American Economic Association
The Effect of the EEC and EFTA on European Trade: A Temporal Cross-Section Analysis
Author(s): Norman D. Aitken
Source: The American Economic Review, Vol. 63, No. 5 (Dec., 1973), pp. 881-892
Published by: American Economic Association
Stable URL: http://www.jstor.org/stable/1813911 .
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The
Effect
of
the
European
EEC
and
A
Trade:
BETA
on
Temporal
Cross-Section Analysis
By
NORMAN
D.
Utilizing a cross-sectional trade flow
model of the type developed by Hans
Linnemann and Jan Tinbergen, this study
attempts to isolate empirically the major
forces which have shaped European trade
relations over the period 1951-67. We first
estimate via the use of dummy variables
the impact of the European Economic
Community (EEC) and the European
Free Trade Association (EFTA) on member trade. For each year of the European
integration period (1959-67), a crosssectional equation is estimated ancl used
to test for the existence and approximate
size of the respective integration effects.
The equation is also calculated for the
eight years prior to the integration period
to obtain a clear picture of the forces
which were at work before the formation
of the EEC. Secondly, a base year equation is used to make projection estimates
of the gross trade creation and European
trade diversion effects of the two communities.
AITKEN*
of a trading community brought about
through integration, regardless of whether
the additional trade replaces domestic
production or whether it replaces nonmember exports. The substitution of imports from member countries (higher cost
imports) for imports from nonmember
countries (lower cost imports) will constitute trade diversion (TD). External
trade creation (ETC) will refer to integration-caused increases in trade between a trading community and countries
outside the trading community. ETC as
an effect of economic integration is possible in the case of a customs union where
high tariff countries have to reduce their
tariffs to the outside world as a part of
the process of achieving a common external tariff. ETC minus TD yields the
net effect of a trading bloc on the outside
world.
All empirical studies which have attempted to measure integration effects
have been faced with the common problem
of isolating the effect of integration on
trade from the effect of income growth
and changes in other variables which
normally affect international trading patterns. The major approaches to this problem have either been to examine changes
in market share of imports (or apparent
consumption)' or to incorporate income
directly into the statistical analysis (by
I. Methodology
As defined by Bela Balassa (1967, p. 5),
gross trade creation (GTC) will refer to
the total increase in trade among members
* Associate professor of economics, University of
Massachusetts. Research for the paper was supported
in part by a grant from the Research Council of the
University of Massachusetts. I am indebted to Ronald
Ehrenberg, Bradley Gale, James Kindahl, Joseph
Kushner, Laurence Mauer, Anthony Scaperlanda,
George Treyz, Brendan Walsh, anonvmous referees, and
the editor for valuable comments on earlier drafts of
the paper. I assume all responsibility for any errors that
remaiin.
1 Various formulations of the market-share approach
are used in the studies by Edwin Truman, P. S. Verdoorn, and F. J. M. Meyer Zu Schlochtern, J. Waelbroeck, M. K. Carney, and The European Free Trade
Association (1969).
881
THE AMERICAN ECONOMIC REVIEW
882
DECEMBER 1973
calculating income elasticities of import
demand for the pre- and postintegration
periods' or by using income as an independent variable in a trade-estimating
the populations of the respective countries,
Dij is the distance between the commercial centers of the two countries, A ij is a
dummy variable for adjacent or neighbor-
model).' Both of these approaches attempt
ing countries, p EEC and
to measure the effects of integration indirectly, or in other words as a residual.
Either by using a base year trade matrix
or a base period relationship between income and imports, estimates are made of
what trade would have been in the absence
of .economic integration and these estimates are then compared to actual trade
to obtain the trade preference effect. One
is thus left with the uncertainty as to
whether or not other factors may have
been responsible for the difference between projected and actual values. Furthermore, the residual approach does not
provide an objective basis for determining
the first year in which integration effects
occurred. The base periods selected by
previous studies of the EEC range from
1958 to 1961 and hence it is clear that
there is no general agreement as to the
timing of the first EEC effect on trade.
The present study has attempted to deal
with these problems by estimating,
through the least squares regression
method, the following variant of the
Linnemann trade flow model for each year
of the 17-year period 1951-67.
variables for trade between partners of the
EEC and EFTA, respectively, and log
refers to common logarithms.4
The variables Yi and NA together determine the potential export supply of
country i, with Y, determining economic
capacity and Ni determining the domestic
market/foreign market production ratio.
(N determines market size and, assuming
economies of scale, the larger N the more
lines of production for which the country
will meet the minimum market size for
efficient market production (see Linnemann, pp. 11-14). The larger N, therefore, the larger the domestic market to
foreign market ratio and the smaller the
potential export supply of the country.)
