q)qi - C(qi), where AQ) is the inverted demand function. Profit maximization requires P+
4i
dPld4i= ci
9
M. Neumann et al., Concentration and foreign %de tmpzt
where ci : =
denotes marginal
c’(qi)
dPldqi
(
=fdPldQ)1+ i
costs,and
dqj ld4i
+ (dxldPNdPldqi)
j=l
) l
(2)
/
Subsequently, the following definitions will be used: E:= -
tl:= ~PIMWP~~ (PIQMQI~P~
Si : = qi /Q,
Zij : = (4i /4iKdqj
ai t =qi/(Q-x),
/dqih
E and q being price elasticities of total demand and supply of the competitive fringe, respectively. ai and si are market shares. The supply elasticity q of the competitive fringe will be assumed to be positive. Aggregate marginal costs of the competitive fringe are thus increasing. The coefficient Zii indicates the elasticity of reaction of rival j as expected by firm i. It is constrained by - 00 SZ& 1. In the Coumot case Zij= 0 for all i, j. On the other hand, perfect collusion implies Zi 1. Perfect competition in a particular market would imply that x =O and z f i dqj /dqi = - 1. That entails & +i ajZij= - ai. Any decrease of supply of one of the firms would be. immediately supplanted by augmented supplies of the rivals. For simplicity it is henceforth assumed that zij =zi for all j. Hence firm i expects all its rivals, irrespective of their individual size, to react in the same way upon an increase in the quantity supplied by firm i. On the other hand, it appears reasonable to assume that the zi’s differ according to the relative size of firm i. A large firm can expect a one percentage increase in its supply to be perceived by the rivals and will therefore elicit some reaction. A smaller firm can expect that a one percentage increase in its quantity supplied: goes largely or even completely undetected and will therefore evoke a comparatively smaller, or even no reaction at all, of the rivals. Thus it will be assumed that Zi=z,(uJ with &(a,) > 0. The elasticity of conjectural reaction held by firm i is thus an increasing function of its market share.’ To take the most simple form of the suggested relationship let j
Zi(Ui)=f+bQi,
=
00,
(3)
where { reflects institutional arrangements conducive to collusive behaviour, such as a legislation permitting cartels, price information agencies, etc. Eq. (2) can be re-arranged to yield dPldqi =
-WWQ~U
+U - ai)/aJzJlf 1+ (~~IE)(xIQ))*
‘We a--e indebted to Stephen Martin for suggesting this relationship.
M. Neumann et al., Concentration and foreign trade impact
4
Upon insertion into eq. (lj one obtains the individual price-cost margin mi:=(p-Ci)/p=K[U,
(4)
+Zi(l-ai)]/(E+dl-K)),
where K: 7 (Q - x)/Q is the k-firm concentration can also be written as
ratio. Since Qi= Si/K eq. (4)
(9
mi=(Si(l-zi)+ZiK)/(E+tf(l--K)).
From the individual price-cost margin mi one can proceed to the average price-cost margin of the k large firms. 2.1. Industry price-cost margin By definition, given in (4), p(1 -mi) = c, . Multiplying both sides with ai and summing over k ~~ITIISyields p(1 -xi mt ai) = & ci ai : = c where c is a weighted mean of the marginal costs of the Jcleading firms. Upon insertion into eq. (4) one obtains m:=(p-c)lp=Cmiai=K
H,+ i
i=1
i
Ziai-
-m II@+tt(l
k zi$
i =1 =
,
where H4: -xf= I a: is the Herfindahl Substituting (3) yields m=(K/(E+q(l
In the
~phtl
-K)))
(l-[+a)H,+[-ai
index of the large firm group.
i a:
i=l
1 .
(6)
case of zi = (5= constant for all i eq. (6) boils down to
m=K[H,(l-Q+fl/(e+q(l-K)).
I
(6)
The special cases derived in the prior literature can be easily recovered from (6’). Setting 5 =0 gives tte formula derived by Encaoua and Jacquemin (1980, p. 99). Setting [ = f yields the formula of Saving (1970). The formula for the pure Coumot case where all firms are on an equal footing which has been propounded by Cowling and Waterson (1976) can be found by setting K = 1 and 5 ~0. The slightly more general case invoked by Clarke and Davies (1982) emerges by setting K = 1 and [ > 0. In the general case, given by (6), the additional term ---oci a; is invoked. It might appear that in this way a dependence of the industry price-cost margin on individual market shares could be established. It does not, however, appear reasonable to assume a dependence of m on any one of
M. Neumann et al., Concentration a.& foreign trade impact
5
the individual market shares because an increase in any Qi implies a corresponding change in at least one other Qj. Yet individual market shares can nevertheless be shown to exert an influence on the industry price-cost margin. To capture this influence it is necessary to introduce an additional summary measure of the size distribution of the large firm group besides the Herfindahl index. We wish to propose a measure of asymmetry defined by
which is a pure number. Generally azH,. The coefficient equals unity in the case of monopoly. It coincides with the Herfindahl index in this case and in the case of equally sized firm~.~ If &a: = (aHjf)li2 is inserted into (6) one obtains m =(K/(E + ~(l- K)))[( 1- 5 + o)H, + [ - aa1/2H;3/‘],
(7a )
According to this equation m depends on the concentration ratio K :snd the Herfindahl index Ha of the leading firm group, given the coeficients QC, O,C,q and c. Since ai = si/K the Herfindahl index H, can be alternatively written as H,=H/K2 where H:=& s2i is a truncated Herfindahl index confined to the set of k large firms. It might be noted that the Herflndahl indices ordinari1.y employed in empirical work are also truncated indices because data on the stratum of smaller firms are usually not available. Substituting H/K2 for H, in eq. (7a) yields
(7b) According to this equation the average price--cost margin of the leading firms depends, inter alia, on K, H/K and H312/K2. 2The coefiicient a is devised in analogy to the coefficients proposed by K. Pearson [see Yule and Kendall (1973, p. lS9)] which utilize the second and third moments of a statistical distribution. Some examples may be useful to demonstrate the properties of the a-measure of asymmetry. Example no.
1
2
3
4
a,
0.80 0.15 0.05 0.665 0.516 0.904
0.70 0.20 0.10 0.540 0.352 0.787
0.50 0.30 0.20 0.380 0.160 0.467
040 0.40 0.20 0.360 0.136 0.396
a2 a3
RI c I.a31 a
it can easily be verified that a? H, and ct= H, in the case of equal market shares.
6
M. Neumann et al., Concentration d_foreign
trade impact
Interpreting these relationships requires a good deal of care since it is not evident a priori whether K, H,., and H are endogenous or exogenous variables. Along the lines suggested by Clarke and Davies (1382) the size distribution within the leading firm group, as reflected by K,, can be shown to be endogenous. In the case Zi = 5 = constant one obtains3
H, = l/K + (1- k((E- O/(1- 0)2(v,21k), where E = [s -+q( 1- K))/K and v, denotes the coefficient of variation of c,/ k c i = 1 ci reflecting the dispersion of the level of marginal costs. The size distribution ot the large firms, as summarized by Ha,thus depends on the dispersion of marginal costs, on various elasticities, on the number of leading firms, and on the k-fm concentration ratio. An analogous relationship can be presumed to exist in the more general case of Zi depending on market shares. Obviously endogeneity also obtains for the coefI%ient of asymmetry, 01. A similar reasoning applies to the number of dominant furns and their individual and cumulative market shares. The existence of a set of dominant firms which exercise a conduct different from the one found within the competitive fringe reflects barriers to entry to the privileged circle of leading firms. These barriers may be due to vari~_~s reasons, one of them being exclusive knowledge of, or access to, superior techniques which gives rise to differing cost functions. Explaining price-cost margins therefore would ideally require to invoke fundamental variables such as cost functions, conditions of entry, factors conducive to collusive behaviour, and demand functions. Those variables however, are very hard to come by and sometimes they are simp:v nonobservable. Therefore, analyses in the wake of the structure-performance paradigm stick to investigating the first order maximum conditions, as given in our case by eqs. (7a) and (7b). These equations should be interpreted as describing relationships between endogenous variables which are ultimately determined by unobservable factors outlined above. Note that also the demand elasticity s and the supply elasticity q are endogenous unless some very specific functional forms of market demand functions and marginal cost functions of the competitive fringe obtain. 3Let E = [E+ q( 1 - K)]/K and ii = [ = constant. In this cas.: eq. (4) can be
(p- Ci)/p= (ai+ c(1-
Ui))/E,
written
which yields
Q:= -[,/(l-i)+(E/(i-C))(l-Cilp). Summing over k and re-arranging gwes P=E~Ci[k(E-~-(1-5)]-‘. Substituting this into (i), which is then squared and summed yields H,= l/k+(l -Ic((E--c)/(l -L#($/k).
(9 (ii)
M. Nwnann
et al., Concentration umi~h-eign W&
impacr
7
Relationships like eqs. (7a) and (7’0) can be considered as lJypcthss ;.s suggesting that maximizing behaviour obtains, that a dominant firm grcrup facing a competitive fringe persists (due to differential entry opportunitizsj and, finally, that collusion_ is viable. Cocfronting these hypotheses with empirical evidence wil; show which one of them can withstand empirical testing. The approach can be further clarified by looting at conjectural variations in particular. Recall that the Coumot solution, perfect collusion, and pure competition, respectively, emerge as special cases, as shown above. It cannot, however, be expected that these special cases will be observed in practice. If the outcome during some period of time is observed some mixt-ure of com*‘ _ptiti*. - ktia;~tii atid collusion can *beexpected to obtain. The explanation of such an outcome notwithstanding, a descl-iption of the outcome can rn;,t conveniently be achieved by invoking the notion of conjectural variations. In this spirit eqs. (‘a) and (7b) are set up by regrouping the variables in such a way that the impact of collusion can be isolated, as shown in detail subsequently. 2.2. In&dual
price-cost margins
One may finally return to the individual price-cost margins. From p( 1- m) = c and p( 1 - mJ = ci one obtains
Hence, given ci/c, the individual price-cost margin is linearly dependent on the industry price-cost margin, and for that matter, on the variables determining m. C&n m, the individual price-cost margin is the higher the lower c’i/G 3. Conioentration and industry prkwmst
margins
Regarding the relationship between concentration and industry price-cost margins we suggest to consider two kinds of models, labelled model A and model B, which are represented by eqs. (7a) and (7b), respectively. Subcases A’ and B’ arise if zi = [ = constant, and Q= 0. In the case of model A one obtains am/aK=(m/K)i1+rlK/(s+11(1-Kj)], dm/aH, = (K/(e i- q( I- K)))[( 1 - [ + a) - (3/2)aa1i2Hti2].
