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Automobile prices and protection: The U.S.-Japan trade restraint
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We examine tireeffectsof the U.S.-Japan trade restraint on automobile prices and quality u@grading,for both Japanese imports and American small Gars.From Aprir 1981 to April 1984 the suggested retail price of aCJapanese models mcreased by 15.8 percent, or ,“1,3percent per year. Ilk f&xi that nearly the entire amount of this rke can be expiained by the upgntding of individual models. This upgrading may benefit consumers who would purchase a luxury import k~any case, but harn~ those who desire the basic imports. In a.ddkiun to upgradings we conclude that a seconli cost of the trade restraint has t>rznto pr~ent the yen depreciation from being passed onto American consumers, in term of lower imported auto prices. For U.S. small cars we fmd a 9.1 percent rise in the suggested retail prioe, or 3 percent per year, with a frmctionof this amunt due to model upgrading. If the yen depreciation had led to lower inqort prices without the trade restraint, then we expect that U.S. auto prices would have been lower, too.
A prominent feature of recent U.S. trade policy has been the restraint on automobile imports from Japan. Beginning in April 19815;this “vo:luntsry export restraint” (VER) has limited Japanese auto sales to 1.68 million annually. The restraint was supposed to expire in March 1984, but instead, under U.S. pressure, the wiling was raised to 1.85 million and extended for another year. This trade arrangement is a particularly important example of the “new protectionism” since the auto industry was initially denied tariff protection under the esapeclause provision of the General Agreement on Tariffs and Trade (GATT). Thus, the rules for international safeguards embodied in the GATT were bypassed in favor of the VEIL The effect of the trade restraint on import and domestic auto prices has been the subject of recent debate in the U.S. A study by Wharton
Recoiled September 1984; acwpted November I 984,.
&onometrics (1983j zpors that from 1980 to 1982 the rise in average new car selling price; was $2,600, and that $I,OOOof this amount was due to the VER..” on the other hand, U.S. automotive pFU&MBrS &ve pc~ir&ti opJlf that the consumer I~F& index of t& &XFMB of lliabor !%&s~~cs (BL!$) has increased by an annuiaxaverage of 4.3 percent ~OF autos over 1.981-84, as compared with 4.7 percent for all ikms.* It should be noted that t$reBLS price index corrects for quality changes in autc)mobiks, such as the addition sf greater size, improved engine features!,etc. Thus, an upgratig of car models and a correspl>nt;iingrise in price will n;;t be seflected in that k&x. This is a very signii‘rcant eorrectkbn since the trade restraint has led to an upgrading of Japanese models, as observed by Feenstra (1984) during its first year? IIn this paper we shall exanke the effect of the trade restraint on automobile prices and quality upgrading, for both Japanese imports and Arnerics:2 small cars,. Summarizing our results, from April 198 P to April 19384the sugg&& ret:d p; ices of alI Japanese models increased by 15.8 percent, or 5.3 percent rer jrear. We fmd that nearly the entire amount of thi.s rise can be explained by the upgrading of individual models. In ciolllarterms, import prices have risen due to upgrading by some $300 each year, #+ving a $KKI rise by 1984. This upgrading may benefit consumers who would purchase a luxury import in any case, since they now have mze models to choose from, but harms consumers who desire the basic imports. In addition to the quality upgrading, there has been a second consumer cost of the VER When the trade restraint began, in April 198 1, the Japanese yen was at one of its strongest levels in recent history. Immediately thereafter, the yen started its well-known depreciation against the dollar. This exchange rate movement would normally
Iead to lower p&es for Japanese imports, but because of the trade restraint, was not refIeci:ed in reduced import prices for autos. We
IWharton Econ ometric Forecastinp Associates (1.983). In Wharton Motor Vehic e Service (January 1983) Specr’nlAnaZy~&~Die Japanese QUO&,the ptic: rise due to thl: VER is reported more precisely as $85 1 for imported autos over the period 1980 to 1382, and $324 for domestic actos. ?kepsc are seasondy unadjustlzd f;sures, average from March 198 1 to March 1WM. The BLS producer price .x+ Yfor autos: rose by an annual average of 4.1 p xent over the same period. *he quality upgrading can be car idered a vclns~~~~enc*~ of the $!1:Wsince it applies to the of UEOS @mted. if mstzsd the trade restraint applied to tke v&e of imports, as with a Mff, then we would not ex~xct producers to shift tow&s the higker price, higher quality molclels. mm&r
conclude that a second cost of thz VER has been to prevent the yen depreciation from bein;;;passed on to American consumers, in terms of 143WW auto prkes. TurrtEne, to U.S. autos, we investigatethz price and quality upgrading of subornpact and act mod&. From April 198f to April 1984, we fbd a 9.1 petcent tie in the suggested retail price, or 3 percent per vear. A f!ration of this amout is due to model upgrading. These figure; can be interpreted in two ways. On one hand, if the yen depreciation had led to tower import prices without the VER, then we expect that U.S. aut:o prices would have been lower, too.4 This point is underscored by noting that b the two years just prior to the VER, the domestic auto industry experienced some:of its largest price increases ever-22 percent and 17 percent, respectively, for U.S. small cars over 1979-80 and 19868 I. Thus, Hhife the recent increase in U.S. small car prices has been moderate ;tt 3% per year, we should recognize that: the histancally high prices of 1981 have been maintained. Without the protection provided by the trade restraint, in?port and domestic auto prices would have been lower. On the other hand, Gnce the actual rise in U.S. au:0 prices has been moderate, it follows that the recent surge in profits of domestic automakers cannot be Ij.ttrjbuted to price increases, but instead must be explained by volume increases and various cost reductions. Since the VER had led to less price Lompetition in the U.S. market, domestic companies have been able to earn record profits from the volume increases and cost reductions rather than lowering prices for consumers. We conclude by arguing that a greater inflow of .lapanzse autos should be permitted when the trade restraint expires in March 1985. This would
ensure that the cost reductions arPd adjustment in the U.S. industry would be passed on to consumers, through expznded trade and lower auto prices. 2, RECENT EXPERIENCE
IN THE USI, AUTO INDUSTRY
On May 1, 19815 the Japanese govr:rnm’ent announced J three-year system of ‘%oIuntary export restraints” (VER) on the export of passenger cars to the U.S. market. Fw- the period from .A March 1982 these exports would not exceed 11.68 ~~~~~~~, sceond year (April I982 to MarGh 19 raised according to the growth in thcb
‘?‘he impact of the VER on U.S. domestic car price Crandall (1984).
is zxnmincd in gn:ater detail in
These-acGons were made against a backgrounti of falEng production and high unemployment in U.S. autos, along with several legislative attempts to curb imlxwts. For eXUnple5b early 1951 Senator%Danforth and Ben&en.introduced : A.! (S.3196)to impose quotas on the import of automoM from 3apan of 1,6 million units during 198143. Indeed, fitis bill,‘was schedulledfor markup (line by line revision) in the Senate P’itlanceCommittee on Map 12, and no doubt cm&but& ts thtsspecific &or1 taken by the Japanese.Other bills included more stktigent import ~ucit;~~s, and domestic content requirements that specify the minimum c~plt tent of American parts for autos sold in the U.S. Supported by the UAW, the domestic content bill passed the House of Representatives in 1982 and 1983 (HlL5133, 1982 and HR.1234, 1983), and has been introduced into the US. Senate iin 1984 (S.707). An ear&r legislative action was thy petition for import r&f by the UAW in June 1980, made to the U.S. Enternational Trade Cor~nission @K). In August 1980 the Commission rweived a petition for sixilas import relief from E;ord Following rules established in the GATT, a x-wmmendation fbr relief ean be given only if the ‘5ncreased imports of an zii& are a substantial cause of se:&us injury, or threat thereof, to the domestic indwtry.” The U.S. tra& Pawdefines the tx n “substantial cause” as “a cause which is imprtrant and not less than any other clause.” The ITC cletern%nedthat, while imports of autos into the U.S. had increased aad the domestic ind stry was in fact injured, the RX~sion k the US. WaS a greater 633 of injury than the increased 5 A~~~~r~~g~~,import relief was not given. It was also found that t In ~~~~~~~~ preferences t~~~~d~ §~~a~,fuel efficient autos (due In part to rklng g;W3”inne s) was an *nt tfian tfie recess~unary i
and
A U T O M O B I L E PRICES A N D P R O T E C T I O N
53
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]InTable 1 we present backgrouuzd data for the awto industry during the perioct to which the VER applies. In the per&d from Apil1980 ta March 2981, tie year immediately ~rec&ing the VEiR,U.& lpraductiun a& emp~syz~a wm at dqressed leveis while the Japanese market share of 21.2 per-t was a historic high. From the second ruw we see the US. p&u&ion has picke~Iup in the most recznt year, wKlti U.S. consum~tiu~ began increasing in the period frownApril 1982 to March 1983. The rise in total consumption has led ts a drop in t2’reJapanese market share. Employment has also increased, though this occurs more slowly than tSrerise in production: as shown in tiJe Anal row, the number of cars produced per worker has increased from 8.1 to 10.2 recently. The rise in the average product of labor is in part a normal occurrence during rhe upswmg of a business cycle, and also reflects new invertment and production trdtniques in the auto industry. Ib determine the impact uf the VER on prices, a sample of Japanese i~r:portand American small (i.e., subcompact and compact) models was ta8t;n from AactomutivePVms Market Dutct.Books for 1979-84. The sample covers the base yrersion (i.e., without options) of every model a~dailablein each year, with the exception of “upility” vehi&s and station -w:Agom?In additisl: to the suggested retail price in March or April of tiz!z year, data were obtained on the quantity sold and model length, width, &eight, hoisepwer, type of transmission, and whether the base version had power steering, power brakes, or air coaditiir,dng. In Section 5 HKshall use these characteristics to estimate the\quality of c.ar models. In 1:heMowing sections we shall report our resust; ,Cor.?apanc:se imports and hmefkan smaII e..A.os,respectively. Before txning to 3ur results, ‘weshould note several qualifications. By using suggested retati prices we are ignoring additional deaIer markups fat& QI consumers, x&ich hav: been substantial in recent years. To this extent oilr price data gives an ul;4erelitimate of the consumer cost of the X3%. In addition, we are ignoring model options which a not provided as standard equipmerlt but which a cOnsumer must purchase to obtain a car. Anccdotd evidence suggests thar: such “‘required options’” have been ~~~q~~~~~y [ised in the plarchase of some recent: Japanese models. Data on thi; practice arc wt available, l-wwever, and WCconsider only the uS;i~ qgrading t.h;lt fras occurred on the base versions (i.e., without options) OF-eacl” model.