The variables Yj and Nj together determine the potential import demand for
country j, using the same arguments that
were applied to potential export supply.
Dij is a proxy variable for natural trade
resistence which in turn is a composite of
transportation cost, transport time, and
economic horizon. Consequently, Dij along
with Ni and Nj is hypothesized to have a
negative effect on Xi.
(1)
log Xij = log bo+ bilog Dij +
b2 log
Yi
+ b3log Yj + b log Nl
+ b5log Nj + b6log A ij
EEC
+ b7log Pi
EFTA
+ b8log Pi,
+ log eii
where Xij is the dollar value of country
i's exports to country j measured according to country.j's import data, Yi and YK
are the nominal GAVPof countries i and j
expressed as a dollar value, Ni and Nj are
2 See Bela Balassa (1967).
3
See MordechaiKreininand J. Waelbroeck.
PiFTA
are dummy
Neighboring countries (A ij) can be ex-
pected to have an additional stimulus to
trade because of similarity of tastes and
an awareness of common interests (see
Balassa (1961, p. 40)). Perhaps more important, however, neighboring countries
are likely to experience significant additional amounts of international trade in
4Sources of data: Directionof Trade; DistanceBeFinancial
tweenPorts;EFTA Trade:1969;International
Statistics; National Accountsof OECD Counties;and
MonthlyBulletin of Statistics.Values of 2 and 1 were
used for the three dummy variables to indicate the
presence and absence of the given characteristic.A
mimeographeddata appendix is available from the
author.
VOL. 63 NO. 5
AITKEN: EUROPEAN
the form of what are essentially locally
traded goods, especially where border
regions are densely populated as in much
of 1EJurope.)
The use of this model permits us to incorporate into the analysis as independent variables the preference area effects
through the use of dummy variables.
Dummy variables are used to represent in
an approximate wav phenomena which
are difficult to measure. I'he approximate
nature of the trade preference variables
means that their measured effect on trade
flows must in turn be considered approximate. On the other hand, there are distinct
a(lvanitages to the use of the cross-sectional
model.
First, by estimating the preference effect
as an independent variable in a multiple
regression equa.tion one is able to hold
constant other major variables which
affect tracle, including not onily potential
(lemand and supply but also to some extent the effects of general changes in
trade liberalization and transportation
cost. Changes in the latter over time would
tend to be picke(d up by the proxy variables Dj, and A, . (The proxy variables
themselves, of course, do not change over
time, but changes in the real variables
Equation (1) represents what Tinbergen has labeled
a "turnover eqluation" in which prices are not specified
(see linnemann pp. 4, 47 for a detailed discussion of
this point). Variables which explicitly measure artificial
trade barriers are also absent from the equation because
of the overwhelming statistical and methodological
problems associated with their direct measurement
(Linnemann, p. 30). The exclusion of these variables
from the equation could be expected to seriously bias
the trade preference coefficients onlv if there were a high
correlation between the excluded variables and the
trade preference variables themselves. If a high correlation existed, however, it would show up in the form of
significant or near significant trade preference coefficients in the preintegration years. Since, as will be seen
below, the preintegration )reference coefficients are
consistently insignificant, it seems reasonable to assume
that the preference variables are not highly correlated
with any important variable that has been excluded
from the statistical analysis.
TRADE
883
would be expected to show up as changes
from year to year in the estimated coefficients of the proxy variables.) Specifically,
since the size of the trade-stimulating
effect of European trade liberalization may
be expected to vary inversely with distance (natural trade resistance) the effect
of trade liberalization may show up as a
change over time in the D?j or A ij coefficients or both.
In addition, the use of a cross-sectional
model allows the estimation of trade preference coefficients for each year in the
integration period and hence a series of
parameter estimates can be obtained
which can then be considered as a whole
in terms of whether t;heirpattern indicates
the expected cumulative growth in the
preference effects.
Ihe preference parameters can in turn
be used to estimate the dollar value of
GTC for each of the two European communities. Because the estimates for each
year are derived from the cross-sectional
equation for that year, each estimate is
independent of the others and the estimating procedure does not require the use of a
base year. In fact, the results of the study
provide information which may be useful
in determining when the first integration
effects on trade occurred.
Using the trade-preference coefficients
to estimate the trade-stimulating effect of
integration in any postintegrationi year
requires one to assume that the size of the
coefficient is being determined solely by
the effect of the trade preference and therefore that it is not in part reflecting some
other special trade relationship which had
existed in the preintegration period. Consequently, it will be necessary to test the
preference coefficients for nonsignificance
in the preintegration period as well as for
significance in the postintegration period.