(9) (10)
Hence, given H,, dm/dK is unambiguously non-negative, its site dcpendicg, among other things, on m which, in turn, depends, inter alia, on (. As can be
M. Neumann et al., Concentration
8
and foreign trade impact
seen from
the larger (5 the higher m. I-Ience a change of the institutional framework in favour of collusive behaviour gives rise to a higher price-cost margin. Whereas the impact of K on m isunambiguous, given Ii,, the influence exerted by H,, given K, can go either way, provided the asymmetry between the leading firms is sufficiently large. In the subcase A’ one could expect dmjaH, > 0. In the case of model B one obtains
amid
= ( i - 5 + ~T)/(E + q(i - K)) > 0,
am/a(H312/K2) = - OCC~/(C + tj( 1 -K)) < 0.
(12) (13)
In the model where all three variables are contained simultaneously am/aK is certainly positive if [ 20, i.e., in the Cournot case and if collusion obtains. It may become negative if CCO. Thus a negative am/X definitely implies competitive behaviour. The same reasoning applies in the subcase B’ where a==. If, a,,ernatively, the presumed market structure, characterized by a set of dominant firms facing a competitive fringe, does not exist, i.e., if K = 1, eqs. (7a) and (10) imply that h/aH, =am/aH may still be positive or negative unless 0 = 0. Provided 0 =0 ii is positive. 4. Imports and exports Generally, imports can be identified as part of the competitive fringe. That may not be true in some cases but it can safely be assumed to hold on average. If that is true an increase of the import ratio exerts an adverse influence on price-cost margins of dominant firms. The increase of the import ratio may be due to trade liberalization which gives rise to an outward shift of the supply curve of the competitive fringe, It may also be due to innovations adopted by foreign competitors which lower their marginal costs. In any case, within the framework of a homogeneous market, prices of domestic firms and foreign firms must be equal after an equilibrium has been attained. Turning to exports we note that total sales Si=pqi fp*q” of firm i are composed of sales to domestic and foreign markets, respectively, starred variables denoting foreign prices and quantities sold to foreign markets. Let
M. Neumann et al., Concentration and foreign trade impact
9
A:= JJ*~:/& denote the share of sales going to foreign markets and mi : = (pqi - cqi)/pqi and rnr : = (p*@ -cqf)/p*gf price-cost margins of firm i obtaining for sales to domestic and foreign markets, respectively. The firm’s total price-cost margin is then given by riTti : = [(pqi - cqi) + Cp*& - cqr)]/Si [see also Pugel(1980)] which can also be written as @=mi(l-A)+Amf. Hence an increase in the export ratio iz yields &iiilait= -(??li -mF)sO
as
Wli$??$.
If domestic markets, in consequence of trading costs and barriers to entry for foreign competitors, are relatively protected so that collusion is more feasible at home than abroad sales to foreign markets are exposed to more competition and hence rnf cm,. That implies a negative impact of an increased export ratio on a domestic firm’s price-cost margin fii* If, on the other hand, a firm is exposed to substantial competition at home and is selling to foreign markets where a higher degree of collusion prevails the difference mi - rnf may be negligible or even negative. That implies an insignificant or even positive impact of an increasing export ratio on the firm’s price-cost margin. 5. Demand shifts Shifts of demand curves which can be expected to occur during the business cycle will ordinarily entail changes of the price elasticity of demand, e. For example, given a linear demand schedule, a parallel shift of the demand curve to the right yields a decrease of e at a given level of price. More generally, any shift of a linear demand curve to the ‘right, except an anti-clockwise rotation of the demand curve around the choke-off price? entails a decrease of E. The impact of E on mi is given by 8mi /as =(cJc)am/& Using alternatively eq. (7a) or (7b) yields ampi5= -m/(6 + rj(1 - K)).
Hence during a business cycle upswing, which entails a decrease of E, mi should be expected to rise. As can be verified frown
a2m/ad4(= -(l/k + ul(1- K)))Cam/aK + mq/(c+ q(1- K))], thz rate of increase of the price-cost margin during a business cycle upswing
M. Neumanroet al., Conecc;;l’i&anand foreign trade impact
10
will be the more pronounced the higher the concentration ratio if the term in square brackets is positive. Utilizing model A, one finds
a2ttqaeaM,= --(am/aw/~~+vl(l -WJ. Hence, an analogous relationship holds for the %cr 4s~riYr ! i,:dex, provided am/aH, > 0. The increase of m upon an expansionary demand shift which lowers demand elasticity gives rise to higher prices if marginal costs are nondecreasing. Given non-decreasing marginal costs, prices will rise the more the higher concentration. Conversely, recessionary demand shifts which engender an increasing demand elasticity entail lower prices, the decrease in price being the larger the higher concentration. 6. Emphical analysis Since data on marginal costs are not available the model, as given ,y eqs. (7a), (7b) and (8), cannot immediately be applied. First cf a!!, ihe mo;,iel has to be modified such that empirically observable variables can be used. In the first step, marginal costs will be replaced by average labour costs. That yields a price-cost margin of firm i, FCMi
:=
vaiue added minus payroll of firm i value added of firm i
l
To expound the relationship between PCMi and the price-cost margin of the theoretical model, ml, we Erst note that the ratio of average costs to marginal COStS, ,Vi, indicates increasing, constant, and decrer?-.g returns to scale, respectively, as vi $1. That implies
where pqi denotes value added. Average costs are given by .
with w = wage rate, f.i = labour input, Ki =capital stock, and ;ip, =(r+ 6) user cost of capital, depending on the interest rate r, the rate of PC/k= depreciation 6, the value of capital goods, pK9and its rate of change PK. After some re-arranging one obtains
M. Neumann et al., Concentration and foreign trade impact
11
Hence PCMi depends on economies of scale, on the capital-output ratio and on those variables determining fii. We shall proceed by relating tii to equity capital because the owners of a firm can be expected to be concerned primarily with the rate of return on equity.4 Denote equity of firm i by Ei, and note that total assets of firm i, PKKi =Ei + Di , equal equity plus debt. Multiplying (16) by pqi/Ei yields the rate of return on equi,ty of firm i, RREi :=(pqi --WLi)/Ei =(l/(l
-LEF))[A+(l
-vi +vitii)(pqi/p~Ki)],
where LE 6 = Di /( Ei + Di). The rate of return thus depends, first, on.. those variables discussed in connection with fii, and for that matter m, second, on leverage LEF, on the ratio of value added to total assets, and on economies of scale. Equity must not be evaluated at current market prices of shares but at historical book value. If equity would be evaluated at current market prices of shares any influence attributable to market structure variables would be absorbed by the concomitant rise of the denominator in (17). To account for the influence of economies of scale we used two variables, i.e., the size of the firm (measured by natural logarithms of total assets) and the rate of growth of sales. Ordinarily one would expect size and economies of scale to be positively associated. If that is true size should exert a negative influence on R REi , as given by ( 17). Let Si denote size and assume V; (Si) > 0. That implies dRREJd& = -(l/(l
- LEK))(PqilPKKijr(l -fiib:tsiJ
Contrary to what might be expected, an inverse relationship between RRE, and size definitely points to the existence of economies of scale, albeit those economies which are not, or cannot be, fully exploited at the actual volume of sales. If economies of scale are associated with size the larger output the greater the ratio of average to marginal costs, given the volume of output. On the other hand, fast growth of output gives rise to higher rates of return because existing economies of scale can be more completely exploited. 41n view of the recent debate concerning accounting rates of return [Fischer and McGowan (1983) Fischer (1984)] we would like to submit the following observation. Whatever the relationship between the accounting rate of return and the economic rate of return of a particuIar firm at a particular point of time may be, any statistically significant association between accounting rates of return and market structure variables established for a cross-section of firms suggests that an economically meaningful relationship exists. This inference is of course subject to subsequent disproval. On the other hand it would not appear legitimate to draw inferences regarding the existence of monopoly power horn the observation of accounting rates of return exceeding some average level at a particular point in time in the case of an iki~viduul lirm.