AUTOMOBILE PRICES AND PWEECTIQN
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Unit va\ue (% change) price index (% chage) Number of models New models unit quzklity (% change) Quality index (% change)
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55
_~~
1979
1980
1981 -
1982
1983
1984
4,946 80.8 21 5,077 92.J -
5,175 (4.6) 83.5 (3.3) 24 3 5,195 (2.3) 93.5 (1.3)
6,211 (20.0) loo.0 (19.8)
6,834 (10.0) 107.8 (7.8) 24 ob 5,942 (6.8) 105.1 (5.1)
7,069 (3.4) 109.5 (1.6) 26 4b 6,316 (6.3) 109.7 (4-4)
7,459 (5.5) 115.8 (5.8) 25 0 G,S56 (3.8) 114.0 (3.9)
24 0 5,564 (7.1) 100.0 (7.0)
uObtain& front Qutomotive Kews, Market Data Book Issue, 1979-84 years. bin 1982-83 the Datsun Sentra, Pulsar, and Stanza replaced the 210, 3 10, and 5 10, respectively, and are not counted as new models.
3. PRICE AND QUALITY OF JAPANESE IMPORTS
In Table 2 we show information for Japanese imports over 1979-84. The unit values shown in the first row are a weighted average cf suggested retail prices of base models, using the quantities sold as weights. The prices for 198 1 were those in effect on April 8, just prior to the annoutcement of the export restraint. Accordingly, we shall treat 193 1 as our ‘&baseyear” for comparisons. The unit values show a rise of 20 percent, from $6,211 in 1981 to $7,459 in 1984. However, this rise includes both the effect of increasing prices for individual models and the shift in quantity demanded towards higher-priced models. A more appropriate measure of the overall price increase is to construct a price index using constant weights between each two years (see Section 6). Reported in the second row, this price index shows an incrl:ase of 15.8 percent in suggested retaiJ prices between 198 1 and 1984,of 5.3 percent per year. The variety of Japanese models o;fered was quite stable over the p~‘r~o~$as shown by the number of existing and new rnodeIs ez\ch year. irA the nljsa7ber of WI&z there was little cha qualities of individual mode
rutted using constant ~~~gh~s irK.kxshalws 23tise sf B4 percent over
retail
prices.Thus, nearlyall of the rise
u&f
in
dmparr prices cm be
explained by t/w upgrading of indtidual modelk. Thr: dollar krease in quality can be computed through multiplying
the change in t5: qu&y index times $6,X 1, the 1981 base price. This gives a quality 15:,eof about $300 per import each year, leading to a $900 increa:;e by f 984 The quality upgrading has a consumer ccst that varks ac-:3ss individua’s. The upgrading may benefit consumers who would purchase a luxury import in any case, since they now have more models to choose from, ~utiharms those who desire the basic madeis. 11is prhaps s~urpriskg that most of the price rise in imports can be attriiut&! to qustity upgrading. We would normally expect the trade restraint to havt: a further upward effecd;on prices, independent of +4&y, by creating an &fkial shortage. We shall nc-1 argue that the VEIf .aas had wch an upward effect on prices, by prevtllting movements irkGre df E;ar/ycn exchange rate from being passed through in import prkc;. 111Figure f we show the quarterly dollar/yen exchange rate over the ast sez3-al ve88:3(o&be to ~~~~~~~) = KIO). In the firsi quarter of 98 9 the y’ n was at me of it s stron:gest bevels in recent history, having a p~~eaated 18 percent from 8he p vioamsyem. ~~d~~d~ this s.pprec;ation ~~~~~~~~~~~very closely f831 ttle z percent ~-kc:in Japanese auto prices fran 1980 to I!318 li (see Tabie 2). ~rnme~at~~yafter the VER was in
AUTOMOBHE PRICES AND PROTECTION
57
place the yen began its depreciatios; against the dollar,,due in part to hligh U.S. interest rates. Relative to the first quarter of 198 19 his depreciatiun eweds 1Q percent fur any of the klllowing fiist-quarter averages, and the @eatest depreciation of 26 per cent is obtained by comparing fmtquarter %981to fourth_quarW 19:U:. Based an this exchange rate behavior, we expect that withcxt the VER Japanese au& prices would have J%%IZsince 198X. This predicticm is reinforced by notm~ :!I& in the year just &fore the VER.,the 1.8percent appreciation of the yen correspond& $0 a nearly equal rise in imported auto prices. It is lcllgicalto expect thz *; the reversal k the yen movement would have led to SWIMS fall k irnp&ed auto prices. However, in the presence of the arMicia1shortage created by the trae restraint, no such deck in auto prims has occurred. We conclude that Qfir&a= impact qf fh43imule restraint has been to prevent t?8eyen depreciutionfrom &ing passed on to American cons’umers,througJtlower imported auto prices., Ilkis consumer cost is in addition to the quality upgrading of impart nxdels tkzt has occurred. 4. PRICE ANR QUALITY OF AMEIWZAN SMALL AUTOS In Table 3 we report the price and quality dzita fcr U.S. subcompact and compact modkls. The unit value, constructed as a weighted average of suggested retail prides of base models, shows a rise CC14.6 percent from 198 I to 1984, However, a considerable portion of fhk rise is due to a demand shift t\(sa7vards larger models. The index of suggested retail
T&k 3: Sample of American §mall .QutoP -Unit ~slue (96 change) Price :ndex (% change) Number of models New models Wnit quality (% change) Quaiity index (% change)
1’97!#
198@
4:186 70.1 24 -
5,067 (21.0) 85.6 (22.1) 22
4,l?5
4,132 (- 93) 47.8 (-0.5)
98.3 -
!
I9tJ .1 --I5,9 f :5 (16 7) 1OG.O
(16.8) 23 7 4,183 (11.2) 18fr.i) (2 2)
1982
1983
X984
6,446 (9.W 107.5 (7.5) 27 4 4,35 I
6,58 1 (2.1) 107.3 (-C.2) 33 4% 4,497 (3.4) lW.4 fi.3)
6,73 I (3.0) 109.1 (11*?b 34 7 4,563 (KS) Ii03.8 (C.4)
(4.0) 102.11 (2. I)
aObtained from Au&mmtive News, Market Data Book Issue, 11979-84 years. bin 1983 the Dodge 400 and Chrysler LeTZaronwere down-sized from %rlier models. and are not counted as laew.
prices,construtied using wnstant weights between irmwses by only 3.1 permnt from 198
I tro
aa& two years, 1984, or 3 percent per year.
This is less than the rise in the overall wnsumtir price tidex ova that petiod (see Se&m 1). We also see that there itrasbeen m active introduction of ne+~models and expiring uf old, Some qutity upgrading of US, models has occurred, though less than the amount of the price rise. Imhe foal rovvof Tab133 we report an increaseof 3.8 percentin the quality index from 1981 tcb1984, which is slightly less than one-half of the rise in the price index. It is also of importance that in the two years just before the trade restraint begqn, U.S. small car prim rose very substantially--Z.!. percent and 16.8 percent, res?ctively, over 1979-80 and 1980-81. This compares with 8 rise in &.Iwnsumer prices of 13.5 percent and 10.4, respectively, over the same periods. Thus, whilethe increase in U.S. small car prices sincl? 1981 has been moderate at 3 percent per year, we can a&natively %e rpret the tidence as saying that the historically high p:icpc of 1981 hwle simply been maintained. This view corresponds with our discussion of the do&¥ exchange rate and import prices. In the absence of the VER, we e,rpct that the yen depreciation would have led to lower import prices. Tlris would have created greater price competition on ‘domesticproducelbs,and some lowering of U.S. auto prices, too. Instead, the trade restrtlin: has insulated the U.S. industry from movements n tie dollar/yen exchange rate, thereby permitting fhe high prices of 1981 to be maintained. However, we should also recognize that the surge in profits of U.S. autcmakc;z cannot be attributed to recent price increases for small cars, swx lhlese increases have beep moderate. Evidence on the extent of the ~;&I: turnaround is 2s folLyas:
Somx:
Autmwtzve
News, various
issues.