By calculating the equation for the eight
years preceding the first tariff reductions
of the EEC, we were thus able to test for
884
THE AMERICAN ECONOMIC REVIEW
the existence of preiintegration preference
effects.
The sample from wlhich the equationis
were calculated inclu(les the original seven
EFTA members plus the five EEC trading
countries (Belgium-Luxembourg being oine
trading country).6 The sample thus contains 20 trade flows between EEC partners,
42 trade flows between EFTA partiners,
and 70 trade flows betweein the members
of the two trade blocs for a total of 132
observations per year. The 70 trade flows
between the two blocs are assumed to constitute "normal" European trade and it is
againist these normal trade flows that trade
among, members of the respective communities is tested for preferential effects.7
Clearly, this assumption would be invalid
for any year in which trade flows between
the two blocs have, on the average, experienced significant amounts of TD.8 Previous studies, however, have found little or
6 See Table 3 for a listing of the individual countries.
Finland was excluded from the sample because its late
entrv into EFTA meant that its inclusion in the sample
would have made it more difficult to discern the first
EFTA effect on the original members.
The decision to limit the sample only to members of
the two trading communities mav be open to question.
Ideally, one would prefer to have "normal tracle" represented by the exports of both EFT.4 and 1E'1C members
to countries who are not memb)ersof either bloc. WAithin
Western Europe, however, only less developed countries
fall into the nonmember category ancdthe possibility of
having normal trade represented by trade flows to the
developed countries outsidle of Europe was rejectecl hecause a signiificant proportion of the observations oil
normal trade woulcI have occupied extremely high
points on the regression lines with respect to the inclependent variables, while all intra--I'KCand intra-I?FTA
observations would have been in the lower range. It was
to the speciconcluded, therefore, that a sample limiaite(d
fied European countries would provide the greatest
opportunity for obtaining b)ias-free preference coeff;cients in the preintegration years andclhence that one
would ob)tain the most sensitive measure for identifying
both the timing andigeneral magnitu(le of the preference
effects. The fact that trade flows between the two blocs
had to b)eused to represeflt normal trade precluded the
introduction of additional dummv variables to measure
the potential trade diversion effect (TID) of each bloc
against the other.
8 The existence of TI) would leadi to an inflated estimate of (TC sinice the value of the trade preference
coefficients would tend to be increased by an integra--
DECEMBER
1973
no net trade diversion by either bl]oc. But
rather thani rely on the findings of previous
studies, it was decided to test for the
existence of TD betweeni the two blocs by
the following procedure: The cross-section
regression results were examine(l to find
an appropriate base year free of integration effects. The equation for the base
year was then used to estimate what trade
would have been between the two blocs in
subsequent years hadl there been no initegration. in Europe. A comparison of the
projection estimates with actual tradle provides an estimate of the degree to which
tra(le betweeni the two blocs has been
reduce(d as a result of trade diversion. The
projection approach was also used to estimate gross trade creation within the EEC
and EFTA as a further check on the
dummy variable estinmates. Consequently,
the second part of the statistical analysis
consists of residual estimates of the dollar
value of gross trade creation and trade
diversioni, but they are residual estimates
based on the information providled by the
regression results as to the timing of the
first EEC effect on EJuropean trade. Furthermore, the assumption of normal trade
between the EEC and E.FTA shouldI be
consideredI only as an initial working hypothesis to be tested against the projection
estimates before any conclusioins are
reache(d concerniing the general magnitude
of GTC withini the two tra(ling communities.
II. Empirical Results
A. Regression Results
Table 1 contains the estimated parameter values for the trade flow equation for
the 17-year period 1951-67. The trade
tion-cause(l reduction in the average trade flow between
the two blocs as well as by an integration-caused increase inl the average trade flow among members. It
should be noted, however, that while the existence of
TD could lead to an inflated estimate of (,TC, it could
not result in the finding of a GTC effect where none
exists since by definition TI) cannot occur unless GTC
occurs as well.