12
M. Neumann et al., Concentrdion and foreign trade impact
’
The foregoing reasoning is predicated on the existence of monopoly power by which output is restricted. The relationship between unexploited economies of scale and monopoly power can be elaborated as follows. The present value of cash flow of a firm is given by Ai=J
0
CpF(Ki,Li)-wLi-(SP~-dg)Ki]e-rcdt,
where F(Ki , Ei) is the production function suitable for a particular SUE of the firm. Assuming a stationary equilibrium with constant prices of output and capital goods, one obtains
Given a homogeneous production function, UiF= F,,Ki + F,,Li is satisfied, where Fgf and FLi denote marginal productivities of capital and labour, respectively. Inserting the first order maximum conditions, p(l-fii)FLi=w
and
P(l-~,)F,,=(64t)pls-IjR,
yields
Since a financial equilibrium requires Ai =pK Ki it implies 1 -Vi + fii 02 = 0. If the production function is homothetic with variable economies of scale, which give rise to a u-shaped average cost curve, vi = l/( 1 -tiJ is an increasing function of C&. Increasing monopoly power yielc3 aapoint fWhrt+ up on the declining branch of the average cost curve w?.ere ui is larger. Hence, given a positive association with the size of the firm .:ind economies of scale entailed by the kind of technique suitable for a paricular size class, unexploited economies are increasing with monopoly power. To complete the model two further variables must be included. It will first be noted that the rate of return on equity also depends on risk which can be depicted, as suggested by the capital asset pricing model, by the variance and covariance of rates of return in a way discussed in Imore detail elsewhere INeumann, Babel and Haid (1979)~. Second, beyond the theoretical framework discussed so far ye have to account for the fact that in many industries heterogeneous products are being sold and that heterogeneity may give rise to barriers to entry. Moreover, heterogeneity ordinarily entails advertising expenses by which price-cost margins are augmented. For depicting heterogeneity we used the number of trademarks, as registered by a firm with the German Patent Ofice, deflated by equity, i.e., the clenominator of eq. (17). The various explanatory variables are hypothesizeI. to exert an influence
M. Neumann et al., Concentration andforeign trade impact
13
on the rate of return which must be depicted P11:3 non-linear equation, not particularly well suited for estimation. For tht; Furpose of a regression ar,.ysis, however, it can be approximated by a linear equation using a P .lor series, the coefficients of which denote the partial derivatives with spect to the variables. 7. The evidence The data base covering 196!5-1977 pertains to 283 West German joint stock companies for which data were available. Individual firm data were collected from annual publications by Hoppenstedt Verlag and from informationa which were made availab1.e to us by the German Patent Office. Industry data were drawn from various official publications. Definitions of the explanatory variables are summarized in the appendix. We report two kinds of results. The first regression, reported in table 1, applies to the average values of the various variables over the total timespan from 1965 through 1977. The second regression, reported in table 2, applies to individual years. Regarding concentration we did not use average values but data on 3-firm concentration ratios and the Herfimdahl’index of 1977, as published by the Monopolkommission, because the publication of the Herfindahl index started only with data for 1977. To ensure comparability between the regression results obtained for C3 and the Herfindahl, respectively, we also took the data on C3 only for 1977? ‘The results reported in table 1 confirm the validity of the theoretical approach. It will be noticed that v/A( =value added over assets) is only significant if the size variable S is deleted. The disappearance of significance for v/A and also the relatively low level of significance attained by the size variable can probably be attributed to multicollinearity since the simple coefficient of correlation between I+4 and S is - 0.61. Concentration. Using C3 and the Herfindahl index alternatively [column (I), (2) and (3) in table I] we find both to exert a significantly favourable influence on the rate of return, the Herflndahl yielding a slightly higher level of significance. Model A [columns (4) and (5) in table 1-j which uses C3 and H,, iapproximated by H/( C3)2, simultaneously yields a significant coeffkien t for C3 whereas Ha turns out to be insignificant, the latter result being in Iine with theoretical expectations. Model B [columns (6) and (7) in table l] fully confirms theoretical expectations. ’ The negative sign associated with C3 indicates that, across %Jsing averages of C3 for 1965-1977 yielded very similar results, which are not reported here for reasons of space. 6As model B yields a slightly superior fit of the equation, as compared to model B’ [column (8) in table l] model B should be preferred to model B’.
M. Neumann et aZ., Concentration and foreign trade impact
14
Table 1
Determinants of rates of return on equity capital for 283 West German fums, 1965-1977, t-ratios in p2rentheses.’
Model B
Model A Explanatory variables
0)
0.007 (3.01)
C3
H H/C3a
(3) -
(4)
(5)
0.007 (2.45)
0.006 (2.19)
-0.11 (0.10)
-
-
-
-
(6)
0.04
(7)
-0.04 -0.04 (3.14) (3.31) -
0.003 (4.03)
-
H/C3 H 3’2/C32
(2)
Model B (8) -0.007 (1.35) -
-
-
-
ww 1.72 (3.41)
-
1.75 (3.46)
- 3.52 -3.51 (2.84) (2.84)
-
0.32 (3.13) -
VAR
0.22 (4.78)
0.22 (4.79)
0.21 (4.70)
0.22 (4.76)
0.22 (4.77)
0.19 (4.21)
0.19 (4.19)
0.19 (4.09)
cov
10.31 (2.25)
11.38 (2.49)
10.58 (2.33)
10.28 (2.24j
11.39 (2.49)
11.33 (2.54)
12.13 (2.74,\
11.47 ( 2 -!x;
VA
8.95 (0.85)
19.52 (2.21)
7.69 (0.73)
8.88 (0.84)
19.53 (2.20)
7.46 (0.72)
14.62 (1.69)
10.09 (0.97)
-0.04 (1.86)
-0.04 (1.81)
S
-0.04 (1.81)
-
-
- 0.03 (1.27)
- 0.03 (1.27)
LEV
1.17 (5.74)
1.15 (5.64)
1.15 (5.72)
1.17 (5.74)
1.15 (5.63)
1.16 (5.90)
1.15 (5.84)
1.18 (5.91)
PD
0.38 (2.96)
0.42 (3.23)
0.37 (2.86)
0.39 (2.95)
0.42 (3.21)
0.34 (2.65)
0.36 (2.82)
0.34 (2.67)
G
1.60 (2.62)
1.57 (2.60)
1.60 (2.62)
1.48 (2.42)
1.76 (2.96)
1.68 (2.85)
1.67 (2.79)
148
(2.43)
IM
- 1.30 - 1.26 - 1.42 (3.67) (4.16) (3.81)
- 1.29 - 1.26 (3.30) (3.37)
- 1.14 -1.11 (3.23) (3.17)
- 1.44 (4.25)
EX
- 1.52 - 1.25 -1.40 (5.23) (5.88) (4.54)
- 1.41 - 1.52 (4.83) (5.30)
- 1.06 - 1.12 (3.80) (4.08)
- 1.07 (3.79)
R2
0.40
0.39
0.41
0.40
0.39
0.44
0.44
0.42
N2)
(0.38)
(0.37)
(0.39)
(0.38)
(0.37)
(0.41)
(0.41)
(0.40)
“t-ratios exceedirrg 1.64 indicate significance at 0.05 level and better, according to a one-tailed test.
industries, a substantial degree of competition prevails. At least in some industries there seems more competition than suggested by the Cournot model. On the other hand, since the coefficient associated with C3 is an average of partial derivatives pertaining to firms, which are operating in various industries characterized by very different market structures, the negative sign does not preclude the existence of collusio;i.
M. Neumann et al., Concentration and foreign trade impact
15
Import penetration and exports. Price-cost margins are adversely affected by both exports and import penetration. The negative sign carried by the export ratio suggests that price-cost margins attainable on foreign markets are lower than those on domestic markets. That implies the presence of some collusion in the domestic economy. It does not, however, necessarily imply that the degree of competition in foreign countries is higher than it is at home. Comparatively low price-cost margins attainable by domestic firms on foreign markets may also be due to the fact that domestic firms are belonging to the competitive fringe in foreign countries and are thus acting as price takers. In any case, however, a negative sign carried by the export variable seems to indicate that domestic firms are not acting as price takers in domestic markets. It seems compelling to infer that some collusion must obtain in the domestic economy. Firm size. Space limitations do not allow discussion of the rest of the explanatory variables in detail, It will, however, be noticed that the absolute size of firms exerts an adverse influence on the rate of return.’ It suggests that unexploited economies of scale, and for that matter, collusive behaviour e+ts, Development of competition. The development of the relationship between market strulture and economic performance can be analyzed by looking at table 2. The most striking result is the close relationship between the impact of concentration on rates of return and the business cycle as depicted by the rates of growth of real gross domestic product (GDP) and the domestic level of prices [seealso Neumann, Babel and Haid (1983)]. The correlation between the rates of growth of GDP and the t-ratios of C3, upon using the regression model listed in column (1) of table 1, turned out to be 0.75 (t =2.60) and the correlation coefficient between rates of growth of GDP and the t-ratio of the Herfindahl index was 0.79 (t = 2.75). Furthermore collusion, as indicated by the coefficient associated with C3 of model B, seems to be more likely during business cycle upswings than during recessions. The correlation coefficient between the t-ratio associated with C3 of mod-lG B [column labelled (6) in table 21 and the rate of change of the domestic price level is 0.35 (t = 1.20). Although the suggested relationship is not strong it cannot be dismissed right away. It will also be noticed that the adverse impact of import penetration on rates of return is comparatively weak during business cycle upswings, in particular in 1965, 1968, 1973, 1974 ano 1977. The relationship between the ‘A very similar result has been obtained [Neumann and Haid (1984)].
for a cross-section
of West German industries
M. Neumann et al., Concentration and foreign trade impact
16
Table 2 Impact of concentration and foreign trade on rates of return. C3 of model R t-ratios Year
C3 (ly
H (4)
1965 1966 1967 1968 1969 1970 1971 1972 1973 1974 1975 1976 1977
2.93 2.70 1.80 3.93 4.54 3.10 3.56 2.59 2.81 1.71 1.56 4.90 0.84
3.34 2.90 2.03 4.67 4.84 3.58 3.90 3.03 3.23 2.25 1.67 5.31 1.38
IM
(1) - 1.51 -2.22 -394 -3.17 - 3.93 - 2.42 - 3.68 -4.09 - 1.70 -0.46 -3.21 -3.51 - 1.10
Coefficient (6l’ - 0.0552 - 0.0573 - 0.0479 -0.0540 - 0.0330 -MI495 -0.0261 -0.0338 -0.0226 - 0.0489 -0.0175 - 0.0084 -0.0469
t-ratio - 3.06 - 3.30 - 267 - 3.72 - 2.42 -299 - 1.40 - 1.82 - 1.23 - 2.23 -0.87 - 0.45 - 1.59
GDP growth rateb 5.5 2.6 -0.1 5.9 7.5 5.1 3.1 4.2 4.6 0.5 - 1.7 5.5 3.1
Pates of changes of prices’ c3 IM in percent 2.5 1.6 -0.9 -0.7 1.8 5.0 4.3 2.7 6.6 13.4 4.6 3.8 27
23.4 23.4 24.8 25.1 25.1 24.8 24.8 24.4 24.3 25.2 23.8 23.8 23.8
13 13 13 14 15 15 16 17 17 17 18 22 21
EX 16 17 18 20 19 19 19 20 22 25 24 29 27
‘Column number of model shown in table 1. bGDP in 1976 prices. cProducers’ prices on domestic markets.