earned by con~p~anies during the first quarter of 1984 have broken all previous records, and even exczeded the annual profits earned in many years. The profit surge of U.S. automakers can be explained by a variety of factors, The volume expansion as the demand f’orcars has increased (see Table 1) will lower average costs due to economies of scale, and raise profits. Direct cost reductions have also been achieved as fi=wrer worke?*s are rehired coming out of the recession, and the average product of labor increases (Table 1). U.S. firms have reported that the;break-even volume for profitability has been reduced in each company, reflecting a decrease in fixed costs. In addition, a numlber of recent U.S. models have proved to be very popular among consumers. Lastly, it shouid be noted that a portion of U.S. automakers’ profits are obtained from truck sales, which have been protected by a 25 percent tariff since 1980. While the cost reductions in the U.S. industry indicate that some adjustment to potential competition has taken place, consumers ha;re not yet benefited from this adjustment. The trade restraint has led to 1~ price competition in the 1J.S. market, enabling domestic companies :o earn record profits from the volume increases and cost reductions rather than lclwering prices. As compared with a tariff, the VER gives greater “market power” to domestic producers since the supply of Japanese imports is constrained in quantity? The recent experience in U.S. autos appears to be an important example of the exercise of market power under a quantitative trade restraint, with U.S. firms increasing their profits in response to higher demand and reduced costs, but not lowering consum prices. 5. ESTIMATION OF PRODUCT QUALITY
For each of the Japanese an4 U.S. small car models, data were collectec. on the suggest%! retail price of base versions (i.e.. ~~~~~~~~ options)., quantit:l sold, and various characteristics. wi3: be used as ;1 measure of qu !ity upgrading, the st~t~st~~a~ technique of “hedonic” xgressions. The basic rtzferenct: cm 0-k tee is ~~~~~~~~~ (B 87
I ).
t979-84 q&w an? able lo eq&&i 92 lpercentQf Is%@. %@J%it~mfi :ilhk-ptim @*=8.92$ ti the fiat c0lu.m). The ~~~~t~ Q~~~~.~~~~~~~~~~~~ can be give a preciseinteqsetatiox the coeB%kntof 0.40 fur width~~fsr e3t8nqk, indkstes that an k3xxsi2 in width af one Rx%would r&e the SW& prioeby 40 perc&nt.The transniission va%db~~t&m #ha v&e of unity ii’ the ma&l hairs a Rve-speed or autamtic trz&stissian as stan&rd qyi~ment, atid zero otherwise;the cc#f%i&xtof 0.15 indiu& &at “&isfeetursrlx&MS the 43stillaf;e$ priceby 15pfxcernt.TIM pmna af air ~sc&itktiqg 3s standard equipment also Ssed the &knz&d price by 15 per=nt. J[nthe iFinalregression “tiehave omitted the wei@ and power brake varia!&s since these estrmated co#kients when included were highly insignificant (with star&& errors nearly twice the coeffkient size). Omitting these variables has only a shght efkct on the t:Gser coet&ients (see bhwj. It canbenoticed that tke length variable has a negative and signif cant coefficient, which is unexpected This may be due to mt&icoUinearity with other &aracteristics or misspecifk&on of the equation. Aside iTom the negative sign on length the 197%-84 regression appears to fit ri.ther weX We have included dumy variables for the various years, equ, tling Gty in that year and zero otherwise. Except for 1980 the estirr ated coeffxcientsof the year dummies are highly significant. The estimale of 0.15 1 in 9984, for example, indicates that 15,l percent of the [unweighted) price rise from 1979 to 1984 is not explained by any chaqt;e in cbaracteristrcs. Y’hus,the yeai- coefficients t;an be interpreted as the qrlahty-adjusted price rke. A measure of quality is obtained by computing the predicted price from the hedonic regression not including the portion explained by the year dunmzks. This calcuktion is rewated for each year, obtaining a mt;3sure of moddquality for each model. The unit quality reported in the se=n.! to the last row of ‘Bble 2 is a weighted average ~“VCSS models, using t, \aquantities sold that year as weights (for 1984 WCME the ! 983 quantities as weights). The quality index reported in the last TQW of Table 2 is computed with constant weights between each two years, as dessrikd in Se&on 6. t is clear that our measure of quaky is sensitive to the estimated coeffitients nf the yeas dummies, since t se coefficients measure the ~~a~ty-~dj~~.;ted ~xice rise, We checked e sensitivity of these coeifficients TV)the mo&4 sp3cifkation hy reestimating the f979- 84 regresskn, wme omitting each time on: of the fokving variables: length, th, weight, and horsepower. ?Selow report the range of point ates k’i,r* Ihe year co these regressions:
1981 I982 1983 1984
0.105-0.120 0.139-0.164 0.128-D. 153 O*i51-0.193
Table 4: hpanese Hedonic Regressions 1939-84 &per&m
1979-80
1981-82 _.
198344
Variable: Japanese
Au& Price (Logarithm) Observations
R2 SSR Intercept Length (feet) Width (feet) Weight (tons) Horsepower (100 HP)
Bummy Variables: Transmission (Five-speed or auto) Power steering Power brakes Air conditioning Year 1980 Year 1981 Year 1982 Year I.983
Year 1984
144 0.924 1.282 6.41° (0.48) - o.045a (0.019) 0.4w (0.11) -
45 0.885 0.425 4.95a (8.85) -0.059 (0.036) 0,71Q (0.2 1) -
48 0.937 0.290 6.83” (0.74) -r).D75’~’ (0.034) 0.42* (0.17) -
51 0.929 0.419 8.28C (0.99) -0.059 (0.039) O.lD (0.20) -
0.77a (0.066)
0.76O (0.15)
o.74a (0.11)
0.79a (0.11)
o.15a (0.023) 0.052 (0.029) -
0.1 la (0.040) 0.12 (0.087) -
8.18O (0.03 5) 0.16 (0.050) -
0.23a (0.062) 0.1053 (0.04 1) _-
O.lSa (0.03 5) Q.0084 (0.030) 0.115” (0.030) 0.15ou (0.03 1)
0.072 (0.098) ODD59 (0.03 3:
0.2sa (0.057)
0.16’ $O.CJS 9)
0.4 2fP f$.O! I ) 0.1i51n (0.03 2)
0.026 (0.028)
By comparing;these ~effickx, ranges yviththe estznates and standard errors ti Tabie 4 (fht column), several results are obtained. First, the meff?cient_:z ngi::sare not more than two standard er:roFs?way. f@m t,he Table 4 estimates, so they ati con.tained &thin ~~.~nv~n~~~~ 95% mnfideace inte~vak This meaxlrsthat the e&ma&% iinrT&k 4 @%t I~knm) are not very slt=nsitiveto the specification of the regrr;ssk~~, which is explained by the high rn~tic~~~ea~~y of the chara&ristics, &con&~it can be noted that for 1983-84 the year estimates in T~blte4 are at the lower end of the coefficient rang=. This sue;geststhat we may be underestimcl’tingthe quality-adjusted price rise in these years. Howeveq the upper ends of the coeffkient ranges in 1983~ 54 are obtained when the width and horsepower variables are on-s&d,, Since these variables are highly sigtificant when included, we shalllnot regard the upper end of the coeffGent ranges for 1983-84 as accumtz estimates. Instead, we shall continue to use the point estimates of the year dummies Wz&xi from the Table 4 regrg:ssion(first column) when estimating product quality. In Table 4 we also show the Japanese regression ,for selected pairs of years. A test for coefficient stability of the quz&y characteristics can be made by separately estimating the regression for each year, obtaining the (unrestricted) cumulative SSR of 1.052. The restricted SSR with c+F.~ coefficients in 1979-84 is i.282, as shown in the first column of T&e 4. The F statistic for testing coefficiena stability is computed as 3.74, with 30 and 102 degrees of freedom.8The 95 percent level of the F distribution is 1.60, and so we can accept the null hmothesis of equal wxxffkientsfor the characteristics in each year. Accordin@y, WC; used the 1979-84 regression 1:oestimate product quality.
In ‘E+ble5 we report the results of regressing the logarithm of US. ;;mall c%rprices (Le., subcompact and compact models) on the various ~harxtxistics for 1979-84 and selected pairs of years, In comparison wth th: Japanese regressions the fits are not as good, a -iich is seen especially from the R2 values for the two-year regressions. The length ariable now has a si~nifkantly positive coefficient, wMe weight has a 7Eegative coefficient (signifkant only at the 90% 1~~1). The width qvariable was omitted from the final regressions since it was highly -_-
-CheF statistic is cmnpuu:d as (( 1.282- I .052)/30]/( I .052/ 102) = 0.74, where six quality ’ f-a?ICtCflitlCSappcaf in 1lis rcgressson of e2c.1 ye3.r gi\‘iQ 5 X6 - :3 r~strictaoas in the 979-84 rqpsiclrr. The yearly regressions have a total of 6X 7 = 42 coefficients, giving 44- 42 = 1rS2dq:rcxs o Ffreedom for the unrestricted SSR
Length (feet) Width (feet) Weight (tons) Horsepower (100 HP)
163 0.879 0.958 7.11cJ (0.13) c).o4w
(0.013) -0.13 (0.077) Q.14a (0.062)
46 0.618 ti.300 a.39a (0.24) 0.025 (0.024) -0.11 (0.15) 0.013 (0.138)
50 0.652 0.239 7.94a (0.2 1) 0.06 la (0.0 19) -0.21 (0.12) 0.10 (0.16)
67 0.859 0.266 8.W (0.2 2) O.ou178 (0.023) -_ 1).3W , (0‘ I 4) 0.12 (OX70)
-E
-b
0.1 la (Of49) -
0.05 1 (0.040) o.12a y251
O.lW (.04 5) o.r:@
Dummy Variabks:
Transmission (Five-speed or auto) Power steering Power brakes Air conditioning Year 1980 Year 1981 Year 1982 Ytxu 1983 Year 19f,4
0.14= (0.050) 0.11= (0.0221 0.091= (0.017) OS laJ (6.060) 0.1 98a (0.024) 0.35w (0.025) 0.408’ (0.025) 0.38s’ (0.026) 0.417O (0.026)
__A
(O‘ot7) 0.08 I”
(0.021) 0.4F (0.055)
0.1 92a (0.027)
0.06 la
(0.022)
%ignificant at 95 percent level. Standard errors are in parenthcscs. bIndicates that no model had this variable as standsrd equipmen&. CThe Cadillac Cimarron was the only base model with air conditioning.