VOL. 63 NO. 5
AITKEN: EUROPEAN
TABLE
1
EQUATIONS FOR EUROPEAN
REGRESSION
885
TRADE
TRADE FLOWS
Coefficients of Independent Variablesa
Year
Constant
D_1
1951
1.958
1952
2.130
1953
2.155
1954
2.052
1955
1.915
1956
2.000
1957
1.900
1958
1.901
1959
1.848
1960
1.617
1961
1.617
1962
1.562
1963
1.589
1964
1.520
1965
1.349
1966
1.252
1967
1.067
- .427
2.53
-.499
2.84
-.509
2.89
- .484
2.63
- .452
2.58
- .476
2.80
- .448
2.57
- .444
2.55
- .449
2.71
- .383
2.34
- .398
2.59
- .410
2.75
- .440
3.06
- .444
3.28
- .392
2.87
- .389
2.93
- .349
2.74
Yi
j
1.137 1.000
8.81
7.76
1.163
.876
8.92
6.72
1.200
.839
9.28
6.49
1.110
.816
8.27
6.08
1.081
.810
8.35
6.26
1.075
.773
8.54
6.14
1.118
.776
8.64
5.99
1.069
.740
8.18
5.66
1.123
.849
8.91
6.74
1.215
.903
9.65
7.17
1.209
.826
10.30
7.03
1.150
.925
10.18
8.18
1.106
.891
10.03
8.08
1.120
.959
10.88
9.31
1.108
.899
10.62
8.62
1.111
.880
10.88
8.62
1.052
.911
10.39
9.00
N
N
pN
i
FATt
A
- .493
3.60
-.567
3.99
-.599
4.17
- .482
3.23
- .463
3.20
- .482
3.41
- .505
3.48
- .481
3.25
-.484
3.44
-.578
4.09
- .551
4.17
-.474
3.72
- .441
3.55
- .439
3.80
- .421
3.64
-.398
3.54
- .331
3.03
- .476
3.48
-.350
2.46
-.350
2.44
- .354
2.37
- .312
2.16
- .276
1.96
- .304
2.09
- .281
1.90
- .396
2.81
- .429
3.04
- .375
2.84
-.456
3.58
-.376
3.02
- .442
3.83
- .396
3.42
-.354
3.15
- .369
3.38
.480
1.85
.490
1.80
.515
1.89
.686
2.40
.759
2.80
.742
2.80
.788
2.89
.766
2.82
.758
2.94
.782
3.07
.798
3.33
.806
3.45
.789
3.47
.747
3.51
.828
3.84
.825
3.94
.892
4.41
pEc
- .141
.46
-.213
.66
-.130
.40
- .146
.43
- .091
.28
- .098
.31
- .019
.06
- .008
.02
.204
.68
.402
1.33
.475
1.68
.527
1.91
.580
2.17
.630
2.52
.743
2.94
.802
3.26
.887
3.75
.053
.23
-.078
.32
- .068
.28
-.111
.44
- .098
.41
- .107
.46
-.136
.56
- .159
.66
- .109
.47
- .103
.45
.045
.21
.076
.36
.172
.85
.326
1.73
.345
1.81
.435
2.35
.572
3.21
R2
S.E.
.766
.277
.760
.292
.769
.292
.755
.306
.777
.291
.782
.284
.779
.292
.770
.291
.802
.276
.815
.273
.831
.257
.846
.251
.854
.244
.870
.229
.863
.232
.871
.225
.874
.217
a Xij is the dependent variable; all variables are expressed in logs; t-values shown in italics, where 1.66 and 2.36 are
significant at the .05 and .01 level, respectively.
preference coefficients come very close to
fitting the expected theoretical pattern. In
all the preintegration years (1951-58) the
PEEC
coefficient is not significantly different from zero and even has a negative
sign.9 In 1959, the first year of integration,
I Despite the consistent negative sign in the preintegration years there is a small perceptible movement
in the pEEC coefficient in 1957 and 1958. This small
movement could be reflecting the effect of any of a
number of different underlying factors, including the
French devaluation, the European Coal and Steel Community, or a possible lagged effect of European trade
liberalization. It seems unlikely, however, that any of
these factors could have been resnonsible for the larg,e
changes in the coefficient which occur during the inte-
there is a sharp increase in the value of the
coefficient (at least relative to earlier
years) as the coefficient becomes positive
gration period. Since rapid progress was made in the
liberalization of intra-European trade during the early
1950s (see Robert Triffin, ch. 5, and the OFEC, pp.
31-32), one would expect the resulting trade stimulation
to be reflected in the regression equations during the
early years of the study period. The Aij coefficient exhibits a definite increasing trend from 1951 through
1957 after which it tends to stabilize at around 0.8
until near the end of the studv period where it shows
another small increase in value. The behavior of the A i,
coefficient, therefore, is more consistent with the expected timing and magnitude of intra-European trade
liberalization than is the smaller and later preintegration movement of the pEEC coefficient.