impact of import penetration and the business cycle is brought out more clearly by looking at the development of domestic prices. The correlation between the t-ratios of import penetration and the rate of change of domestic prices is 0.60 (t=2.08). Hence, the larger the increase in prices, which indicates demand pressures in the domestic economy, the lower (the absolute value of) the t-ratio associated with import penetration. The picture is, however, somewhat marred since, first, the import ratio reflects not only imports competing with domestic supplies but also noncompeting imports of energy and raw materials. It, secondly, reflects intrafirm trade of transnational companies. It should, thirdly, be recalled that applying the regression model to single years entails bringing in a lot of noise which engenders a decline of the explanatory power of the model. The evidence adduced so far supports the infererice drawn above that a considerable degree of collusiveness is present. One therefore wonders whether it has changed during the time-span under consideration, especially in view of the substantial increase of foreign trade. As can be seen from table 2 domestic concentration did not change very much. The ratios of exports and import 10l:netration, however, increased by more than 51)percent. If the degree cf collusiveness obtaining in domestic markets would have ,,htiilged one should firstly expect a relative decline in the significance of the
M Neumann et al., Concentration and foreign trade impact
17
explanatory power of the concentration ratio as compared to that of the Herfindahl index. To examine whether such a trend existed we looked at the ratio of the t-values associated with C3 and the Herfindahl in&x, respectively. Using the Kendall rank correlation method the hypothesis of a decline of the ratio t &ttr must be rejected. The ordinary correlation coefficient of that ratio with time turned out to be -0.44, which at first sight seems to support the hypothesis of a downward trend. Actually, however, the significance of this correlation depends solely on the development in a single year, i.e., in 1977. If that year is deleted the correlation coefficient drops to insignificance. In a second line of inquiry one should look at the development of the ,tratios of C3 of model B [column labelled (6) in table 23. In this case one observes a strong time trend of increasing t-ratios (r = 0.8 1, t = 1.81, and Kendall’s S = 50, t= 3.05) and increasing coefficients associated with C3 (r = 0.63, t = 2.17, with time). That clearly suggests an increasing degree of collusiveness. Therefore it seems fair to conclude that in spite of the substantial mcrease of import penetration and a pronounced increase in the share oi domestic production being exported there is no visible decrease in the degree of collusiveness, and for that matter of the intensity of competition in domestic markets. That does not, of course, exclude the feeling of a dramatic increase of being exposed to foreign competition prevailing among domestic manufacturers. Since the share of domestic production being sold abroad with lower priciecost margins than those prevailing at home has increased the concern of domestic manufacturers about the exposure to more competition is completely valid. Nevertheless, at the same time, domestic markets can remain as collusive as before. 8,
c0nc1uslon
In a competitive fringe model the impact of the concentration ratios on priwost margins can be shown to be ambiguous. It may be either positive or negative depending on what kind of model is employed and on whether the collusive elements outweigh the competitive ones nor not. Therefore looking at the relationship between rates of return and the concentratior, ratio is not sufficient for perceiving collusive elements. We have set up a model that permits discerning collusion more clearly. Additional evidence concerning collusion can be gained by analyzing the influence of exports and imports on price-cost margins, Applying this approach to West German evidence yields the interesting result that the collusiveness prevailing in domestic markets has not visibly decreased over the time period under consideration in spite of a substantial increase in foreign trade.
18
M. Neumann et al., Concentration and foreign trade impact
Appendix A.!. Definition oj*variables =rate of return on equity: (total revenue minus current expenses for labour and raw material) over equity capital including reserves, = concentration: three-firm concentration ratio, c3 = Hefindahl index, H =export ratio: exports of commodities of the industry divided by EX total sales of the industry, =import penetration: imports of commodities of the industry bM diyided by total sales of the industry minus exports plus imports, =value added divided by total assets, VA PD =product differentiation: number of trade marks registered by the firm with the German Patent Office over equity, VAR,COV= trend free variance and covariance (with profits of all West German firms), respectively, of profits, S =size of the firm: natural logarithm of total assets, G =rate of growth of sales of the firm: slope of exponential time trend of sales of the firm for 1965-1977 (table 1), or for three consecutive years (table 2), respectively, LEV =leverage: debt over equity plus debt.
RRE
A.2. Data sources
Bayerische Hypotheken- und Wechselbank, ed., 1966-1979, Wegweiser durch deutsche Aktiengesellschaften, various volumes 1965-1978 (Mfinchen). I Bundeskartellamt, 1974, Bericht des Bundeskartellamtes fiber seine Tgtigkeit im Jahre 1973 sowie iiber die Lage und Entwicklung auf seinem Arbeitsgebiet ($50 GWB), 16undestagsdrucksache 7/2250 (Bonn). Monopolkommission, 1978, Fortschreitende Konzen:ration bei GroDuntemehmen, Hauptgutachten 1976/77 (Nomos, Baden-Baden). Monopolkommission, 1982, Fortschritte bei der Konzentrationserfassung, Hauptgutacb ten 1980/8”1(Nomos, Saden-Baden). Verlag Hoypenstedt, ed., 1976, 1977, BGrsenfiihrer - Stock guide (Darmstadt).
Clarke, R. arid S.W. Davies, 1982, Market structur;: and ptice-costmargins, Economica 49, 277287. Cowling ‘K. and M. Waterson, 1976, Price-cost margins and market structure, Economica 43, 267-274.
M. Neumann et al., Concentration and foreign trade impact
19
Encaoua, D. and A. Jacquemin, 1980, Degree of monopoly, indiczs of concentrr;tion and threat to entry, International Economic Review 21,87-105. Esposito, L. and F.F. Esposito, 1911, Foreign competition and domestic Lndustry profitaMity, Review of Economics and Statistics 53,343-353. Fischer, F.M., 1984, The misuse of accounting rates of rettim Reply, American Economic Review 74,509-517. Fiscner, F.M. and J.J. McGowan, 1983, On the misuse of accounting rates of return to infer monopoly profits, American Economic Review 73,82-97. Jacquemin, A., E. de GhelIinck and C. Huveneers, 1980, Concentration and profitability in a small open economy, Journal of Industrial Economics 29,131-144. Martin, S., 1983, Market, firm and economic performance, The Monograph Series in Finance and Economics of the Schools of Business of New York University, no. 1 (New York). Neumann, M., 1983, Concentration and economic performance, Mimeo. Neumann, M. and A. Haid, 1984, Concentration and economic performance, a cross-section analysis of West German industries, paper prepared for the Workshop on Industrial Organization of the International Institute of Management (Berlin). Neumann, M., I. Btibel and A. Haid, 1979, Profitability, risk and market structure in West German inoustries, Journal of Industrial Economics 27,227-242. Neumann, M., I. Biibel and A. Haid, 1983, Business cycle and industrial market power: An empirical investigation for West German industries, 1965-1977, Journal of Industrial . Econtinncs 32, 187-196. Pugel, T.A., 1980, Foreign trade and US market performance, Journal of Industrial Economics 29, 119-129. Ravenscraft, D.J., 1983, Structure-profit relationships at the line of business and industry lever, Review of Economics and Statistics 65,22-3 1. Saving, T., 1970, Concentration ratios and the degree of monopoly, International Econotic Review 11,139-146. Shepherd, W.G., 1982, Causes of increased competition in the U.S. economy, 1939-1980, Review of Economics and Statistics 64,613-626. Yule, G.U. and M.G. Kendall, 1973, An introduction to the theory of statistics, 14th ed. (Griffin, London).
DOMESTIC CONCENTRATION, FOREIGN TR4DE ECONOMIC PERF6RMANCE
AJPiD
Manfred NEUMANN, Ingo BABEL and Alfred HAID* Universityof Erlangen-Nhberg,
O-8500 Nfirnberg I, FRG
Final version received September 1984 In a competitive fringe model the impact of concentration on price-cost margins can be shown to be ambiguous. Therefore, looking at the relationship “between rates of return and concentration ratios is not sufficient for perceiving collusive elements. In this paper a model is set up that permits discerning collusion more clearly. Additional evidence concerning collusion can be gained by analyzing the influence of exports and imports on pric+cost margins. Applying this approach to West Germany yields the result that the collusiveness prevailing in domestic markets has not visibly decreased in spite of a substantial increase in foreign trade.