0.032 (0.01;“)
5 (first c&~mn). Htfallows that the 19’79-84 estin ates in Table 5 are not sensitive to the specification of the regression. VV’IMXI the &m&an and Japanese data are portled over 5979-84, the
b~gth crxffitient drops to 0.001.6 (sigrticant at t:le 90% level) white ihe width cxx$ficientis highly insignificant. In the pocled regression we %ave “qua1 cot#ficients for all quality characteristics, but allow Laurent year dummies for American and Japanese models. The SSR of the pooled regression is 3.45, as compared with 2.24 from adding the SSR or’the sepal-ate natilonal regressions. The s .tatistic f-Jlrtesting whether the Ameriw n and Japanese regressior, irave equrzl coefkients for the quality characteristics is computed as 18.8, with 8 and 179 degrees of freec!om? This compares with a 95 percent levt:lfor the F distrillution of 1.94, so we sxuxlly reject the hypothesis of equal coefficients for‘ the quality sharacteristics across tht: national regressions. In Table 5 we also report the American hedonic regressions for s3lectecl pairs of years. A test for coefficient stability of the quality ci~aractxistics is performed by separately estimating the regression for each year, obtaining the (unrestricted) cumulative SSR of 0.697. The restricted SSR with equal coefficients for the quality characteristics in 1979-84 is 0.958, as shown in the first column of Table 5. The Ii statistic for testing coefficient stability is computed as 1.87, with 25 and 125 degrees !offreedom.loThe 95 percent level of the F distribution is 1.619 so we initi;tlly reject the hypothesis of equal coefkients for the quality characteristics in each year. ‘:Toinvestigate stability rnoFeclosely, we performed an F test for equal coefficients of the characteristvzs ix1!:a& pair of years, with the folio King resdls: YMF!i
k smistic
i 54
1979-80 1980-8 1
3. i!5
1981-82 1982-83 1983-84
I,39 O.r26 0.82
DegHes Qf freedem
4, 4, 5, 5, 7,
36 34 38 46 Sl
--
“The 1: statistic is computed as [(3.45--2.24 B//(2.24/279) = lk!.S. where the coefficients of eil:;lt quality characteristics are restricted to be equal in the pxjlcd regression. The separy te American and Japanese regPt’ssic.ns have a toLli of 307 coefficients, giving, 307-- ‘8 = 279 degrees of freedom. “The: k’ statistic is computed as IO.958-0.C 97)/25 i/(0.697/ I 2.5) = 1.87. The number of quaht! characteristics appearing in the yearI.. regressions difkrs. hecause in some yenrs IN) model has a particuiar charactensttc such as air-c~~diti~n~g. ‘ik total number of estimxed coeffkients in the vearly regressions is 38, while the 1979-84 regression has i 3, giving 38- 13 = 25 restrictiorls and It63-38 = 125 degress of freedom for the unrestricted SS R.
Tht: ortly pair of years in which we reject the hypothesis of equal coefficents is I980-8 IIs where the F statistic of 3,lS exceeds the 95 per-t level of the P ~tr~bu~o~ given by 2,165.In alt other pairs of years, we ~CXX@the null hypothes~ of equal =Ecients for the ~hara~~~~~~~. In view of these results we decided to use the 1979-84 ~gr~ssio~, shown in the fist culDmn of Table 5, to compute product qua&v. One explanatron for the unstabie hedonic regression over 1980-81 may be the consumer response to gasoline price changes. Ohta and Griliches (I 983) examine in detail the stability of hedonic regrcssi3n 3 for U.S. used autos during 1970-81, and they find that the regression eoeficients chang:: between April and October 1979 if gasoline czsts are not taken into account, They also propose a correction for the change in gasoline prices, so that the estimated regression is stable.’ ’ When examitling eoeficient stab%ity,Ohta and Griliches use both the conventional F test and a weaker Bayesian criterion due to _tiamer (1978). Learner’s test is derived on the assumption of diffuse p&M, and the criticai level for the F statistic is therLcomputed as LB = (M- k)(rrGin - 1)/q. In this expression k is the number of narameters in the unrestricted regression, q is the number of parameters restricted by the null hypothesis, and n is the number of observations. For testing coefficient stability of the quality characteristics over W9--84, the critical level LB = 5.92, as compared with the observed F staitisticsf 1.87 reported above. ‘* Thus, according to Lea.mer’stest AXcan acei>pt the hypothesis of equal coefficients for the characteristics in each year. This reinforces our decision to use the 1979-84 regression to compute product quality. The method ot calculating quaiity is the same as that s reported above for the Japanese models, 6. PRICE AND QUALITY INDEXES The price and quality indexes reported in Tables 2 and 3 are computed with constant weiglits between each pair of years, Specifically, between each two years the Laspeyres and Paasche price indexes were computed, where the former uses the first-period quantities as weights aSIdthe latter e Fisher Ideal inderl wib;i t uses the second-period quantities.
variable in the present study.
66
Robert Feenstra
computed a:; the geometric mean (square root of the prodraet).of the
I_aspeyres and Paasche indexes, This method is recomtimd~ in Dim& (197d).13 The same calmlation wasparf$med formodel qualities, where these were obtained as the predicted price froti the hedonic regressions ttot including the contribution of that year’s dummy variable (see Siection5.1). The Fisher Ideal indexes for price and quality wL;re separately computed for Japanese and American models, as s=eportedin Tables 2 and 3. When new models are introdmed, as occurred in our sacanple,the cakulat.ion of the prim and quality indexes is affected. Since the Paasche imkx uses second-period quantities as weights, we need tc include a hypothetical price arid quality ftsr the year before the model first appears. This probletn also arisezi;,with the Laspeyres index when a model is drogpedr in tGs case the ca!&ation reqkes a hypothetical price and quality for the year afier the model last appeared. Fortunately, our use of hedonic regressions permits a natural solution to this “new goods” problem. In any year whcQ a model is not available, we can prckt its hypothetica! price and quality from the hedonic regression by ustitg its actual characteristics in the next, or previous, year (depending w-- *Nhethera model is being added or dropped). Of course, in a year when a model is not available its quantity sold is zero. Ikis methodology allows us to construct the price and quality indexes while the range of models offered changes. According to this approach, a new model would show a fall i2 its price the first year it is available if the actual price charged is below that predicted by the hedonic regression. This is because the predicted price is used as its hypothetical price in the year before it w;as first available. This price fall would affect the value of the Paasche price index. Sirce the quality of a model is always computed as is predicted price from the hedonic regressions (not including the contt’3ution af that year’s dummy variable), a new model will always Aow a constant quality between the year before and rst year it is available. Even though the quality is constant, :the existence of this new nodell will still affect Lhevalue of the Paasche ql.!ality index, by having a +ght tendencyto make the quality index itself constant. The s!.atements me have made for a Paasche index apply to the Laspeyres index when a ~-NM is dropped.
’ ?It can be noted thal: ‘.x Laspeyres md Paasche indexe% II-Iour sampie differed by less thaiI 1 pC.CeRt fOof _ach pair of giears. so fm-ning the Fisher Ideal index was a miror -idifi_C%t;!>jJ*
The YER has imposed several costs on U.S. consumers. First, the substantial depreciation of the yen sine 1981 has not been passed through in lower imported auto prices, as would be expected without the trade restraint. This represents a loss to all purchasers of Japanese or competing domestic models. In addition, there has been a significant quality upgrading of import models, amounting to $300 per import each year. This upgrading may benefit consumers who would purchase a luxury import in any case, since they now have more models to choose from, but harms those who desire the basic models. For U.S. small cars, we found a 9.1 percent increase in the suggested retail prices from 198IIto 1984, or 3 percent per year. While this increase is moderate, we must also recognize that U.S. small car priors rose rapidly just prior to the VER, and that these high prices have been maintained. The recent profit surge of U.S. producers should be attributed to volume increases and various cost reductions. While the cost reductions indicate that some adjustment to potential r:ompettion has taken place within the US. industry, consumers have not yet benefited from this adjustment. The VER has led to less prize competition in the U.S. market, enabling domestic companies to earn record profits from the volume incre.ases and cost reduction rather than lowering prices. Under these conditions, eliminating the VER would lead to substantial gains to U.S. consumers, through reduced prices for Japanese and American autos. The low value or”the yen relative to the dollar XrJeuldbe quickly reflected in reduced import prices. The quality upgrading of import models during the trade restraint may also be reversed, though this would occur more slowly. The current profits in the U.S. industry indicate that increased imports need not lead to serious disruption at home: domestic auto prices can be reduced somewhat while still maintaining profitability. The VE R with Japan is now scheduled to expire in March 1985, and it would be to the benefit of American consumers not to extend it beyoncl that date.