886
THE AMERICAN ECONOMIC REVIEW
for the first time. The value of the coefficient continues to increase in subsequent
years, reaching a significance level of .1 in
1960 and becoming statistically significant
at the .05 level in 1961. The change in sign
and large change in value of the PEEC
coefficient (relative to earlier years) from
1958 to 1959 is consistent with the hypothesis that the first EEC effect on member
trade occurred in 1959, but the hypothesis
of no EEC effect cannot be rejected at the
standard .05 confidence level until 1961.10
In dealing with the methodological issue
of selecting a base year for making projection estimates, however, the appropriate
methodological question is not whether
the null hypothesis of no EEC effect can
be rejected at the .05 level, but rather
whether or not the hypothesis of no EEC
effect can be accepted. To make the latter
decision, the hypothesis to be tested for
acceptance (i.e., no EEC effect) becomes
the alternative hypothesis and the hypothesis that the EEC has had a positive effect
on trade becomes the equivalent of the
null hypothesis (i.e., the hypothesis to be
tested for rejection). Clearly, the positive
pEEC
coefficients for 1959 and 1960 and
the change in sign and large increase in
value from 1958 to 1959 do not permit the
rejection of the hypothesis of a positive
EEC effect and hence the hypothesis of no
EEC effect cannot be accepted for those
years. The year 1958, therefore, constitutes the last date for which the regression
10 The standard t-test for the significance of the coefficient constitutes only part of the statistical evidence
which should be considered in evaluating the null hypothesis for 1959 and 1960, since it does not take into
consideration all the information provided by the total
set of yearly regression results. A runs test on the yearly
sign of the PEEC coefficient for the 17-year period yields
a probability of .00008 that the sign pattern could have
occurred by chance. A test of the sign of the yearly
change in the value of the coefficient, to test whether or
not the coefficient is changing in a random manner,
gives P> .97 that the coefficient is changing in a random
manner for the preintegration period (1951-58), but
P<.001 (ten consecutive increases) for the integration
period (1958-67).
DECEMBER
1973
results allow one to assume that there was
no EEC effect on member trade.-1
The coefficient for the EFTA preference area follows a pattern similar to that
of the EEC. In the preintegration years
(1951 through 1959) the coefficient is insignificant and in all but the first year it
carries a negative sign. In 1960, the first
year of integration, the sign is still negative, but since the tariff reductions took
place in July of that year, there would
have been very little time for the new
tariff levels to affect trade flows. The coefficient becomes positive (although it is
extremely small) in the following year,
grows slowly through 1963, and becomes
statistically significant (.05 level) in 1964.
The behavior of the PEFTA coefficient is
thus consistent with the hypothesis that
the first EFTA effect on trade took place
in 1961, although the null hypothesis of no
EFTA effect cannot be rejected at the .05
confidence level until 1964.12 All the other
estimated parameters in the equations
have the correct sign and all are significant at at least the .05 level in every year.
11 Studies whiclh have used a base year beyond 1958
assume that the EEC could not have affected trade in
1959 and/or 1960 either because the 1959 tariff cuts
were extended to all GA TT members or because of an
assumed lag in the response of firms to the tariff reductions. The 1959 internal tariff cuts of the EEC, however,
were specified in the Treaty of Rome, signed in March
1957, while the decision to extend the 1959 tariff reductions to non-EEC members was made by the Council of
Ministers only on Dec. 3, 1958. Furthermore, the extension of the 1959 tariff reductions to GATT members was
subject to the important proviso that no tariff would be
reduced below the eventual common external tariff
(CET) for the product. Since the tariffs of Germany
and the Benelux countries were already below the C/ET,
the tariff reductions to other GA TT members were
limited, for the most part, to French and Italian imports. (See the Three Banks Review.) It is by no means
clear, therefore, that there could not have been an EEC
effect on member trade as early as 1959 and certainly
the question is not above empirical investigation.
co12 P= .0007 that the sign pattern of the pEFTA
efficient could have occurred by chance. P> .92 that the
coefficient is changing in a random manner for the preintegration period (1951-59); P <.01 that change is
random during the integration period.
VOL. 63 NO. 5
AITKEN: EUROPEAN TRADE
B. Dummy Variable and
Projection Estimates
The tra(le preference coefficients provide a measure of the factor by which
normal trade among members has been
increased as a result of the formation of
EEC an(d EFTA. Estimates of GTC for
each tra(ling community, therefore, can
be derived from the coefficients for each
year of the respective integration periods."3
These estimates are presented and discussed below in conjunction with the
projection estimates.
As noted above, 1958 is the last year for
which it can safely be assumed that there
was no EEC effect on European trade.
Accordingly, 1958 was selected as the
base year for making the projection estimate of what trade would have been in the
absence of economic integration.
In order to make the projections, the
1958 equation was recalculated leaving
out the two trade preferencevariables and
the following results were obtained:
(2)
log Xij = 1.978 - .487 log Di3
(3.76)
+ 1.062log Yi + .733 log Yj
(8.33)
-.459
(5.75)
logXN - .259 log lY
(3.36)
(1.90)
+ .718logAi, + log ei1
(2.84)
W2 =
.776
S.E. = .289
The t-values are all significant above the
.05 level.14
13 Since raw data values of 1 and 2 were used for the
dummy preference variables, 2b7provides an estimate of
the factor by which the average trade flow among EEC
members has been increased as a result of integration.