1. Introduction It seems to be widely believed that the more recent d.evelopment of foreign trade has contributed to intensifying competition in most industrialized countries. Two pieces of evidence are usually adduced for drawing this conclusion. First, in most structure-performance studies where foreign trade variables have been included the import ratio d:splayed an adverse effect on price-cost margins [see, for example, Esposito and Esposito (197 l), Neumann, Bobel and Haid (1979), Jacquemin, de Ghellinck and Huvencers (198C), Ravenscraft (1983), Martin (1983)]. Second, over the last decades the share of imports in most industrialized countries has substantially increased. From these observations it seems to follow that anticompetitive effects residing in domestic concentration have been counteracted by competition of foreign supply. Moreover, since domestic concentration has increased only at a glacial pace and has apparently been checked in many countries by anti-trust policies the conclusion seems to be warranted that the overall intensit:r of competition has increased [for the U.S., see Shepherd (1982)]. This inference, however, appears to rest on shaky ground. Looking more closely at the evidence will reveal that the relationships involved are a good deal more *We are grateful to Bernard Gahlen, Stephen Martin, Dennis C. Mueller, Roger Sugden,
and to the anonymous referees of this Journal for helpftil comments and suggestions. We- are also indebted to participants in workshops of the International Institute of Management, Berlin, and the OECD. This usual disclaimer applies. 0167-7187/85/$3.30 0 1985, Elsevier Science Pd~lishcrs
R.V. (North-Hollamc’)
M. Neu-
2
et al., Concentrationand foreigts m& impact
complex. To make the claim of an increased intensity of competition convincing other pieces of evidence must fit into the picture consistently. It will be shown in this paper that such is not the case. The additional evidence we wish to call upon is concerned with the explanatory power of alternative measures of concentration and the interpretation of the statistical impact of foreign trade variables on primost margins. It is well known that in a price leadership or competitive fringe model the g firm group are favourably affected by the price-cost margins of the lea leading firms, i.e., the concentration ratio, cumulative market share of provided the leading firm group acts as a. cartel [Saving (1970)]. It has furthermore been shown that in? a Coumot model it is the Herfindahl index which, instead of the concentrai ion ratio, has a favourable impact on pricecost margins [Cowling and W erson ( 1976)]. Looking at these polar cases suggests that the predictive ower of the concentration ratio and the IIerfindahl index should be ;:smpared. It could further be investigated whether collusitin or competitive behaviour prevails to a larger extent. In the next step it should be examined whether the prevalence of either mode of behaviour has changed more recently. The remainder of the paper is organized as follows. In section 2 the basic model is set up along the lines suggested by Encaoua and Jacquemin (1980), Clarke and Davies (! 982) and Neumann (1983). The implic&ians regarding concentration, imports and exports, and demand shifts are discussed in sections 3 through 5. In the subsequent section 6 empirical estimates are given for West German industries. They may, first, serve as an example for the testing procedure outlined above. ‘I&y are, however, also tieresting on their own because they throw some light on the recent development of conipctition in Germany.
2 The basic model
We shall start from the presu=mption of a market structure characterized by a group of dominant firms and a competitive fringe where both kinds of firms are assumed to behave differently. This point of vantage has been chosen because the dominant firm model constitutes a general case from which cases of homogeneous firms with the same type of behaviour can be derived as special cases. Let total demand Q =zf= 1 qi +x be served by k large firms and a cokmpetitive fringe. Firms belonging to the fringe act as price takers. Profits of’any one of the leading ohms are given by ni =P(Q)qi - C(qi), where AQ) is the inverted demand function. Profit maximization requires P+
4i
dPld4i= ci
9
M. Neumann et al., Concentration and foreign %de tmpzt
where ci : =
denotes marginal
c’(qi)
dPldqi
(
=fdPldQ)1+ i
costs,and
dqj ld4i
+ (dxldPNdPldqi)
j=l
) l
(2)
/
Subsequently, the following definitions will be used: E:= -
tl:= ~PIMWP~~ (PIQMQI~P~
Si : = qi /Q,
Zij : = (4i /4iKdqj
ai t =qi/(Q-x),
/dqih
E and q being price elasticities of total demand and supply of the competitive fringe, respectively. ai and si are market shares. The supply elasticity q of the competitive fringe will be assumed to be positive. Aggregate marginal costs of the competitive fringe are thus increasing. The coefficient Zii indicates the elasticity of reaction of rival j as expected by firm i. It is constrained by - 00 SZ& 1. In the Coumot case Zij= 0 for all i, j. On the other hand, perfect collusion implies Zi 1. Perfect competition in a particular market would imply that x =O and z f i dqj /dqi = - 1. That entails & +i ajZij= - ai. Any decrease of supply of one of the firms would be. immediately supplanted by augmented supplies of the rivals. For simplicity it is henceforth assumed that zij =zi for all j. Hence firm i expects all its rivals, irrespective of their individual size, to react in the same way upon an increase in the quantity supplied by firm i. On the other hand, it appears reasonable to assume that the zi’s differ according to the relative size of firm i. A large firm can expect a one percentage increase in its supply to be perceived by the rivals and will therefore elicit some reaction. A smaller firm can expect that a one percentage increase in its quantity supplied: goes largely or even completely undetected and will therefore evoke a comparatively smaller, or even no reaction at all, of the rivals. Thus it will be assumed that Zi=z,(uJ with &(a,) > 0. The elasticity of conjectural reaction held by firm i is thus an increasing function of its market share.’ To take the most simple form of the suggested relationship let j
Zi(Ui)=f+bQi,
=
00,
(3)
where { reflects institutional arrangements conducive to collusive behaviour, such as a legislation permitting cartels, price information agencies, etc. Eq. (2) can be re-arranged to yield dPldqi =
-WWQ~U
+U - ai)/aJzJlf 1+ (~~IE)(xIQ))*
‘We a--e indebted to Stephen Martin for suggesting this relationship.
M. Neumann et al., Concentration and foreign trade impact
4
Upon insertion into eq. (lj one obtains the individual price-cost margin mi:=(p-Ci)/p=K[U,
(4)
+Zi(l-ai)]/(E+dl-K)),
where K: 7 (Q - x)/Q is the k-firm concentration can also be written as
ratio. Since Qi= Si/K eq. (4)
(9
mi=(Si(l-zi)+ZiK)/(E+tf(l--K)).
From the individual price-cost margin mi one can proceed to the average price-cost margin of the k large firms. 2.1. Industry price-cost margin By definition, given in (4), p(1 -mi) = c, . Multiplying both sides with ai and summing over k ~~ITIISyields p(1 -xi mt ai) = & ci ai : = c where c is a weighted mean of the marginal costs of the Jcleading firms. Upon insertion into eq. (4) one obtains m:=(p-c)lp=Cmiai=K
H,+ i
i=1
i
Ziai-
-m II@+tt(l
k zi$
i =1 =
,
where H4: -xf= I a: is the Herfindahl Substituting (3) yields m=(K/(E+q(l
In the
~phtl
-K)))
(l-[+a)H,+[-ai
index of the large firm group.
i a:
i=l
1 .
(6)
case of zi = (5= constant for all i eq. (6) boils down to
m=K[H,(l-Q+fl/(e+q(l-K)).
I
(6)
The special cases derived in the prior literature can be easily recovered from (6’). Setting 5 =0 gives tte formula derived by Encaoua and Jacquemin (1980, p. 99). Setting [ = f yields the formula of Saving (1970). The formula for the pure Coumot case where all firms are on an equal footing which has been propounded by Cowling and Waterson (1976) can be found by setting K = 1 and 5 ~0. The slightly more general case invoked by Clarke and Davies (1982) emerges by setting K = 1 and [ > 0. In the general case, given by (6), the additional term ---oci a; is invoked. It might appear that in this way a dependence of the industry price-cost margin on individual market shares could be established. It does not, however, appear reasonable to assume a dependence of m on any one of
M. Neumann et al., Concentration a.& foreign trade impact
5
the individual market shares because an increase in any Qi implies a corresponding change in at least one other Qj. Yet individual market shares can nevertheless be shown to exert an influence on the industry price-cost margin. To capture this influence it is necessary to introduce an additional summary measure of the size distribution of the large firm group besides the Herfindahl index. We wish to propose a measure of asymmetry defined by
which is a pure number. Generally azH,. The coefficient equals unity in the case of monopoly. It coincides with the Herfindahl index in this case and in the case of equally sized firm~.~ If &a: = (aHjf)li2 is inserted into (6) one obtains m =(K/(E + ~(l- K)))[( 1- 5 + o)H, + [ - aa1/2H;3/‘],
(7a )
According to this equation m depends on the concentration ratio K :snd the Herfindahl index Ha of the leading firm group, given the coeficients QC, O,C,q and c. Since ai = si/K the Herfindahl index H, can be alternatively written as H,=H/K2 where H:=& s2i is a truncated Herfindahl index confined to the set of k large firms. It might be noted that the Herflndahl indices ordinari1.y employed in empirical work are also truncated indices because data on the stratum of smaller firms are usually not available. Substituting H/K2 for H, in eq. (7a) yields
(7b) According to this equation the average price--cost margin of the leading firms depends, inter alia, on K, H/K and H312/K2. 2The coefiicient a is devised in analogy to the coefficients proposed by K. Pearson [see Yule and Kendall (1973, p. lS9)] which utilize the second and third moments of a statistical distribution. Some examples may be useful to demonstrate the properties of the a-measure of asymmetry. Example no.
1
2
3
4
a,
0.80 0.15 0.05 0.665 0.516 0.904
0.70 0.20 0.10 0.540 0.352 0.787
0.50 0.30 0.20 0.380 0.160 0.467
040 0.40 0.20 0.360 0.136 0.396
a2 a3
RI c I.a31 a
it can easily be verified that a? H, and ct= H, in the case of equal market shares.
6
M. Neumann et al., Concentration d_foreign
trade impact
Interpreting these relationships requires a good deal of care since it is not evident a priori whether K, H,., and H are endogenous or exogenous variables. Along the lines suggested by Clarke and Davies (1382) the size distribution within the leading firm group, as reflected by K,, can be shown to be endogenous. In the case Zi = 5 = constant one obtains3
H, = l/K + (1- k((E- O/(1- 0)2(v,21k), where E = [s -+q( 1- K))/K and v, denotes the coefficient of variation of c,/ k c i = 1 ci reflecting the dispersion of the level of marginal costs. The size distribution ot the large firms, as summarized by Ha,thus depends on the dispersion of marginal costs, on various elasticities, on the number of leading firms, and on the k-fm concentration ratio. An analogous relationship can be presumed to exist in the more general case of Zi depending on market shares. Obviously endogeneity also obtains for the coefI%ient of asymmetry, 01. A similar reasoning applies to the number of dominant furns and their individual and cumulative market shares. The existence of a set of dominant firms which exercise a conduct different from the one found within the competitive fringe reflects barriers to entry to the privileged circle of leading firms. These barriers may be due to vari~_~s reasons, one of them being exclusive knowledge of, or access to, superior techniques which gives rise to differing cost functions. Explaining price-cost margins therefore would ideally require to invoke fundamental variables such as cost functions, conditions of entry, factors conducive to collusive behaviour, and demand functions. Those variables however, are very hard to come by and sometimes they are simp:v nonobservable. Therefore, analyses in the wake of the structure-performance paradigm stick to investigating the first order maximum conditions, as given in our case by eqs. (7a) and (7b). These equations should be interpreted as describing relationships between endogenous variables which are ultimately determined by unobservable factors outlined above. Note that also the demand elasticity s and the supply elasticity q are endogenous unless some very specific functional forms of market demand functions and marginal cost functions of the competitive fringe obtain. 3Let E = [E+ q( 1 - K)]/K and ii = [ = constant. In this cas.: eq. (4) can be
(p- Ci)/p= (ai+ c(1-
Ui))/E,
written
which yields
Q:= -[,/(l-i)+(E/(i-C))(l-Cilp). Summing over k and re-arranging gwes P=E~Ci[k(E-~-(1-5)]-‘. Substituting this into (i), which is then squared and summed yields H,= l/k+(l -Ic((E--c)/(l -L#($/k).