68
Robert Feenstra
,
-”
.%,
We examine tireeffectsof the U.S.-Japan trade restraint on automobile prices and quality u@grading,for both Japanese imports and American small Gars.From Aprir 1981 to April 1984 the suggested retail price of aCJapanese models mcreased by 15.8 percent, or ,“1,3percent per year. Ilk f&xi that nearly the entire amount of this rke can be expiained by the upgntding of individual models. This upgrading may benefit consumers who would purchase a luxury import k~any case, but harn~ those who desire the basic imports. In a.ddkiun to upgradings we conclude that a seconli cost of the trade restraint has t>rznto pr~ent the yen depreciation from being passed onto American consumers, in term of lower imported auto prices. For U.S. small cars we fmd a 9.1 percent rise in the suggested retail prioe, or 3 percent per year, with a frmctionof this amunt due to model upgrading. If the yen depreciation had led to lower inqort prices without the trade restraint, then we expect that U.S. auto prices would have been lower, too.
A prominent feature of recent U.S. trade policy has been the restraint on automobile imports from Japan. Beginning in April 19815;this “vo:luntsry export restraint” (VER) has limited Japanese auto sales to 1.68 million annually. The restraint was supposed to expire in March 1984, but instead, under U.S. pressure, the wiling was raised to 1.85 million and extended for another year. This trade arrangement is a particularly important example of the “new protectionism” since the auto industry was initially denied tariff protection under the esapeclause provision of the General Agreement on Tariffs and Trade (GATT). Thus, the rules for international safeguards embodied in the GATT were bypassed in favor of the VEIL The effect of the trade restraint on import and domestic auto prices has been the subject of recent debate in the U.S. A study by Wharton
Recoiled September 1984; acwpted November I 984,.
&onometrics (1983j zpors that from 1980 to 1982 the rise in average new car selling price; was $2,600, and that $I,OOOof this amount was due to the VER..” on the other hand, U.S. automotive pFU&MBrS &ve pc~ir&ti opJlf that the consumer I~F& index of t& &XFMB of lliabor !%&s~~cs (BL!$) has increased by an annuiaxaverage of 4.3 percent ~OF autos over 1.981-84, as compared with 4.7 percent for all ikms.* It should be noted that t$reBLS price index corrects for quality changes in autc)mobiks, such as the addition sf greater size, improved engine features!,etc. Thus, an upgratig of car models and a correspl>nt;iingrise in price will n;;t be seflected in that k&x. This is a very signii‘rcant eorrectkbn since the trade restraint has led to an upgrading of Japanese models, as observed by Feenstra (1984) during its first year? IIn this paper we shall exanke the effect of the trade restraint on automobile prices and quality upgrading, for both Japanese imports and Arnerics:2 small cars,. Summarizing our results, from April 198 P to April 19384the sugg&& ret:d p; ices of alI Japanese models increased by 15.8 percent, or 5.3 percent rer jrear. We fmd that nearly the entire amount of thi.s rise can be explained by the upgrading of individual models. In ciolllarterms, import prices have risen due to upgrading by some $300 each year, #+ving a $KKI rise by 1984. This upgrading may benefit consumers who would purchase a luxury import in any case, since they now have mze models to choose from, but harms consumers who desire the basic imports. In addition to the quality upgrading, there has been a second consumer cost of the VER When the trade restraint began, in April 198 1, the Japanese yen was at one of its strongest levels in recent history. Immediately thereafter, the yen started its well-known depreciation against the dollar. This exchange rate movement would normally
Iead to lower p&es for Japanese imports, but because of the trade restraint, was not refIeci:ed in reduced import prices for autos. We
IWharton Econ ometric Forecastinp Associates (1.983). In Wharton Motor Vehic e Service (January 1983) Specr’nlAnaZy~&~Die Japanese QUO&,the ptic: rise due to thl: VER is reported more precisely as $85 1 for imported autos over the period 1980 to 1382, and $324 for domestic actos. ?kepsc are seasondy unadjustlzd f;sures, average from March 198 1 to March 1WM. The BLS producer price .x+ Yfor autos: rose by an annual average of 4.1 p xent over the same period. *he quality upgrading can be car idered a vclns~~~~enc*~ of the $!1:Wsince it applies to the of UEOS @mted. if mstzsd the trade restraint applied to tke v&e of imports, as with a Mff, then we would not ex~xct producers to shift tow&s the higker price, higher quality molclels. mm&r
conclude that a second cost of thz VER has been to prevent the yen depreciation from bein;;;passed on to American consumers, in terms of 143WW auto prkes. TurrtEne, to U.S. autos, we investigatethz price and quality upgrading of subornpact and act mod&. From April 198f to April 1984, we fbd a 9.1 petcent tie in the suggested retail price, or 3 percent per vear. A f!ration of this amout is due to model upgrading. These figure; can be interpreted in two ways. On one hand, if the yen depreciation had led to tower import prices without the VER, then we expect that U.S. aut:o prices would have been lower, too.4 This point is underscored by noting that b the two years just prior to the VER, the domestic auto industry experienced some:of its largest price increases ever-22 percent and 17 percent, respectively, for U.S. small cars over 1979-80 and 19868 I. Thus, Hhife the recent increase in U.S. small car prices has been moderate ;tt 3% per year, we should recognize that: the histancally high prices of 1981 have been maintained. Without the protection provided by the trade restraint, in?port and domestic auto prices would have been lower. On the other hand, Gnce the actual rise in U.S. au:0 prices has been moderate, it follows that the recent surge in profits of domestic automakers cannot be Ij.ttrjbuted to price increases, but instead must be explained by volume increases and various cost reductions. Since the VER had led to less price Lompetition in the U.S. market, domestic companies have been able to earn record profits from the volume increases and cost reductions rather than lowering prices for consumers. We conclude by arguing that a greater inflow of .lapanzse autos should be permitted when the trade restraint expires in March 1985. This would
ensure that the cost reductions arPd adjustment in the U.S. industry would be passed on to consumers, through expznded trade and lower auto prices. 2, RECENT EXPERIENCE
IN THE USI, AUTO INDUSTRY
On May 1, 19815 the Japanese govr:rnm’ent announced J three-year system of ‘%oIuntary export restraints” (VER) on the export of passenger cars to the U.S. market. Fw- the period from .A March 1982 these exports would not exceed 11.68 ~~~~~~~, sceond year (April I982 to MarGh 19 raised according to the growth in thcb
‘?‘he impact of the VER on U.S. domestic car price Crandall (1984).
is zxnmincd in gn:ater detail in
These-acGons were made against a backgrounti of falEng production and high unemployment in U.S. autos, along with several legislative attempts to curb imlxwts. For eXUnple5b early 1951 Senator%Danforth and Ben&en.introduced : A.! (S.3196)to impose quotas on the import of automoM from 3apan of 1,6 million units during 198143. Indeed, fitis bill,‘was schedulledfor markup (line by line revision) in the Senate P’itlanceCommittee on Map 12, and no doubt cm&but& ts thtsspecific &or1 taken by the Japanese.Other bills included more stktigent import ~ucit;~~s, and domestic content requirements that specify the minimum c~plt tent of American parts for autos sold in the U.S. Supported by the UAW, the domestic content bill passed the House of Representatives in 1982 and 1983 (HlL5133, 1982 and HR.1234, 1983), and has been introduced into the US. Senate iin 1984 (S.707). An ear&r legislative action was thy petition for import r&f by the UAW in June 1980, made to the U.S. Enternational Trade Cor~nission @K). In August 1980 the Commission rweived a petition for sixilas import relief from E;ord Following rules established in the GATT, a x-wmmendation fbr relief ean be given only if the ‘5ncreased imports of an zii& are a substantial cause of se:&us injury, or threat thereof, to the domestic indwtry.” The U.S. tra& Pawdefines the tx n “substantial cause” as “a cause which is imprtrant and not less than any other clause.” The ITC cletern%nedthat, while imports of autos into the U.S. had increased aad the domestic ind stry was in fact injured, the RX~sion k the US. WaS a greater 633 of injury than the increased 5 A~~~~r~~g~~,import relief was not given. It was also found that t In ~~~~~~~~ preferences t~~~~d~ §~~a~,fuel efficient autos (due In part to rklng g;W3”inne s) was an *nt tfian tfie recess~unary i
and
A U T O M O B I L E PRICES A N D P R O T E C T I O N
53
I ~
~o ~
r,O
•~ ' ° _ o " ~ ~ ~Z ~ Z
~
~c~ -~
c~ .~
-~
~
~
~
~
~,~
L~
]InTable 1 we present backgrouuzd data for the awto industry during the perioct to which the VER applies. In the per&d from Apil1980 ta March 2981, tie year immediately ~rec&ing the VEiR,U.& lpraductiun a& emp~syz~a wm at dqressed leveis while the Japanese market share of 21.2 per-t was a historic high. From the second ruw we see the US. p&u&ion has picke~Iup in the most recznt year, wKlti U.S. consum~tiu~ began increasing in the period frownApril 1982 to March 1983. The rise in total consumption has led ts a drop in t2’reJapanese market share. Employment has also increased, though this occurs more slowly than tSrerise in production: as shown in tiJe Anal row, the number of cars produced per worker has increased from 8.1 to 10.2 recently. The rise in the average product of labor is in part a normal occurrence during rhe upswmg of a business cycle, and also reflects new invertment and production trdtniques in the auto industry. Ib determine the impact uf the VER on prices, a sample of Japanese i~r:portand American small (i.e., subcompact and compact) models was ta8t;n from AactomutivePVms Market Dutct.Books for 1979-84. The sample covers the base yrersion (i.e., without options) of every model a~dailablein each year, with the exception of “upility” vehi&s and station -w:Agom?In additisl: to the suggested retail price in March or April of tiz!z year, data were obtained on the quantity sold and model length, width, &eight, hoisepwer, type of transmission, and whether the base version had power steering, power brakes, or air coaditiir,dng. In Section 5 HKshall use these characteristics to estimate the\quality of c.ar models. In 1:heMowing sections we shall report our resust; ,Cor.?apanc:se imports and hmefkan smaII e..A.os,respectively. Before txning to 3ur results, ‘weshould note several qualifications. By using suggested retati prices we are ignoring additional deaIer markups fat& QI consumers, x&ich hav: been substantial in recent years. To this extent oilr price data gives an ul;4erelitimate of the consumer cost of the X3%. In addition, we are ignoring model options which a not provided as standard equipmerlt but which a cOnsumer must purchase to obtain a car. Anccdotd evidence suggests thar: such “‘required options’” have been ~~~q~~~~~y [ised in the plarchase of some recent: Japanese models. Data on thi; practice arc wt available, l-wwever, and WCconsider only the uS;i~ qgrading t.h;lt fras occurred on the base versions (i.e., without options) OF-eacl” model.