Actual intra-EEC trade divided by 2b7,therefore, yields
an estimate of what intra-EEC trade would have been
in the absence of integration. This value in turn, is subtracted from actual trade to obtain the estimated GTC
of the EEC for the given year.
14The error term was included in the equation because the cross-section residuals tended to be stable
887
Equation (2) cannot account for the
trade effects of changes in competitive
position among the countries in the
sample15nor can it account for any general trade liberalization effects which may
have occurred after 1958. The projection
estimates are made, therefore, on the basis
of the usual assumption that the effect of
these factors on trade has been small relative to the effects of integration. For the
estimation of GTC, this assumption is
supported by the regression results of the
present study, where the pronounced increase in the trade preference coefficients
indicates that economic integration provided the major impetus for additional
trade (with the effect of income held constant) among the respective members of
the two blocs. For trade between the two
blocs, however, the model cannot provide
information as to the general magnitude of
the TD (or ETC) effect.
The exponential form of equation (2)
was used to estimate intra- and inter-EEC
and EFTA trade and the estimated trade
values were then subtracted from actual
trade to obtain the residual estimates of
the trade effects of the EEC and EFTA.
Since equation (2) estimates the dollar
value of trade in 1958 prices, estimated
trade was multiplied by country i's export
over time. Real GNP in 1958 prices was used to make
the projection estimates since it provided the best measure of how the economic capacity of each country has
changed relative to the base year. A detailed explanation of these points is provided in an extended mimeographed copy of the paper.
15 The fact that there was significant dispersion in the
export price trends of the individual countries within
each trading community over the projection period (see
International Financial Statistics, Aug. 1968, p. 30)
means that the projection estimates for each country
are likely to be subject to error, but it also means that
there will be at least some tendency for the errors to
cancel out in the aggregate results for each community
as a whole (i.e. not all competitive effect errors will be
in the same direction). Consequently, less confidence
can be placed in the individual country estimates than
in the aggregate community results. Even in the latter
case, however, the results should be considered only as
indications of approximate magnitude.
DECEMBER
THE AMERICAN ECONOMIC REVIEW
888
TABLE
2
1973
NET EFFECTS OF INTEGRATION ON E,A1CAND EFTA TRADEDUMMY VARIABLE AND PROJECTION ESTIMATES
(Millions of Dollars at Current Prices)
Net EFTA Effect on:
Net EEC Effect on:
EEC ExportSa
Year
11959
1960
1961
1962
1963
1964
1965
1966
1967
EFTA Exportsb
Dummy
Variable
Estimated
Projection
Estimatee
Projection
Estimatee
1,067
2,468
3,284
4,114
5,203
6,388
8,228
9,784
11,127
925
1,639
2,254
3,213
4,731
5,695
6,941
8,612
9,189
50
31
67
393
541
202
- 41
-157
-629
EFTA Exportsa
Dummy
Variable
E"stimated
Of
Of
126
222
545
1,151
1,326
1,773
2,425
lE C Exportsc
Projection
Estimatee
Projection
Estimatee
-8
140
149
243
389
573
690
919
1,264
66
48
-102
-201
-262
-289
-259
-205
-202
Estimates of Gross Trade Creation (GTC).
Estimates of the net external trade creation (ETC) or trade diversion (TD) effect of the EEC on the exports of
EFTA countries.
e Estimates of the net TI) effect of EFTA on the exports of EEC countries.
d Dummy variable estimates of GTC derived from the pEEC and pEFTA
coefficients reported in Table 1.
e Actual trade in current prices minus trade estimated by equation (2) converted to current prices.
f Zero values are given for 1959 and 1960 since the pEFTA coefficient is negative for those years.
a
b
price index (dollar prices, 1958 base) in
order to obtain estimates of the effects of
integration in current prices."6 The projection estimates therefore can be compared
directly with the dummy variable (DV)
estimates which reflect the prices of the
given year for which the regression equation was calculated. The projection estimates for each of the four subgroupings of
the sample are contained in Table 2 along
with the DV estimates of GTC.
Because the economic integration of
both trading communities has been a
cumulative process (i.e., internal tariffs
being reduced by stages over the integration periods covered by the study), one
should expect to find the estimates of
16 The use of total export price indexes introduces
error into the results since composition of exports to the
EEC and EFTA may differ from total export composition (although each country's exports to the respective
trading communities would, in general, cover a broad
spectrum of its total commodity exports).