(9 (ii)
M. Nwnann
et al., Concentration umi~h-eign W&
impacr
7
Relationships like eqs. (7a) and (7’0) can be considered as lJypcthss ;.s suggesting that maximizing behaviour obtains, that a dominant firm grcrup facing a competitive fringe persists (due to differential entry opportunitizsj and, finally, that collusion_ is viable. Cocfronting these hypotheses with empirical evidence wil; show which one of them can withstand empirical testing. The approach can be further clarified by looting at conjectural variations in particular. Recall that the Coumot solution, perfect collusion, and pure competition, respectively, emerge as special cases, as shown above. It cannot, however, be expected that these special cases will be observed in practice. If the outcome during some period of time is observed some mixt-ure of com*‘ _ptiti*. - ktia;~tii atid collusion can *beexpected to obtain. The explanation of such an outcome notwithstanding, a descl-iption of the outcome can rn;,t conveniently be achieved by invoking the notion of conjectural variations. In this spirit eqs. (‘a) and (7b) are set up by regrouping the variables in such a way that the impact of collusion can be isolated, as shown in detail subsequently. 2.2. In&dual
price-cost margins
One may finally return to the individual price-cost margins. From p( 1- m) = c and p( 1 - mJ = ci one obtains
Hence, given ci/c, the individual price-cost margin is linearly dependent on the industry price-cost margin, and for that matter, on the variables determining m. C&n m, the individual price-cost margin is the higher the lower c’i/G 3. Conioentration and industry prkwmst
margins
Regarding the relationship between concentration and industry price-cost margins we suggest to consider two kinds of models, labelled model A and model B, which are represented by eqs. (7a) and (7b), respectively. Subcases A’ and B’ arise if zi = [ = constant, and Q= 0. In the case of model A one obtains am/aK=(m/K)i1+rlK/(s+11(1-Kj)], dm/aH, = (K/(e i- q( I- K)))[( 1 - [ + a) - (3/2)aa1i2Hti2].
(9) (10)
Hence, given H,, dm/dK is unambiguously non-negative, its site dcpendicg, among other things, on m which, in turn, depends, inter alia, on (. As can be
M. Neumann et al., Concentration
8
and foreign trade impact
seen from
the larger (5 the higher m. I-Ience a change of the institutional framework in favour of collusive behaviour gives rise to a higher price-cost margin. Whereas the impact of K on m isunambiguous, given Ii,, the influence exerted by H,, given K, can go either way, provided the asymmetry between the leading firms is sufficiently large. In the subcase A’ one could expect dmjaH, > 0. In the case of model B one obtains
amid
= ( i - 5 + ~T)/(E + q(i - K)) > 0,
am/a(H312/K2) = - OCC~/(C + tj( 1 -K)) < 0.
(12) (13)
In the model where all three variables are contained simultaneously am/aK is certainly positive if [ 20, i.e., in the Cournot case and if collusion obtains. It may become negative if CCO. Thus a negative am/X definitely implies competitive behaviour. The same reasoning applies in the subcase B’ where a==. If, a,,ernatively, the presumed market structure, characterized by a set of dominant firms facing a competitive fringe, does not exist, i.e., if K = 1, eqs. (7a) and (10) imply that h/aH, =am/aH may still be positive or negative unless 0 = 0. Provided 0 =0 ii is positive. 4. Imports and exports Generally, imports can be identified as part of the competitive fringe. That may not be true in some cases but it can safely be assumed to hold on average. If that is true an increase of the import ratio exerts an adverse influence on price-cost margins of dominant firms. The increase of the import ratio may be due to trade liberalization which gives rise to an outward shift of the supply curve of the competitive fringe, It may also be due to innovations adopted by foreign competitors which lower their marginal costs. In any case, within the framework of a homogeneous market, prices of domestic firms and foreign firms must be equal after an equilibrium has been attained. Turning to exports we note that total sales Si=pqi fp*q” of firm i are composed of sales to domestic and foreign markets, respectively, starred variables denoting foreign prices and quantities sold to foreign markets. Let
M. Neumann et al., Concentration and foreign trade impact
9
A:= JJ*~:/& denote the share of sales going to foreign markets and mi : = (pqi - cqi)/pqi and rnr : = (p*@ -cqf)/p*gf price-cost margins of firm i obtaining for sales to domestic and foreign markets, respectively. The firm’s total price-cost margin is then given by riTti : = [(pqi - cqi) + Cp*& - cqr)]/Si [see also Pugel(1980)] which can also be written as @=mi(l-A)+Amf. Hence an increase in the export ratio iz yields &iiilait= -(??li -mF)sO
as
Wli$??$.
If domestic markets, in consequence of trading costs and barriers to entry for foreign competitors, are relatively protected so that collusion is more feasible at home than abroad sales to foreign markets are exposed to more competition and hence rnf cm,. That implies a negative impact of an increased export ratio on a domestic firm’s price-cost margin fii* If, on the other hand, a firm is exposed to substantial competition at home and is selling to foreign markets where a higher degree of collusion prevails the difference mi - rnf may be negligible or even negative. That implies an insignificant or even positive impact of an increasing export ratio on the firm’s price-cost margin. 5. Demand shifts Shifts of demand curves which can be expected to occur during the business cycle will ordinarily entail changes of the price elasticity of demand, e. For example, given a linear demand schedule, a parallel shift of the demand curve to the right yields a decrease of e at a given level of price. More generally, any shift of a linear demand curve to the ‘right, except an anti-clockwise rotation of the demand curve around the choke-off price? entails a decrease of E. The impact of E on mi is given by 8mi /as =(cJc)am/& Using alternatively eq. (7a) or (7b) yields ampi5= -m/(6 + rj(1 - K)).
Hence during a business cycle upswing, which entails a decrease of E, mi should be expected to rise. As can be verified frown
a2m/ad4(= -(l/k + ul(1- K)))Cam/aK + mq/(c+ q(1- K))], thz rate of increase of the price-cost margin during a business cycle upswing
M. Neumanroet al., Conecc;;l’i&anand foreign trade impact
10
will be the more pronounced the higher the concentration ratio if the term in square brackets is positive. Utilizing model A, one finds
a2ttqaeaM,= --(am/aw/~~+vl(l -WJ. Hence, an analogous relationship holds for the %cr 4s~riYr ! i,:dex, provided am/aH, > 0. The increase of m upon an expansionary demand shift which lowers demand elasticity gives rise to higher prices if marginal costs are nondecreasing. Given non-decreasing marginal costs, prices will rise the more the higher concentration. Conversely, recessionary demand shifts which engender an increasing demand elasticity entail lower prices, the decrease in price being the larger the higher concentration. 6. Emphical analysis Since data on marginal costs are not available the model, as given ,y eqs. (7a), (7b) and (8), cannot immediately be applied. First cf a!!, ihe mo;,iel has to be modified such that empirically observable variables can be used. In the first step, marginal costs will be replaced by average labour costs. That yields a price-cost margin of firm i, FCMi
:=
vaiue added minus payroll of firm i value added of firm i
l
To expound the relationship between PCMi and the price-cost margin of the theoretical model, ml, we Erst note that the ratio of average costs to marginal COStS, ,Vi, indicates increasing, constant, and decrer?-.g returns to scale, respectively, as vi $1. That implies
where pqi denotes value added. Average costs are given by .
with w = wage rate, f.i = labour input, Ki =capital stock, and ;ip, =(r+ 6) user cost of capital, depending on the interest rate r, the rate of PC/k= depreciation 6, the value of capital goods, pK9and its rate of change PK. After some re-arranging one obtains
M. Neumann et al., Concentration and foreign trade impact
11
Hence PCMi depends on economies of scale, on the capital-output ratio and on those variables determining fii. We shall proceed by relating tii to equity capital because the owners of a firm can be expected to be concerned primarily with the rate of return on equity.4 Denote equity of firm i by Ei, and note that total assets of firm i, PKKi =Ei + Di , equal equity plus debt. Multiplying (16) by pqi/Ei yields the rate of return on equi,ty of firm i, RREi :=(pqi --WLi)/Ei =(l/(l
-LEF))[A+(l
-vi +vitii)(pqi/p~Ki)],
where LE 6 = Di /( Ei + Di). The rate of return thus depends, first, on.. those variables discussed in connection with fii, and for that matter m, second, on leverage LEF, on the ratio of value added to total assets, and on economies of scale. Equity must not be evaluated at current market prices of shares but at historical book value. If equity would be evaluated at current market prices of shares any influence attributable to market structure variables would be absorbed by the concomitant rise of the denominator in (17). To account for the influence of economies of scale we used two variables, i.e., the size of the firm (measured by natural logarithms of total assets) and the rate of growth of sales. Ordinarily one would expect size and economies of scale to be positively associated. If that is true size should exert a negative influence on R REi , as given by ( 17). Let Si denote size and assume V; (Si) > 0. That implies dRREJd& = -(l/(l
- LEK))(PqilPKKijr(l -fiib:tsiJ
Contrary to what might be expected, an inverse relationship between RRE, and size definitely points to the existence of economies of scale, albeit those economies which are not, or cannot be, fully exploited at the actual volume of sales. If economies of scale are associated with size the larger output the greater the ratio of average to marginal costs, given the volume of output. On the other hand, fast growth of output gives rise to higher rates of return because existing economies of scale can be more completely exploited. 41n view of the recent debate concerning accounting rates of return [Fischer and McGowan (1983) Fischer (1984)] we would like to submit the following observation. Whatever the relationship between the accounting rate of return and the economic rate of return of a particuIar firm at a particular point of time may be, any statistically significant association between accounting rates of return and market structure variables established for a cross-section of firms suggests that an economically meaningful relationship exists. This inference is of course subject to subsequent disproval. On the other hand it would not appear legitimate to draw inferences regarding the existence of monopoly power horn the observation of accounting rates of return exceeding some average level at a particular point in time in the case of an iki~viduul lirm.