AUTOMOBILE PRICES AND PWEECTIQN
_ ~_
_ ___ __ _
Unit va\ue (% change) price index (% chage) Number of models New models unit quzklity (% change) Quality index (% change)
-
55
_~~
1979
1980
1981 -
1982
1983
1984
4,946 80.8 21 5,077 92.J -
5,175 (4.6) 83.5 (3.3) 24 3 5,195 (2.3) 93.5 (1.3)
6,211 (20.0) loo.0 (19.8)
6,834 (10.0) 107.8 (7.8) 24 ob 5,942 (6.8) 105.1 (5.1)
7,069 (3.4) 109.5 (1.6) 26 4b 6,316 (6.3) 109.7 (4-4)
7,459 (5.5) 115.8 (5.8) 25 0 G,S56 (3.8) 114.0 (3.9)
24 0 5,564 (7.1) 100.0 (7.0)
uObtain& front Qutomotive Kews, Market Data Book Issue, 1979-84 years. bin 1982-83 the Datsun Sentra, Pulsar, and Stanza replaced the 210, 3 10, and 5 10, respectively, and are not counted as new models.
3. PRICE AND QUALITY OF JAPANESE IMPORTS
In Table 2 we show information for Japanese imports over 1979-84. The unit values shown in the first row are a weighted average cf suggested retail prices of base models, using the quantities sold as weights. The prices for 198 1 were those in effect on April 8, just prior to the annoutcement of the export restraint. Accordingly, we shall treat 193 1 as our ‘&baseyear” for comparisons. The unit values show a rise of 20 percent, from $6,211 in 1981 to $7,459 in 1984. However, this rise includes both the effect of increasing prices for individual models and the shift in quantity demanded towards higher-priced models. A more appropriate measure of the overall price increase is to construct a price index using constant weights between each two years (see Section 6). Reported in the second row, this price index shows an incrl:ase of 15.8 percent in suggested retaiJ prices between 198 1 and 1984,of 5.3 percent per year. The variety of Japanese models o;fered was quite stable over the p~‘r~o~$as shown by the number of existing and new rnodeIs ez\ch year. irA the nljsa7ber of WI&z there was little cha qualities of individual mode
rutted using constant ~~~gh~s irK.kxshalws 23tise sf B4 percent over
retail
prices.Thus, nearlyall of the rise
u&f
in
dmparr prices cm be
explained by t/w upgrading of indtidual modelk. Thr: dollar krease in quality can be computed through multiplying
the change in t5: qu&y index times $6,X 1, the 1981 base price. This gives a quality 15:,eof about $300 per import each year, leading to a $900 increa:;e by f 984 The quality upgrading has a consumer ccst that varks ac-:3ss individua’s. The upgrading may benefit consumers who would purchase a luxury import in any case, since they now have more models to choose from, ~utiharms those who desire the basic madeis. 11is prhaps s~urpriskg that most of the price rise in imports can be attriiut&! to qustity upgrading. We would normally expect the trade restraint to havt: a further upward effecd;on prices, independent of +4&y, by creating an &fkial shortage. We shall nc-1 argue that the VEIf .aas had wch an upward effect on prices, by prevtllting movements irkGre df E;ar/ycn exchange rate from being passed through in import prkc;. 111Figure f we show the quarterly dollar/yen exchange rate over the ast sez3-al ve88:3(o&be to ~~~~~~~) = KIO). In the firsi quarter of 98 9 the y’ n was at me of it s stron:gest bevels in recent history, having a p~~eaated 18 percent from 8he p vioamsyem. ~~d~~d~ this s.pprec;ation ~~~~~~~~~~~very closely f831 ttle z percent ~-kc:in Japanese auto prices fran 1980 to I!318 li (see Tabie 2). ~rnme~at~~yafter the VER was in
AUTOMOBHE PRICES AND PROTECTION
57
place the yen began its depreciatios; against the dollar,,due in part to hligh U.S. interest rates. Relative to the first quarter of 198 19 his depreciatiun eweds 1Q percent fur any of the klllowing fiist-quarter averages, and the @eatest depreciation of 26 per cent is obtained by comparing fmtquarter %981to fourth_quarW 19:U:. Based an this exchange rate behavior, we expect that withcxt the VER Japanese au& prices would have J%%IZsince 198X. This predicticm is reinforced by notm~ :!I& in the year just &fore the VER.,the 1.8percent appreciation of the yen correspond& $0 a nearly equal rise in imported auto prices. It is lcllgicalto expect thz *; the reversal k the yen movement would have led to SWIMS fall k irnp&ed auto prices. However, in the presence of the arMicia1shortage created by the trae restraint, no such deck in auto prims has occurred. We conclude that Qfir&a= impact qf fh43imule restraint has been to prevent t?8eyen depreciutionfrom &ing passed on to American cons’umers,througJtlower imported auto prices., Ilkis consumer cost is in addition to the quality upgrading of impart nxdels tkzt has occurred. 4. PRICE ANR QUALITY OF AMEIWZAN SMALL AUTOS In Table 3 we report the price and quality dzita fcr U.S. subcompact and compact modkls. The unit value, constructed as a weighted average of suggested retail prides of base models, shows a rise CC14.6 percent from 198 I to 1984, However, a considerable portion of fhk rise is due to a demand shift t\(sa7vards larger models. The index of suggested retail
T&k 3: Sample of American §mall .QutoP -Unit ~slue (96 change) Price :ndex (% change) Number of models New models Wnit quality (% change) Quaiity index (% change)
1’97!#
198@
4:186 70.1 24 -
5,067 (21.0) 85.6 (22.1) 22
4,l?5
4,132 (- 93) 47.8 (-0.5)
98.3 -
!
I9tJ .1 --I5,9 f :5 (16 7) 1OG.O
(16.8) 23 7 4,183 (11.2) 18fr.i) (2 2)
1982
1983
X984
6,446 (9.W 107.5 (7.5) 27 4 4,35 I
6,58 1 (2.1) 107.3 (-C.2) 33 4% 4,497 (3.4) lW.4 fi.3)
6,73 I (3.0) 109.1 (11*?b 34 7 4,563 (KS) Ii03.8 (C.4)
(4.0) 102.11 (2. I)
aObtained from Au&mmtive News, Market Data Book Issue, 11979-84 years. bin 1983 the Dodge 400 and Chrysler LeTZaronwere down-sized from %rlier models. and are not counted as laew.
prices,construtied using wnstant weights between irmwses by only 3.1 permnt from 198
I tro
aa& two years, 1984, or 3 percent per year.
This is less than the rise in the overall wnsumtir price tidex ova that petiod (see Se&m 1). We also see that there itrasbeen m active introduction of ne+~models and expiring uf old, Some qutity upgrading of US, models has occurred, though less than the amount of the price rise. Imhe foal rovvof Tab133 we report an increaseof 3.8 percentin the quality index from 1981 tcb1984, which is slightly less than one-half of the rise in the price index. It is also of importance that in the two years just before the trade restraint begqn, U.S. small car prim rose very substantially--Z.!. percent and 16.8 percent, res?ctively, over 1979-80 and 1980-81. This compares with 8 rise in &.Iwnsumer prices of 13.5 percent and 10.4, respectively, over the same periods. Thus, whilethe increase in U.S. small car prices sincl? 1981 has been moderate at 3 percent per year, we can a&natively %e rpret the tidence as saying that the historically high p:icpc of 1981 hwle simply been maintained. This view corresponds with our discussion of the do&¥ exchange rate and import prices. In the absence of the VER, we e,rpct that the yen depreciation would have led to lower import prices. Tlris would have created greater price competition on ‘domesticproducelbs,and some lowering of U.S. auto prices, too. Instead, the trade restrtlin: has insulated the U.S. industry from movements n tie dollar/yen exchange rate, thereby permitting fhe high prices of 1981 to be maintained. However, we should also recognize that the surge in profits of U.S. autcmakc;z cannot be attributed to recent price increases for small cars, swx lhlese increases have beep moderate. Evidence on the extent of the ~;&I: turnaround is 2s folLyas:
Somx:
Autmwtzve
News, various
issues.
