Australia’s coking coal exports to Japan Price–quality relationships under benchmark and fair treatment pricing Anthony Swan, Sally Thorpe and Lindsay Hogan* Australian Bureau of Agricultural and Resource Economics 42nd Annual Conference of the Australian Agricultural and Resource Economics Society University of New England, Armidale, 19–21 January 1998 ABARE CONFERENCE PAPER 98.3 1 ABARE ABARE project 1387 Given Japan’s dominant position in the regional coal market and the continuing relatively low profitability of Australia’s coal industry, the influence of the Japanese steel mills on coal pricing arrangements between Australia and Japan remains a policy issue in Australia. The Japanese steel mills introduced the ‘fair treatment’ pricing system in Japanese fiscal year (JFY) 1996, whereby coal contract information would be confidential but it was argued that coal would be priced according to its value in use. In this paper, Quandt’s switching regime model is used to test for structural change in price–quality relationships in the important Australia–Japan coking coal trade. There is statistical evidence that price–quality relationships changed fundamentally after JFY 1994 for semisoft coking coals when the soft coking coal category was merged with the semisoft coking coal category, and in JFY 1996 for hard coking coals when the fair treatment pricing system was introduced. It is concluded that coking coal price–quality relationships have become substantially less transparent in recent years. * The authors wish to thank Andrew Dickson of ABARE for useful comments.
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Australia’s coking coal exports to JapanPrice–quality relationships under
benchmark and fair treatment pricingAnthony Swan, Sally Thorpe and Lindsay Hogan*
Australian Bureau of Agricultural and Resource Economics
42nd Annual Conference of the Australian Agricultural and Resource Economics Society
University of New England, Armidale, 19–21 January 1998
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ABARE project 1387
Given Japan’s dominant position in the regional coal market and the
continuing relatively low profitability of Australia’s coal industry, the
influence of the Japanese steel mills on coal pricing arrangements between
Australia and Japan remains a policy issue in Australia. The Japanese steel
mills introduced the ‘fair treatment’ pricing system in Japanese fiscal year
(JFY) 1996, whereby coal contract information would be confidential but it
was argued that coal would be priced according to its value in use.
In this paper, Quandt’s switching regime model is used to test for structural
change in price–quality relationships in the important Australia–Japan coking
coal trade. There is statistical evidence that price–quality relationships
changed fundamentally after JFY 1994 for semisoft coking coals when the soft
coking coal category was merged with the semisoft coking coal category, and
in JFY 1996 for hard coking coals when the fair treatment pricing system was
introduced. It is concluded that coking coal price–quality relationships have
become substantially less transparent in recent years.
* The authors wish to thank Andrew Dickson of ABARE for useful comments.
IntroductionAustralia is the world’s largest coal exporter, accounting for around 29 per cent of world
coal exports in 1996, and Japan is the world’s largest coal importer, accounting for around
26 per cent of world coal exports (IEA 1997). In 1996-97, Australia’s coal exports were
valued at A$7.9 billion or 9 per cent of Australia’s total merchandise exports. Japan
remains Australia’s largest coal export market, accounting for 41 per cent and 53 per cent
of Australia’s metallurgical and thermal coal exports, respectively, in 1996-97.
Given the importance of coal in Australia’s merchandise exports, Japan’s dominant market
position and the continuing relatively low profitability of the Australian coal industry, the
efficiency of the Asia Pacific regional coal market remains a policy issue in Australia
(Hogan, Thorpe, Graham and Middleton 1997). In recent years, there have been two major
inquiries into the economic performance of Australia’s black coal industry. In the final
report of the 1994 inquiry, Taylor (1994) noted that it is difficult to assess whether the
collective buying practices of the Japanese steel mills and power utilities have resulted in
lower coal prices to this market. The recommendations in the Taylor report were directed
at increasing market transparency, increasing productivity and international competitive-
ness, and implementing an export diversification strategy (Taylor 1994). The recent inquiry
by the Industry Commission, which is to report in early 1998, has emphasised an inter-
national benchmarking approach to assess productivity and international competitiveness
issues more fully.