GTC increasing from year to year with no
reversals. Both the DV and projection
estimates of the respective communities'
effect on member trade are consistent
with this expectation. For the EEC, the
projection estimates are consistently below the DV estimates, with the gap increasing at the end of the study period.
Despite this discrepancy, both estimates
show a large EEC effect occurring in 1959
and a strong cumulative growth in the
yearly values of GTC. For EFTA, both
estimates also show a cumulative growth
in GTC, but the DV estimates are below
(but reasonably close to) the projection
estimates through 1962. Starting in 1963,
however, the DV estimate is greater, with
the gap increasing progressively in subsequent years. Even if allowance is made for
a large margin of error, the results clearly
indicate that the GTC of the EEC has
been substantially greater than that of
VOL. 63 NO. 5
AITKEN: EUROPEAN
EFTA. I-n percentage terms, the projection estimate of GTC accounts for 38 percent of actual intra-EEC trade (import
data) and 16 percent of total EEC exports
in 1967. For EFTA, the projection estimate of GTC for 1967 accounts for 16 percent of intra-EFTA trade and only 4 percent of total EFTA exports.
.In considering the effects of each trading community on the other, EFTA could
only be expected to have a negative effect
on EEC exports since there was no reduction in the external tariffs of EFTA
countries. Furthermore, the size of the
TD effect should increase progressively
throughout the inte.gration period since
increasing trade discrimination against
nonmembers would result from the progressive reduction in internal tariffs. The
estimated EFTA effect on EEC exports,
therefore, is only partially consistent with
theoretical expectation. The fact that the
estimated net effect becomes negative in
1961 (the first full year of integration) and
the existence of increasing negative values
through 1964 is consistent with the TD
hypothesis, while the indicated reduction
in the size of the TD effect during the last
three years of the period is not. Disaggregation of the EFTA effect on EEC
exports by individual EFITA importing
country revealed that the decrease in the
"TD effect" over the specified period was
accounted for almost entirely by the
United Kingdom, with the net effect on
EEC exports increasing from -$38 million in 1963 to $530 million in 1967. Since
the period 1964-67 coincides with the
general decline in the U.K. balance of
trade position culminating in devaluation
at the end of 1967, the divergence of the
U.K. effect can be attributed to the serious
deterioration of the U.K. competitive
position. The net effect on EEC exports of
all EFTA countries other than the United
Kingdom over the same period was found
to be consistent with the expected increas-
TRADE
889
ing TD effect (the estimated net TD effect
increased progressively from -$223 million in 1963 to -$731 in 1967).
Because of the reductions in external
tariffs by the original high tariff EEC
countries (i.e., France and Italy), both
ETC and TD are theoretically possible
effects of the EEC on EFTA trade. The
TD effect would he expected to dominate
eventually, however, since the reduction
of internal tariffs toward zero would eventually result in trade discrimination
against nonmembers, even in the markets
of the original high tariff countries. The
estimates of the net EEC effect on EFTA,
therefore, are not inconsistent with theoretical expectation, since they show a general increasing net ETC effect through
1963,17 followed by the emergence of a
growing net TD effect over the last three
years of the period.
Finatly, the fact that TD is dominating
trade between the two communities over
the latter years of the period explains the
growing (livergence between the DV and
projection estimates for these years. It can
be concluded, therefore, that the DV estimates are definitely inflated for the years
1965--67 and possibly for earlier years as
well. Hence, the DV estimates must be
rejected in favor of the projection estimates for at least the last three years of
the period.
C. Individual Country Estimates
IH[aving found that the results for the
two trading communities as a whole are
consistent with theoretical expectation,
we now proceed to the question of whether
the expected results hold for each of the
individual countries in the sample for the
17
Disaggregation of the EEC effect by original high
and low tariff EFC countries provided additional support for the 1E1TC hypothesis since the results showed
that the increasing ETC effect for the period 1959-63
was entirely accounted for by the original high tariff
EEC countries.
THE AMERICAN ECONOMIC REVIEW
890
3-NET
EFFECT OF EEC AND EFTA ON
TRADE OF INDIVIDUAL COUNTRIES--PROJECTION
TABLE
1967
ESTIMATES
(Millions of Dollars at Current Prices)
Exporting Country
Belgium-Lux.
France
Germany
Italy
Netherlands
Total EEC
Austria
Denmark
Norway
Portugal
Sweden
Switzerland
United Kingdom
Total EFTA
Net EEC
Effect
1649
2170
2473
1958
939
9189
-123
-350
- 30
- 42
-100
-229
245
-629
Net EFTA
Effect
-
39
379
-390
260
-412
-202
234
144
211
107
397
173
2
1264
last year of the study period."8 This
represents a more stringent test of the
GTC and TD hypotheses, since there is
less opportunity for errors caused by
competitive effects to cancel out in the
estimates for the individual countries
than there is in the aggregated community
results. Table 3 contains the individual
country projection estimates for 1967.