earned by con~p~anies during the first quarter of 1984 have broken all previous records, and even exczeded the annual profits earned in many years. The profit surge of U.S. automakers can be explained by a variety of factors, The volume expansion as the demand f’orcars has increased (see Table 1) will lower average costs due to economies of scale, and raise profits. Direct cost reductions have also been achieved as fi=wrer worke?*s are rehired coming out of the recession, and the average product of labor increases (Table 1). U.S. firms have reported that the;break-even volume for profitability has been reduced in each company, reflecting a decrease in fixed costs. In addition, a numlber of recent U.S. models have proved to be very popular among consumers. Lastly, it shouid be noted that a portion of U.S. automakers’ profits are obtained from truck sales, which have been protected by a 25 percent tariff since 1980. While the cost reductions in the U.S. industry indicate that some adjustment to potential competition has taken place, consumers ha;re not yet benefited from this adjustment. The trade restraint has led to 1~ price competition in the 1J.S. market, enabling domestic companies :o earn record profits from the volume increases and cost reductions rather than lclwering prices. As compared with a tariff, the VER gives greater “market power” to domestic producers since the supply of Japanese imports is constrained in quantity? The recent experience in U.S. autos appears to be an important example of the exercise of market power under a quantitative trade restraint, with U.S. firms increasing their profits in response to higher demand and reduced costs, but not lowering consum prices. 5. ESTIMATION OF PRODUCT QUALITY
For each of the Japanese an4 U.S. small car models, data were collectec. on the suggest%! retail price of base versions (i.e.. ~~~~~~~~ options)., quantit:l sold, and various characteristics. wi3: be used as ;1 measure of qu !ity upgrading, the st~t~st~~a~ technique of “hedonic” xgressions. The basic rtzferenct: cm 0-k tee is ~~~~~~~~~ (B 87
I ).
t979-84 q&w an? able lo eq&&i 92 lpercentQf Is%@. %@J%it~mfi :ilhk-ptim @*=8.92$ ti the fiat c0lu.m). The ~~~~t~ Q~~~~.~~~~~~~~~~~~ can be give a preciseinteqsetatiox the coeB%kntof 0.40 fur width~~fsr e3t8nqk, indkstes that an k3xxsi2 in width af one Rx%would r&e the SW& prioeby 40 perc&nt.The transniission va%db~~t&m #ha v&e of unity ii’ the ma&l hairs a Rve-speed or autamtic trz&stissian as stan&rd qyi~ment, atid zero otherwise;the cc#f%i&xtof 0.15 indiu& &at “&isfeetursrlx&MS the 43stillaf;e$ priceby 15pfxcernt.TIM pmna af air ~sc&itktiqg 3s standard equipment also Ssed the &knz&d price by 15 per=nt. J[nthe iFinalregression “tiehave omitted the wei@ and power brake varia!&s since these estrmated co#kients when included were highly insignificant (with star&& errors nearly twice the coeffkient size). Omitting these variables has only a shght efkct on the t:Gser coet&ients (see bhwj. It canbenoticed that tke length variable has a negative and signif cant coefficient, which is unexpected This may be due to mt&icoUinearity with other &aracteristics or misspecifk&on of the equation. Aside iTom the negative sign on length the 197%-84 regression appears to fit ri.ther weX We have included dumy variables for the various years, equ, tling Gty in that year and zero otherwise. Except for 1980 the estirr ated coeffxcientsof the year dummies are highly significant. The estimale of 0.15 1 in 9984, for example, indicates that 15,l percent of the [unweighted) price rise from 1979 to 1984 is not explained by any chaqt;e in cbaracteristrcs. Y’hus,the yeai- coefficients t;an be interpreted as the qrlahty-adjusted price rke. A measure of quality is obtained by computing the predicted price from the hedonic regression not including the portion explained by the year dunmzks. This calcuktion is rewated for each year, obtaining a mt;3sure of moddquality for each model. The unit quality reported in the se=n.! to the last row of ‘Bble 2 is a weighted average ~“VCSS models, using t, \aquantities sold that year as weights (for 1984 WCME the ! 983 quantities as weights). The quality index reported in the last TQW of Table 2 is computed with constant weights between each two years, as dessrikd in Se&on 6. t is clear that our measure of quaky is sensitive to the estimated coeffitients nf the yeas dummies, since t se coefficients measure the ~~a~ty-~dj~~.;ted ~xice rise, We checked e sensitivity of these coeifficients TV)the mo&4 sp3cifkation hy reestimating the f979- 84 regresskn, wme omitting each time on: of the fokving variables: length, th, weight, and horsepower. ?Selow report the range of point ates k’i,r* Ihe year co these regressions:
1981 I982 1983 1984
0.105-0.120 0.139-0.164 0.128-D. 153 O*i51-0.193
Table 4: hpanese Hedonic Regressions 1939-84 &per&m
1979-80
1981-82 _.
198344
Variable: Japanese
Au& Price (Logarithm) Observations
R2 SSR Intercept Length (feet) Width (feet) Weight (tons) Horsepower (100 HP)
Bummy Variables: Transmission (Five-speed or auto) Power steering Power brakes Air conditioning Year 1980 Year 1981 Year 1982 Year I.983
Year 1984
144 0.924 1.282 6.41° (0.48) - o.045a (0.019) 0.4w (0.11) -
45 0.885 0.425 4.95a (8.85) -0.059 (0.036) 0,71Q (0.2 1) -
48 0.937 0.290 6.83” (0.74) -r).D75’~’ (0.034) 0.42* (0.17) -
51 0.929 0.419 8.28C (0.99) -0.059 (0.039) O.lD (0.20) -
0.77a (0.066)
0.76O (0.15)
o.74a (0.11)
0.79a (0.11)
o.15a (0.023) 0.052 (0.029) -
0.1 la (0.040) 0.12 (0.087) -
8.18O (0.03 5) 0.16 (0.050) -
0.23a (0.062) 0.1053 (0.04 1) _-
O.lSa (0.03 5) Q.0084 (0.030) 0.115” (0.030) 0.15ou (0.03 1)
0.072 (0.098) ODD59 (0.03 3:
0.2sa (0.057)
0.16’ $O.CJS 9)
0.4 2fP f$.O! I ) 0.1i51n (0.03 2)
0.026 (0.028)
By comparing;these ~effickx, ranges yviththe estznates and standard errors ti Tabie 4 (fht column), several results are obtained. First, the meff?cient_:z ngi::sare not more than two standard er:roFs?way. f@m t,he Table 4 estimates, so they ati con.tained &thin ~~.~nv~n~~~~ 95% mnfideace inte~vak This meaxlrsthat the e&ma&% iinrT&k 4 @%t I~knm) are not very slt=nsitiveto the specification of the regrr;ssk~~, which is explained by the high rn~tic~~~ea~~y of the chara&ristics, &con&~it can be noted that for 1983-84 the year estimates in T~blte4 are at the lower end of the coefficient rang=. This sue;geststhat we may be underestimcl’tingthe quality-adjusted price rise in these years. Howeveq the upper ends of the coeffkient ranges in 1983~ 54 are obtained when the width and horsepower variables are on-s&d,, Since these variables are highly sigtificant when included, we shalllnot regard the upper end of the coeffGent ranges for 1983-84 as accumtz estimates. Instead, we shall continue to use the point estimates of the year dummies Wz&xi from the Table 4 regrg:ssion(first column) when estimating product quality. In Table 4 we also show the Japanese regression ,for selected pairs of years. A test for coefficient stability of the quz&y characteristics can be made by separately estimating the regression for each year, obtaining the (unrestricted) cumulative SSR of 1.052. The restricted SSR with c+F.~ coefficients in 1979-84 is i.282, as shown in the first column of T&e 4. The F statistic for testing coefficiena stability is computed as 3.74, with 30 and 102 degrees of freedom.8The 95 percent level of the F distribution is 1.60, and so we can accept the null hmothesis of equal wxxffkientsfor the characteristics in each year. Accordin@y, WC; used the 1979-84 regression 1:oestimate product quality.
In ‘E+ble5 we report the results of regressing the logarithm of US. ;;mall c%rprices (Le., subcompact and compact models) on the various ~harxtxistics for 1979-84 and selected pairs of years, In comparison wth th: Japanese regressions the fits are not as good, a -iich is seen especially from the R2 values for the two-year regressions. The length ariable now has a si~nifkantly positive coefficient, wMe weight has a 7Eegative coefficient (signifkant only at the 90% 1~~1). The width qvariable was omitted from the final regressions since it was highly -_-
-CheF statistic is cmnpuu:d as (( 1.282- I .052)/30]/( I .052/ 102) = 0.74, where six quality ’ f-a?ICtCflitlCSappcaf in 1lis rcgressson of e2c.1 ye3.r gi\‘iQ 5 X6 - :3 r~strictaoas in the 979-84 rqpsiclrr. The yearly regressions have a total of 6X 7 = 42 coefficients, giving 44- 42 = 1rS2dq:rcxs o Ffreedom for the unrestricted SSR
Length (feet) Width (feet) Weight (tons) Horsepower (100 HP)
163 0.879 0.958 7.11cJ (0.13) c).o4w
(0.013) -0.13 (0.077) Q.14a (0.062)
46 0.618 ti.300 a.39a (0.24) 0.025 (0.024) -0.11 (0.15) 0.013 (0.138)
50 0.652 0.239 7.94a (0.2 1) 0.06 la (0.0 19) -0.21 (0.12) 0.10 (0.16)
67 0.859 0.266 8.W (0.2 2) O.ou178 (0.023) -_ 1).3W , (0‘ I 4) 0.12 (OX70)
-E
-b
0.1 la (Of49) -
0.05 1 (0.040) o.12a y251
O.lW (.04 5) o.r:@
Dummy Variabks:
Transmission (Five-speed or auto) Power steering Power brakes Air conditioning Year 1980 Year 1981 Year 1982 Ytxu 1983 Year 19f,4
0.14= (0.050) 0.11= (0.0221 0.091= (0.017) OS laJ (6.060) 0.1 98a (0.024) 0.35w (0.025) 0.408’ (0.025) 0.38s’ (0.026) 0.417O (0.026)
__A
(O‘ot7) 0.08 I”
(0.021) 0.4F (0.055)
0.1 92a (0.027)
0.06 la
(0.022)
%ignificant at 95 percent level. Standard errors are in parenthcscs. bIndicates that no model had this variable as standsrd equipmen&. CThe Cadillac Cimarron was the only base model with air conditioning.