The focus in this paper is on the recommendations of the Taylor report relating to the need
to increase transparency in coal price determination, particularly with respect to the
influence of the Japanese steel mills on coal pricing arrangements between Australia and
Japan. From the mid-1980s, coking coal prices paid by Japanese steel mills, and broadly
followed by Asian steel mills more generally, were based on the Japanese benchmark
pricing system. Under this system, coking coals were grouped and priced by coal category,
where coal categories included hard, soft and semisoft coking coal. The absolute price of
a coal brand within a given coal category was negotiated relative to the benchmark coal
with known quality characteristics. After Japanese fiscal year (JFY) 1994, the soft coking
coal category was merged into the traditionally lower priced semisoft coking coal category.
In JFY1996, Japan replaced the benchmark pricing system for coking coals with the ‘fair
treatment’ pricing system whereby, it was argued, each coal brand would be valued
according to the quality requirements of specific Japanese steel mills. Under this
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arrangement, however, coal prices and other contract details would remain confidential.
The fair treatment system was introduced after substantial coking coal price increases in
the JFY1995 negotiations, following several years in which coking coal prices were
consistently reduced in real terms.
Issues of market inefficiency arise under both the benchmark and ‘fair treatment’ systems
because, first, coking coal categories are not uniquely defined in terms of coal quality
attributes but prices are nonoverlapping between categories and, second, the international
price negotiations are conducted sequentially by coal category and by coal importing
nation which effectively reduces the number of buyers in the market at any point in time
(Hogan, Thorpe, Graham and Middleton 1997). Confidentiality of price–quality–quantity
information under the fair treatment system is an issue since coal price discovery during
the annual negotiations, particularly for coal exporters, is made difficult. More generally,
the information content of price signals is critical for efficient resource allocation in
international coal trade.
The objective in this paper is to test for structural change in the price–quality relationships
implicit in the Australia–Japan coking coal trade between JFY1992 and JFY1997 using
Quandt’s exogenous switching regime regression technique, described in Goldfeld and
Quandt (1976) and Johnston (1991). Price–quality data are obtained from the Australian
Coal Report (various issues), including the data for JFY 1996 and 1997 which represent
estimates of actual settlements.
Recent hedonic regression analyses of hard coking coal price–quality relationships include
Chang (1995), Koerner (1996) and Hogan, Thorpe and Middleton (1997), although the
last paper also includes other coal categories. The consistency of coal price–quality
relationships by two or more major suppliers into Japan are examined in the first two
papers, and coal price–quality relationships in Australia’s exports to Japan and other
markets are examined in the third paper. Chang (1995) pools data for JFY 1993 and JFY
1994, Koerner (1996) undertakes separate regression analyses for JFY 1992 and JFY 1994,
while Hogan, Thorpe and Middleton (1997) pool data for the period JFY 1989 to JFY
1996. None of these studies explicitly test for structural changes in coal price–quality
relationships over time.
The next section contains a brief description of coal pricing arrangements and the quality
attributes of coking coals. Recent previous studies are described in the third section,
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followed by a description of the research method and data. Results are presented in the
following section and concluding comments are given in the final section.
Coal quality and pricing arrangements
Coal quality
Coal is not a homogeneous product, and is classified into broad groups — hard, soft and
semisoft coking and thermal — according to the end use and the manner in which the coal
performs in that end use. The main quality characteristics of coking coal are total moisture
Coal type Hard Hard Hard Hard Soft SemisoftEstimation period a 1993–94 1992 1994 1989–96 1989–94 1989–96No. observations 34 38 38 1255 1255 1255Adjusted R squared b 0.89 0.95 0.93 0.96 0.96 0.96
a Based on Japanese fiscal years. b Unadjusted R squared for Koerner (1996). Only one adjusted R squared is generated for thepooled regression in Hogan, Thorpe and Middleton (1997). c Insignificant denotes a variable that is not significantly different fromzero at the 5% significance level.
studies. The coal price is measured in nominal US dollars a tonne in Chang (fob) and
Koerner (cif), and in JFY 1995 prices in Hogan, Thorpe and Middleton (fob).
The four quality variables included in Chang’s regression analysis are VM, SULP, CSN
and LFLUID. With the exception of SULP, all quality parameters are statistically
significant from zero at the 5 per cent level and have the correct sign. The four quality
variables included in Koerner’s JFY 1992 regression equation are ASH, SULP, REFL and
LFLUID. With the exception of ASH, all quality parameters are statistically significant at
the 5 per cent level and have the correct sign. A similar regression analysis is also
conducted for JFY 1994 and, except for an insignificant coefficient for sulphur, the results
are similar.