12
M. Neumann et al., Concentrdion and foreign trade impact
’
The foregoing reasoning is predicated on the existence of monopoly power by which output is restricted. The relationship between unexploited economies of scale and monopoly power can be elaborated as follows. The present value of cash flow of a firm is given by Ai=J
0
CpF(Ki,Li)-wLi-(SP~-dg)Ki]e-rcdt,
where F(Ki , Ei) is the production function suitable for a particular SUE of the firm. Assuming a stationary equilibrium with constant prices of output and capital goods, one obtains
Given a homogeneous production function, UiF= F,,Ki + F,,Li is satisfied, where Fgf and FLi denote marginal productivities of capital and labour, respectively. Inserting the first order maximum conditions, p(l-fii)FLi=w
and
P(l-~,)F,,=(64t)pls-IjR,
yields
Since a financial equilibrium requires Ai =pK Ki it implies 1 -Vi + fii 02 = 0. If the production function is homothetic with variable economies of scale, which give rise to a u-shaped average cost curve, vi = l/( 1 -tiJ is an increasing function of C&. Increasing monopoly power yielc3 aapoint fWhrt+ up on the declining branch of the average cost curve w?.ere ui is larger. Hence, given a positive association with the size of the firm .:ind economies of scale entailed by the kind of technique suitable for a paricular size class, unexploited economies are increasing with monopoly power. To complete the model two further variables must be included. It will first be noted that the rate of return on equity also depends on risk which can be depicted, as suggested by the capital asset pricing model, by the variance and covariance of rates of return in a way discussed in Imore detail elsewhere INeumann, Babel and Haid (1979)~. Second, beyond the theoretical framework discussed so far ye have to account for the fact that in many industries heterogeneous products are being sold and that heterogeneity may give rise to barriers to entry. Moreover, heterogeneity ordinarily entails advertising expenses by which price-cost margins are augmented. For depicting heterogeneity we used the number of trademarks, as registered by a firm with the German Patent Ofice, deflated by equity, i.e., the clenominator of eq. (17). The various explanatory variables are hypothesizeI. to exert an influence
M. Neumann et al., Concentration andforeign trade impact
13
on the rate of return which must be depicted P11:3 non-linear equation, not particularly well suited for estimation. For tht; Furpose of a regression ar,.ysis, however, it can be approximated by a linear equation using a P .lor series, the coefficients of which denote the partial derivatives with spect to the variables. 7. The evidence The data base covering 196!5-1977 pertains to 283 West German joint stock companies for which data were available. Individual firm data were collected from annual publications by Hoppenstedt Verlag and from informationa which were made availab1.e to us by the German Patent Office. Industry data were drawn from various official publications. Definitions of the explanatory variables are summarized in the appendix. We report two kinds of results. The first regression, reported in table 1, applies to the average values of the various variables over the total timespan from 1965 through 1977. The second regression, reported in table 2, applies to individual years. Regarding concentration we did not use average values but data on 3-firm concentration ratios and the Herfimdahl’index of 1977, as published by the Monopolkommission, because the publication of the Herfindahl index started only with data for 1977. To ensure comparability between the regression results obtained for C3 and the Herfindahl, respectively, we also took the data on C3 only for 1977? ‘The results reported in table 1 confirm the validity of the theoretical approach. It will be noticed that v/A( =value added over assets) is only significant if the size variable S is deleted. The disappearance of significance for v/A and also the relatively low level of significance attained by the size variable can probably be attributed to multicollinearity since the simple coefficient of correlation between I+4 and S is - 0.61. Concentration. Using C3 and the Herfindahl index alternatively [column (I), (2) and (3) in table I] we find both to exert a significantly favourable influence on the rate of return, the Herflndahl yielding a slightly higher level of significance. Model A [columns (4) and (5) in table 1-j which uses C3 and H,, iapproximated by H/( C3)2, simultaneously yields a significant coeffkien t for C3 whereas Ha turns out to be insignificant, the latter result being in Iine with theoretical expectations. Model B [columns (6) and (7) in table l] fully confirms theoretical expectations. ’ The negative sign associated with C3 indicates that, across %Jsing averages of C3 for 1965-1977 yielded very similar results, which are not reported here for reasons of space. 6As model B yields a slightly superior fit of the equation, as compared to model B’ [column (8) in table l] model B should be preferred to model B’.
M. Neumann et aZ., Concentration and foreign trade impact
14
Table 1
Determinants of rates of return on equity capital for 283 West German fums, 1965-1977, t-ratios in p2rentheses.’
Model B
Model A Explanatory variables
0)
0.007 (3.01)
C3
H H/C3a
(3) -
(4)
(5)
0.007 (2.45)
0.006 (2.19)
-0.11 (0.10)
-
-
-
-
(6)
0.04
(7)
-0.04 -0.04 (3.14) (3.31) -
0.003 (4.03)
-
H/C3 H 3’2/C32
(2)
Model B (8) -0.007 (1.35) -
-
-
-
ww 1.72 (3.41)
-
1.75 (3.46)
- 3.52 -3.51 (2.84) (2.84)
-
0.32 (3.13) -
VAR
0.22 (4.78)
0.22 (4.79)
0.21 (4.70)
0.22 (4.76)
0.22 (4.77)
0.19 (4.21)
0.19 (4.19)
0.19 (4.09)
cov
10.31 (2.25)
11.38 (2.49)
10.58 (2.33)
10.28 (2.24j
11.39 (2.49)
11.33 (2.54)
12.13 (2.74,\
11.47 ( 2 -!x;
VA
8.95 (0.85)
19.52 (2.21)
7.69 (0.73)
8.88 (0.84)
19.53 (2.20)
7.46 (0.72)
14.62 (1.69)
10.09 (0.97)
-0.04 (1.86)
-0.04 (1.81)
S
-0.04 (1.81)
-
-
- 0.03 (1.27)
- 0.03 (1.27)
LEV
1.17 (5.74)
1.15 (5.64)
1.15 (5.72)
1.17 (5.74)
1.15 (5.63)
1.16 (5.90)
1.15 (5.84)
1.18 (5.91)
PD
0.38 (2.96)
0.42 (3.23)
0.37 (2.86)
0.39 (2.95)
0.42 (3.21)
0.34 (2.65)
0.36 (2.82)
0.34 (2.67)
G
1.60 (2.62)
1.57 (2.60)
1.60 (2.62)
1.48 (2.42)
1.76 (2.96)
1.68 (2.85)
1.67 (2.79)
148
(2.43)
IM
- 1.30 - 1.26 - 1.42 (3.67) (4.16) (3.81)
- 1.29 - 1.26 (3.30) (3.37)
- 1.14 -1.11 (3.23) (3.17)
- 1.44 (4.25)
EX
- 1.52 - 1.25 -1.40 (5.23) (5.88) (4.54)
- 1.41 - 1.52 (4.83) (5.30)
- 1.06 - 1.12 (3.80) (4.08)
- 1.07 (3.79)
R2
0.40
0.39
0.41
0.40
0.39
0.44
0.44
0.42
N2)
(0.38)
(0.37)
(0.39)
(0.38)
(0.37)
(0.41)
(0.41)
(0.40)
“t-ratios exceedirrg 1.64 indicate significance at 0.05 level and better, according to a one-tailed test.
industries, a substantial degree of competition prevails. At least in some industries there seems more competition than suggested by the Cournot model. On the other hand, since the coefficient associated with C3 is an average of partial derivatives pertaining to firms, which are operating in various industries characterized by very different market structures, the negative sign does not preclude the existence of collusio;i.
M. Neumann et al., Concentration and foreign trade impact
15
Import penetration and exports. Price-cost margins are adversely affected by both exports and import penetration. The negative sign carried by the export ratio suggests that price-cost margins attainable on foreign markets are lower than those on domestic markets. That implies the presence of some collusion in the domestic economy. It does not, however, necessarily imply that the degree of competition in foreign countries is higher than it is at home. Comparatively low price-cost margins attainable by domestic firms on foreign markets may also be due to the fact that domestic firms are belonging to the competitive fringe in foreign countries and are thus acting as price takers. In any case, however, a negative sign carried by the export variable seems to indicate that domestic firms are not acting as price takers in domestic markets. It seems compelling to infer that some collusion must obtain in the domestic economy. Firm size. Space limitations do not allow discussion of the rest of the explanatory variables in detail, It will, however, be noticed that the absolute size of firms exerts an adverse influence on the rate of return.’ It suggests that unexploited economies of scale, and for that matter, collusive behaviour e+ts, Development of competition. The development of the relationship between market strulture and economic performance can be analyzed by looking at table 2. The most striking result is the close relationship between the impact of concentration on rates of return and the business cycle as depicted by the rates of growth of real gross domestic product (GDP) and the domestic level of prices [seealso Neumann, Babel and Haid (1983)]. The correlation between the rates of growth of GDP and the t-ratios of C3, upon using the regression model listed in column (1) of table 1, turned out to be 0.75 (t =2.60) and the correlation coefficient between rates of growth of GDP and the t-ratio of the Herfindahl index was 0.79 (t = 2.75). Furthermore collusion, as indicated by the coefficient associated with C3 of model B, seems to be more likely during business cycle upswings than during recessions. The correlation coefficient between the t-ratio associated with C3 of mod-lG B [column labelled (6) in table 21 and the rate of change of the domestic price level is 0.35 (t = 1.20). Although the suggested relationship is not strong it cannot be dismissed right away. It will also be noticed that the adverse impact of import penetration on rates of return is comparatively weak during business cycle upswings, in particular in 1965, 1968, 1973, 1974 ano 1977. The relationship between the ‘A very similar result has been obtained [Neumann and Haid (1984)].