Because of the possibility of error, intercountry comparisons in the size of the estimated effects should be avoi(led. The
results should be viewed primarily in
terms of whether or not they are consistent with the GTC, TD, or ETC hypotheses.'9 WVith the exception of the
small negative value for the United Kingdom, the estimated effects of the EEC and
EFTA on the trade of the respective member countries are all positive and hence
consistent with the expected GTC effect.
18 Copies of individual country estimates for earlier
years are available from the author.
19 Because of the large magnitude of the difference
in GTC estimates between the respective members of
the two communities, it is possible to conclude that the
GTC of each PFECmember has been greater than the
GTC of any EFTA member.
DECEMBER
1973
(Again, the discrepancy for the United Kingdom can be attributed to the deterioration
of the U.K. competitive position.)20
The estimated net EEC effect on the
individual EFTA countries indicates that
all EFTA countries other than the United
Kingdom were experiencing net TD as of
1967.
Disaggregation of the 1967 EFTA
country estimates by original high and low
tariff EEC countries showed that the high
tariff countries accounted for $238 million
of the total EEC effect on the United
Kingdom and hence the positive effect for
the United Kingdom can be attributed to
ETC.21 Sweden ($82 million) and Switzerland ($7 million) also showed evidence of
an ETC effect by the high tariff EEC
countries, but it was more than offset by
a much larger TD effect from the low
tariff countries.22 Consequently, the results indicate that while TD was the domiiinant effect of the EEC towards EFTA in
1967, the three major industrial counitries
of EFTA continued to benefit from ETC
in the French-Italian market.
The estimated EFTA effect on EEC
countries shows positive values for both
France and I taly which are contrary to
theoretical expectation, since only a ID
effect could be expected.23 While it is pos20
The estimated EEC effect on each of the five member countries was positive for all years of the periocd
1959-67 and the estimated EFTA effect on each of the
seven member countries was consistently positive after
1962 with the exception of the United Kingdom in 1967.
21 While ETC by France and Italy explains virtually
all of the positive UT.K.value for 1967, the fact remains
that there is no evidence of a net TI) effect hvrthe low
tariff E'EC countries against the United Kingdom as of
1967. (This was true for earlier years as well.) This finding is surprising in view of the fact that TI and the
worsening U.K. competitive position should 1)oth be
operating to produce a negative effect.
22 A net TD effect by both high and low tariff E'EC
countries prevaile(l in the case of the four remaining-less industrialized--countries of 1E'FTA.
23 The net effect of all EFTA countries other than the
United Kingdom was as follows: Belgium-Luxembourg
(-131), F'rance (230), Germany (-539), Italy (125),
Netherlands (-416). Consequently, while the removal
VOL. 63 NO. 5
AITKEN: EUROPEAN
sible that these two countries experienced
only minimal TD by EFTA, one certainly
cannot rule out the possibility that the
estimates may have been largely influenced by competitive effects. Consequently, while the net effect of EFTA on
the EEC as a whole is consistent with the
TD hypothesis, there are sufficient discrepancies among the individual country
estimates to temper this conclusion with
reservation.
III. Summary and Conclusion
The empirical findings of the study were
found to be generally consistent with the
expectations of customs union theory. The
results showed that both the EEC and
EFTA have experienced a cumulative
growth in GTC over their respective integration periods, with the GTC of the EEC
being substantially greater than the GTC
of EFTA. The projection estimates for
1967 placed the size of the GTC effect for
the EEC and EFTA at approximately
$9.2 billion and $1.3 billion, respectively.
The estimated effect of EFTA on the EEC
as a whole showed a consistent TD effect
for the period 1961-67, but there were two
positive deviations from the expected
negative effects of EFTA on the exports of
the five individual EEC countries. The
EEC was found to have had a net ETC
effect on EFTA through 1964, but this
was replaced by a growing net TD effect
from 1965 through 1967. The estimated
effects of the EEC on both membercountry trade and EFTA-country exports
were found to be consistent with theoretical expectation.
The regression results of the study
yielded the important methodological conclusion that 1958 is the last year for which
it can safely be assumed that European
of the United Kingdom from the EFTA effect moves
the estimates more in line with the TD) hypothesis,
positive values still remain for France and Italy.
TRADE
891
trade was unaffected by the formation of
the EEC. This finding, in turn, suggests
that an underestimation of the EEC effect
on member trade could result if a base year
beyond 1958 is used to represent normal
preintegration trade.
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