Notably, Koerner does not include CSN in the regression model arguing, among other
things, that it is an alternative physical measure to LFLUID of the coking characteristic of
a coking coal. Koerner also argues that LFLUID is related to VM, which may explain why
he does not also include that variable in the regression. Unlike Chang, Koerner also
includes REFL in the regression model arguing that this variable is a measure of the
available carbon in a coking coal.
Hogan, Thorpe and Middleton (1997) have information on six quality characteristics for
Australia’s coking coal shipments, including TM, IM, VM, ASH, SULP and CSN. They
impose a linear functional form on the coking coal regression equation arguing that the
high degree of correlation between each of the quality variables and squared and cross-
product terms rules out the use of more general functional forms in ordinary least squares
regression. Notably, Hogan, Thorpe and Middleton (1997) conduct RESET tests for
general model misspecification, particularly in functional form, and find no evidence of
this. RESET tests pick up curvature in the underlying model. Hogan, Thorpe and
Middleton (1997) also undertake several tests for departures from homoskedasticity and
find no evidence of this.
Hogan, Thorpe and Middleton (1997) find that IM, ASH and CSN are significant
explanators of hard, soft and semisoft coking coal prices. SULP is also an important
explanator of soft coking coal prices, and a kinked relationship for VM, whereby the
implicit price of semisoft coking coal is positive for coals with a VM less than 30 but is
negative thereafter, is an important explanator of semisoft coking coal prices. The negative
kink is introduced through variable VM-KINK, a dummy variable relationship for VM that
is defined in section 4. Intercept dummies for soft and semisoft coking coals are also found
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to be significant. This suggests that the categorisation of coal by type is a key explanator
of coking coal price differentials.
Hogan, Thorpe and Middleton (1997) provide some caveats to their research. Omitted
variables are not a problem if they contain no new information. However, Hogan, Thorpe
and Middleton note that their model may suffer from omitted variable bias because
variables considered important to the end use performance of coking coals are excluded
from the regression. The omitted variables are REFL, LFLUID and CSR. These variables
are not recorded in the export controls database. Notably, Chang omits REFL and both
Koerner and Chang omit CSR from the regression models.
The two key quality characteristics that the Japanese steel mills nominated under the fair
treatment system are CSR and fluidity (Kahraman, Coin and Reifenstein 1997). Based on
unpublished research, Kahraman, Coin and Reifenstein (1997) state that both factors have
been significant determinants of price for several years, even though there is no substantive
evidence for high fluidity levels to necessarily imply a high level of value in use. As Berndt
(1991) and Kahraman, Coin and Reifenstein (1997) note, however, hedonic regressions
are concerned with consumer perceptions of end use quality attributes rather than actual
performance attributes.
Research method
Hedonic regression equations and data
Separate hedonic regression equations are estimated for hard and semisoft coking coals.
In each case, the dependent variable is a measure of the real coal price. The set of
explanatory variables comprise coal quality characteristics and a set of annual dummy
variables. Following Hogan, Thorpe and Middleton (1997), a linear functional form based
on equation 1 is adopted to avoid multicollinearity problems associated with the inclusion
of quadratic or higher order polynomial terms.
The dataset used in this study comprises a cross-section of prices and quality characteristics
for hard and semisoft coking coal brands exported from Australia to Japan in the years JFY
1992 to JFY 1997. The data were obtained from the Australian Coal Report’s Coal year
(1992 and 1994–97). Coal prices are fob and quoted in US$ a tonne. Real prices are derived
by deflating the published nominal prices by the US consumer price index with a base year
of JFY 1995. Annual dummy variables (D1992, D1993, …, D1997) are included to cater
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for general changes in market conditions over time. Since coal prices are in JFY 1995 US
dollars, DJFY95 was excluded to avoid perfect multicollinearity.
The quality variables included in the regression equations, with notation and expected sign
shown in brackets, are excess moisture (EM, –), inherent moisture (IM, –), ash (ASH, –),
in the analysis since only 15 observations were available for this category over the period
of analysis. Summary statistics of the quality and price variables used in this analysis are
listed in tables 2 and 3 for hard and semisoft coking coals, respectively.
Structural change tests
The key objective in this paper is to test for structural change or price–quality regime
switches over the period JFY 1992 to JFY 1997. A maximum likelihood estimation
technique, based on Quandt’s exogenous switching regime method, is used to test for
structural change (Goldfeld and Quandt 1976; Johnston 1991). With this method, the
coefficients for the quality characteristics may vary between hypothesised regimes and the
distributions of the error terms are assumed to be independent normal random variables
with zero mean and constant but different variances.