0.032 (0.01;“)
5 (first c&~mn). Htfallows that the 19’79-84 estin ates in Table 5 are not sensitive to the specification of the regression. VV’IMXI the &m&an and Japanese data are portled over 5979-84, the
b~gth crxffitient drops to 0.001.6 (sigrticant at t:le 90% level) white ihe width cxx$ficientis highly insignificant. In the pocled regression we %ave “qua1 cot#ficients for all quality characteristics, but allow Laurent year dummies for American and Japanese models. The SSR of the pooled regression is 3.45, as compared with 2.24 from adding the SSR or’the sepal-ate natilonal regressions. The s .tatistic f-Jlrtesting whether the Ameriw n and Japanese regressior, irave equrzl coefkients for the quality characteristics is computed as 18.8, with 8 and 179 degrees of freec!om? This compares with a 95 percent levt:lfor the F distrillution of 1.94, so we sxuxlly reject the hypothesis of equal coefficients for‘ the quality sharacteristics across tht: national regressions. In Table 5 we also report the American hedonic regressions for s3lectecl pairs of years. A test for coefficient stability of the quality ci~aractxistics is performed by separately estimating the regression for each year, obtaining the (unrestricted) cumulative SSR of 0.697. The restricted SSR with equal coefficients for the quality characteristics in 1979-84 is 0.958, as shown in the first column of Table 5. The Ii statistic for testing coefficient stability is computed as 1.87, with 25 and 125 degrees !offreedom.loThe 95 percent level of the F distribution is 1.619 so we initi;tlly reject the hypothesis of equal coefkients for the quality characteristics in each year. ‘:Toinvestigate stability rnoFeclosely, we performed an F test for equal coefficients of the characteristvzs ix1!:a& pair of years, with the folio King resdls: YMF!i
k smistic
i 54
1979-80 1980-8 1
3. i!5
1981-82 1982-83 1983-84
I,39 O.r26 0.82
DegHes Qf freedem
4, 4, 5, 5, 7,
36 34 38 46 Sl
--
“The 1: statistic is computed as [(3.45--2.24 B//(2.24/279) = lk!.S. where the coefficients of eil:;lt quality characteristics are restricted to be equal in the pxjlcd regression. The separy te American and Japanese regPt’ssic.ns have a toLli of 307 coefficients, giving, 307-- ‘8 = 279 degrees of freedom. “The: k’ statistic is computed as IO.958-0.C 97)/25 i/(0.697/ I 2.5) = 1.87. The number of quaht! characteristics appearing in the yearI.. regressions difkrs. hecause in some yenrs IN) model has a particuiar charactensttc such as air-c~~diti~n~g. ‘ik total number of estimxed coeffkients in the vearly regressions is 38, while the 1979-84 regression has i 3, giving 38- 13 = 25 restrictiorls and It63-38 = 125 degress of freedom for the unrestricted SS R.
Tht: ortly pair of years in which we reject the hypothesis of equal coefficents is I980-8 IIs where the F statistic of 3,lS exceeds the 95 per-t level of the P ~tr~bu~o~ given by 2,165.In alt other pairs of years, we ~CXX@the null hypothes~ of equal =Ecients for the ~hara~~~~~~~. In view of these results we decided to use the 1979-84 ~gr~ssio~, shown in the fist culDmn of Table 5, to compute product qua&v. One explanatron for the unstabie hedonic regression over 1980-81 may be the consumer response to gasoline price changes. Ohta and Griliches (I 983) examine in detail the stability of hedonic regrcssi3n 3 for U.S. used autos during 1970-81, and they find that the regression eoeficients chang:: between April and October 1979 if gasoline czsts are not taken into account, They also propose a correction for the change in gasoline prices, so that the estimated regression is stable.’ ’ When examitling eoeficient stab%ity,Ohta and Griliches use both the conventional F test and a weaker Bayesian criterion due to _tiamer (1978). Learner’s test is derived on the assumption of diffuse p&M, and the criticai level for the F statistic is therLcomputed as LB = (M- k)(rrGin - 1)/q. In this expression k is the number of narameters in the unrestricted regression, q is the number of parameters restricted by the null hypothesis, and n is the number of observations. For testing coefficient stability of the quality characteristics over W9--84, the critical level LB = 5.92, as compared with the observed F staitisticsf 1.87 reported above. ‘* Thus, according to Lea.mer’stest AXcan acei>pt the hypothesis of equal coefficients for the characteristics in each year. This reinforces our decision to use the 1979-84 regression to compute product quality. The method ot calculating quaiity is the same as that s reported above for the Japanese models, 6. PRICE AND QUALITY INDEXES The price and quality indexes reported in Tables 2 and 3 are computed with constant weiglits between each pair of years, Specifically, between each two years the Laspeyres and Paasche price indexes were computed, where the former uses the first-period quantities as weights aSIdthe latter e Fisher Ideal inderl wib;i t uses the second-period quantities.
variable in the present study.
66
Robert Feenstra
computed a:; the geometric mean (square root of the prodraet).of the
I_aspeyres and Paasche indexes, This method is recomtimd~ in Dim& (197d).13 The same calmlation wasparf$med formodel qualities, where these were obtained as the predicted price froti the hedonic regressions ttot including the contribution of that year’s dummy variable (see Siection5.1). The Fisher Ideal indexes for price and quality wL;re separately computed for Japanese and American models, as s=eportedin Tables 2 and 3. When new models are introdmed, as occurred in our sacanple,the cakulat.ion of the prim and quality indexes is affected. Since the Paasche imkx uses second-period quantities as weights, we need tc include a hypothetical price arid quality ftsr the year before the model first appears. This probletn also arisezi;,with the Laspeyres index when a model is drogpedr in tGs case the ca!&ation reqkes a hypothetical price and quality for the year afier the model last appeared. Fortunately, our use of hedonic regressions permits a natural solution to this “new goods” problem. In any year whcQ a model is not available, we can prckt its hypothetica! price and quality from the hedonic regression by ustitg its actual characteristics in the next, or previous, year (depending w-- *Nhethera model is being added or dropped). Of course, in a year when a model is not available its quantity sold is zero. Ikis methodology allows us to construct the price and quality indexes while the range of models offered changes. According to this approach, a new model would show a fall i2 its price the first year it is available if the actual price charged is below that predicted by the hedonic regression. This is because the predicted price is used as its hypothetical price in the year before it w;as first available. This price fall would affect the value of the Paasche price index. Sirce the quality of a model is always computed as is predicted price from the hedonic regressions (not including the contt’3ution af that year’s dummy variable), a new model will always Aow a constant quality between the year before and rst year it is available. Even though the quality is constant, :the existence of this new nodell will still affect Lhevalue of the Paasche ql.!ality index, by having a +ght tendencyto make the quality index itself constant. The s!.atements me have made for a Paasche index apply to the Laspeyres index when a ~-NM is dropped.
’ ?It can be noted thal: ‘.x Laspeyres md Paasche indexe% II-Iour sampie differed by less thaiI 1 pC.CeRt fOof _ach pair of giears. so fm-ning the Fisher Ideal index was a miror -idifi_C%t;!>jJ*
The YER has imposed several costs on U.S. consumers. First, the substantial depreciation of the yen sine 1981 has not been passed through in lower imported auto prices, as would be expected without the trade restraint. This represents a loss to all purchasers of Japanese or competing domestic models. In addition, there has been a significant quality upgrading of import models, amounting to $300 per import each year. This upgrading may benefit consumers who would purchase a luxury import in any case, since they now have more models to choose from, but harms those who desire the basic models. For U.S. small cars, we found a 9.1 percent increase in the suggested retail prices from 198IIto 1984, or 3 percent per year. While this increase is moderate, we must also recognize that U.S. small car priors rose rapidly just prior to the VER, and that these high prices have been maintained. The recent profit surge of U.S. producers should be attributed to volume increases and various cost reductions. While the cost reductions indicate that some adjustment to potential r:ompettion has taken place within the US. industry, consumers have not yet benefited from this adjustment. The VER has led to less prize competition in the U.S. market, enabling domestic companies to earn record profits from the volume incre.ases and cost reduction rather than lowering prices. Under these conditions, eliminating the VER would lead to substantial gains to U.S. consumers, through reduced prices for Japanese and American autos. The low value or”the yen relative to the dollar XrJeuldbe quickly reflected in reduced import prices. The quality upgrading of import models during the trade restraint may also be reversed, though this would occur more slowly. The current profits in the U.S. industry indicate that increased imports need not lead to serious disruption at home: domestic auto prices can be reduced somewhat while still maintaining profitability. The VE R with Japan is now scheduled to expire in March 1985, and it would be to the benefit of American consumers not to extend it beyoncl that date.
68
Robert Feenstra