for a cross-section
of West German industries
M. Neumann et al., Concentration and foreign trade impact
16
Table 2 Impact of concentration and foreign trade on rates of return. C3 of model R t-ratios Year
C3 (ly
H (4)
1965 1966 1967 1968 1969 1970 1971 1972 1973 1974 1975 1976 1977
2.93 2.70 1.80 3.93 4.54 3.10 3.56 2.59 2.81 1.71 1.56 4.90 0.84
3.34 2.90 2.03 4.67 4.84 3.58 3.90 3.03 3.23 2.25 1.67 5.31 1.38
IM
(1) - 1.51 -2.22 -394 -3.17 - 3.93 - 2.42 - 3.68 -4.09 - 1.70 -0.46 -3.21 -3.51 - 1.10
Coefficient (6l’ - 0.0552 - 0.0573 - 0.0479 -0.0540 - 0.0330 -MI495 -0.0261 -0.0338 -0.0226 - 0.0489 -0.0175 - 0.0084 -0.0469
t-ratio - 3.06 - 3.30 - 267 - 3.72 - 2.42 -299 - 1.40 - 1.82 - 1.23 - 2.23 -0.87 - 0.45 - 1.59
GDP growth rateb 5.5 2.6 -0.1 5.9 7.5 5.1 3.1 4.2 4.6 0.5 - 1.7 5.5 3.1
Pates of changes of prices’ c3 IM in percent 2.5 1.6 -0.9 -0.7 1.8 5.0 4.3 2.7 6.6 13.4 4.6 3.8 27
23.4 23.4 24.8 25.1 25.1 24.8 24.8 24.4 24.3 25.2 23.8 23.8 23.8
13 13 13 14 15 15 16 17 17 17 18 22 21
EX 16 17 18 20 19 19 19 20 22 25 24 29 27
‘Column number of model shown in table 1. bGDP in 1976 prices. cProducers’ prices on domestic markets.
impact of import penetration and the business cycle is brought out more clearly by looking at the development of domestic prices. The correlation between the t-ratios of import penetration and the rate of change of domestic prices is 0.60 (t=2.08). Hence, the larger the increase in prices, which indicates demand pressures in the domestic economy, the lower (the absolute value of) the t-ratio associated with import penetration. The picture is, however, somewhat marred since, first, the import ratio reflects not only imports competing with domestic supplies but also noncompeting imports of energy and raw materials. It, secondly, reflects intrafirm trade of transnational companies. It should, thirdly, be recalled that applying the regression model to single years entails bringing in a lot of noise which engenders a decline of the explanatory power of the model. The evidence adduced so far supports the infererice drawn above that a considerable degree of collusiveness is present. One therefore wonders whether it has changed during the time-span under consideration, especially in view of the substantial increase of foreign trade. As can be seen from table 2 domestic concentration did not change very much. The ratios of exports and import 10l:netration, however, increased by more than 51)percent. If the degree cf collusiveness obtaining in domestic markets would have ,,htiilged one should firstly expect a relative decline in the significance of the
M Neumann et al., Concentration and foreign trade impact
17
explanatory power of the concentration ratio as compared to that of the Herfindahl index. To examine whether such a trend existed we looked at the ratio of the t-values associated with C3 and the Herfindahl in&x, respectively. Using the Kendall rank correlation method the hypothesis of a decline of the ratio t &ttr must be rejected. The ordinary correlation coefficient of that ratio with time turned out to be -0.44, which at first sight seems to support the hypothesis of a downward trend. Actually, however, the significance of this correlation depends solely on the development in a single year, i.e., in 1977. If that year is deleted the correlation coefficient drops to insignificance. In a second line of inquiry one should look at the development of the ,tratios of C3 of model B [column labelled (6) in table 23. In this case one observes a strong time trend of increasing t-ratios (r = 0.8 1, t = 1.81, and Kendall’s S = 50, t= 3.05) and increasing coefficients associated with C3 (r = 0.63, t = 2.17, with time). That clearly suggests an increasing degree of collusiveness. Therefore it seems fair to conclude that in spite of the substantial mcrease of import penetration and a pronounced increase in the share oi domestic production being exported there is no visible decrease in the degree of collusiveness, and for that matter of the intensity of competition in domestic markets. That does not, of course, exclude the feeling of a dramatic increase of being exposed to foreign competition prevailing among domestic manufacturers. Since the share of domestic production being sold abroad with lower priciecost margins than those prevailing at home has increased the concern of domestic manufacturers about the exposure to more competition is completely valid. Nevertheless, at the same time, domestic markets can remain as collusive as before. 8,
c0nc1uslon
In a competitive fringe model the impact of the concentration ratios on priwost margins can be shown to be ambiguous. It may be either positive or negative depending on what kind of model is employed and on whether the collusive elements outweigh the competitive ones nor not. Therefore looking at the relationship between rates of return and the concentratior, ratio is not sufficient for perceiving collusive elements. We have set up a model that permits discerning collusion more clearly. Additional evidence concerning collusion can be gained by analyzing the influence of exports and imports on price-cost margins, Applying this approach to West German evidence yields the interesting result that the collusiveness prevailing in domestic markets has not visibly decreased over the time period under consideration in spite of a substantial increase in foreign trade.
18
M. Neumann et al., Concentration and foreign trade impact
Appendix A.!. Definition oj*variables =rate of return on equity: (total revenue minus current expenses for labour and raw material) over equity capital including reserves, = concentration: three-firm concentration ratio, c3 = Hefindahl index, H =export ratio: exports of commodities of the industry divided by EX total sales of the industry, =import penetration: imports of commodities of the industry bM diyided by total sales of the industry minus exports plus imports, =value added divided by total assets, VA PD =product differentiation: number of trade marks registered by the firm with the German Patent Office over equity, VAR,COV= trend free variance and covariance (with profits of all West German firms), respectively, of profits, S =size of the firm: natural logarithm of total assets, G =rate of growth of sales of the firm: slope of exponential time trend of sales of the firm for 1965-1977 (table 1), or for three consecutive years (table 2), respectively, LEV =leverage: debt over equity plus debt.
RRE
A.2. Data sources
Bayerische Hypotheken- und Wechselbank, ed., 1966-1979, Wegweiser durch deutsche Aktiengesellschaften, various volumes 1965-1978 (Mfinchen). I Bundeskartellamt, 1974, Bericht des Bundeskartellamtes fiber seine Tgtigkeit im Jahre 1973 sowie iiber die Lage und Entwicklung auf seinem Arbeitsgebiet ($50 GWB), 16undestagsdrucksache 7/2250 (Bonn). Monopolkommission, 1978, Fortschreitende Konzen:ration bei GroDuntemehmen, Hauptgutachten 1976/77 (Nomos, Baden-Baden). Monopolkommission, 1982, Fortschritte bei der Konzentrationserfassung, Hauptgutacb ten 1980/8”1(Nomos, Saden-Baden). Verlag Hoypenstedt, ed., 1976, 1977, BGrsenfiihrer - Stock guide (Darmstadt).
Clarke, R. arid S.W. Davies, 1982, Market structur;: and ptice-costmargins, Economica 49, 277287. Cowling ‘K. and M. Waterson, 1976, Price-cost margins and market structure, Economica 43, 267-274.
M. Neumann et al., Concentration and foreign trade impact
19
Encaoua, D. and A. Jacquemin, 1980, Degree of monopoly, indiczs of concentrr;tion and threat to entry, International Economic Review 21,87-105. Esposito, L. and F.F. Esposito, 1911, Foreign competition and domestic Lndustry profitaMity, Review of Economics and Statistics 53,343-353. Fischer, F.M., 1984, The misuse of accounting rates of rettim Reply, American Economic Review 74,509-517. Fiscner, F.M. and J.J. McGowan, 1983, On the misuse of accounting rates of return to infer monopoly profits, American Economic Review 73,82-97. Jacquemin, A., E. de GhelIinck and C. Huveneers, 1980, Concentration and profitability in a small open economy, Journal of Industrial Economics 29,131-144. Martin, S., 1983, Market, firm and economic performance, The Monograph Series in Finance and Economics of the Schools of Business of New York University, no. 1 (New York). Neumann, M., 1983, Concentration and economic performance, Mimeo. Neumann, M. and A. Haid, 1984, Concentration and economic performance, a cross-section analysis of West German industries, paper prepared for the Workshop on Industrial Organization of the International Institute of Management (Berlin). Neumann, M., I. Btibel and A. Haid, 1979, Profitability, risk and market structure in West German inoustries, Journal of Industrial Economics 27,227-242. Neumann, M., I. Biibel and A. Haid, 1983, Business cycle and industrial market power: An empirical investigation for West German industries, 1965-1977, Journal of Industrial . Econtinncs 32, 187-196. Pugel, T.A., 1980, Foreign trade and US market performance, Journal of Industrial Economics 29, 119-129. Ravenscraft, D.J., 1983, Structure-profit relationships at the line of business and industry lever, Review of Economics and Statistics 65,22-3 1. Saving, T., 1970, Concentration ratios and the degree of monopoly, International Econotic Review 11,139-146. Shepherd, W.G., 1982, Causes of increased competition in the U.S. economy, 1939-1980, Review of Economics and Statistics 64,613-626. Yule, G.U. and M.G. Kendall, 1973, An introduction to the theory of statistics, 14th ed. (Griffin, London).