Given the hedonic regression equations described above, the maximum likelihood method
involves choosing the regime switch year, including the possibility of no switch, that
maximises the likelihood of observing the data sample as measured by the log likelihood
function. For completeness, tests are conducted for structural change at the beginning of
JFY 1994, 1995 and 1996. Log likelihood function estimates are derived using OLS
estimates of the relevant parameters for each of these hypothesised regime switches.
A statistically definitive likelihood ratio test of the null hypothesis of no structural change
or regime switch in the dataset is not available. However, Goldfeld and Quandt (1976)
report that the use of a Chi-square (3) distribution has been found to be an acceptable
approximation in some applications. The results for this approximate test are reported in
section 5.
A Chow test is commonly used to test the null hypothesis that the regression parameters
are jointly identical between regimes. However, the validity of this test rests on the
assumption that the error variances are common and constant for the regimes (Greene
1993). Greene (1993) describes a test which can be applied to Quandt’s switching regime
regression model of heteroscedasticity across regimes, with homoscedasticity within each
regime. Greene’s likelihood ratio statistic for a test of the null hypothesis of homoscedas-
ticity is identical to the likelihood ratio test for Quandt’s switching regime regression
model. In this case, the likelihood ratio test statistic is asymptotically distributed as a Chi-
square (1) distribution. The results from this test are also reported later.
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Other diagnostic testsFollowing Hogan, Thorpe and Middleton (1997), a set of Ramsey RESET tests are
conducted to test for general specification error, and a set of LM tests are also conducted
to test for the presence of nonconstant error variances.
RESET test statistics (2), (3) and (4) are derived from artificial regressions of the coal price
on the exogenous variables and an increasing number of polynomial terms in the fitted
value of the dependent variable (SHAZAM 1993). A test of the null hypothesis of no model
misspecification error is an F test that all of the coefficients on the fitted values of the
dependent variable in the artifical regression are jointly zero. Due to the power terms in
the regression, RESET tests tend to be very useful in detecting a misspecified functional
form.
The LM tests for departures from homoscedasticity are also based on artificial regressions.
Those considered here are based on regressing the estimated squared errors on the fitted
value of the dependent variable. In the LM (1), (2) and (3) tests, the fitted variable is
expressed in linear, quadratic or log quadratic form, respectively. In each case, the LM test
statistic, which has an asymptotic Chi-square (1) distribution, is computed as the number
of observations multiplied by the unadjusted R squared from the artificial regression.
Results
Structural change test results
The log likelihood function estimates for the assumptions of no structural change and
structural change at the beginning of JFY
1994, 1995 and 1996 are given in table 4.
From these results, it is likely that there
has been a structural shift in price–quality
relationships for both hard and semisoft
coking coal, although the timing of these
shifts differs. JFY 1996 is the most likely
switch point for the change in valuation
regime for hard coking coal, while JFY
1995 is estimated to be the most likely
switch point for the change in valuation
regime for semisoft coking coal.
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Table 4: Log likelihood function estimatesusing Quandt’s exogenous switching regimemethod for hard and semisoft coking coalequations a
Hard SemisoftSwitching point b coking coal coking coal
No switch –98.21 –83.71
JFY 1994 –95.78 –77.80
JFY 1995 –92.53 –69.96
JFY 1996 –83.51 –71.94
a Bold indicates the maximum log likelihood function estimate.b The hypothesised regime switching point or timing of thestructural change is at the beginning of the year indicated.
The likelihood ratio test statistic is 29.39 for the hard coking coal equation and 27.50 for
the semisoft coking coal equation. Based on the Chi-square (3) distribution, the null
hypothesis of no structural change is rejected at the 5 per cent significance level for both
hard and semisoft coking coal equations. Based on Greene’s test using the Chi-square (1)
distribution, the null hypothesis of homoscedasticity across regression groups is rejected
at the 5 per cent significance level for both hard and semisoft coking coal equations. Thus,
the application of Quandt’s switching regime model is appropriate in this study, rather than
Chow’s test for structural change.
Overall, these results indicate that the structure of price–quality relationships for the
Australia–Japan coking coal trade changed fundamentally for semisoft coking coal with
the merging of the soft coking coal category into the semisoft coking coal category after
JFY 1994 and for hard coking coal with the subsequent adoption of the fair treatment
system in JFY 1996.
Estimated regression equations
Following Chang (1995) and Koerner (1996), the full regression results are reported in
this section including variables not significantly different from zero at the 5 per cent level.
A major reason for this approach is that multicollinearity is a potential problem in this
type of study which, if present, lowers the t-values and increases the risk of excluding
significant variables. This may particularly be the case for the hard coking coal equation
in the second period. Omitted variables bias is considered to be a greater problem than the
inclusion of irrelevant variables. The full regression results are also strictly consistent with
the structural change tests and associated diagnostic tests reported above.
Hard coking coal equations
The regression results for the hard coking coal equations are given in table 5. Consistent
with the structural change test results, there are two sets of regression results corresponding
to the estimation periods JFY 1992 to 1995 and JFY 1996 to 1997. Notably, the adjusted
R squared is reduced from 0.97 in the first period to 0.74 in the second period. All RESET
and LM tests for both equations are accepted at the 5 per cent significant level.
For the first period, six quality characteristics are found to have a significant impact on
Number of observations 69 34Adjusted R squared 0.97 0.74
RESET testsRESET test 2 F(1,55) 0.43 F(1,22) 4.06RESET test 3 F(1,54) 0.21 F(1,21) 2.67RESET test 4 F(1,53) 0.38 F(1,20) 1.99
LM testsLM test 1 Chi square(1) 0.05 Chi square(1) 0.28LM test 2 Chi square(1) 0.05 Chi square(1) 0.29LM test 3 Chi square(1) 0.05 Chi square(1) 0.27
* Statistically significant at the 5 per cent significance level (2 tail test). ** Statistically significant at the 10 per cent significancelevel (2 tail test).
The interpretation of the estimated coefficients for the quality variables is summarised as
follows.
• For each one percentage point increase in excess moisture, the hard coking coal price
(in JFY 1995 US dollars) falls by an estimated US$0.64 a tonne in the first period. For
each one percentage point increase in inherent moisture, the hard coking coal price falls
by an estimated US$1.90 a tonne in the first period and US$4.30 a tonne in the second
period.
• For each one percentage point increase in coke strength after reaction, the hard coking
coal price is estimated to increase by US$0.06 a tonne in the first period and, at the 10
per cent significance level, US$0.07 a tonne in the second period.
• For each one unit increase in the crucible swelling number, the hard coking coal price
is estimated to increase by US$0.56 a tonne in the first period and, at the 10 per cent
significance level, US$0.41 a tonne in the second period.
• For each one unit increase in log fluidity, the hard coking coal price is estimated to
increase by US$0.73 a tonne in the first period.
• For each one percentage point increase in sulphur, the hard coking coal price is
estimated to fall by US$2.26 a tonne in the first period and US$2.43 a tonne in the
second period.
• For each one percentage point increase in volatile matter, the hard coking coal price is
estimated to increase by US$0.48 a tonne in the second period.
Overall, price–quality relationships for the Australia–Japan hard coking coal trade have
changed substantially since JFY 1992. Compared with the first estimation period JFY 1992
to 1995, in JFY 1996 and 1997, the extent to which differences in coal quality explain
differences in coal prices has been reduced markedly. The number of quality characteristics
that are found to be significant determinants of hard coking coal prices has also been
reduced, and the mix of quality characteristics has changed.
Compared with Chang (1995), Koerner (1996) and Hogan, Thorpe and Middleton (1997),
a larger number of quality characteristics are found to be significant determinants of hard
coking coal prices, particularly in the first estimation period which is the most comparable
time period. The quality characteristics included in the three previous econometric
exercises vary across the studies, but each represents a subset of the quality characteristics
used in the current study.
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The one exception to this is the use of total moisture rather than excess moisture in Hogan,
Thorpe and Middleton (1997), and total moisture was not included in their final equation.
Based on goodness of fit, excess moisture was preferred to total moisture in the current
study.
The estimated equations in the three previous studies may therefore have problems
associated with omitted variables bias. However, given the relatively high adjusted R
squared in these studies, particularly Koerner (0.95 and 0.93) and Hogan, Thorpe and
Middleton (0.96), there is a possibility that some properties of the coal are measured by
more than a single variable. For example, crucible swelling number, coke strength after
reaction and vitrinite reflectance are all indicators of coking properties of the coal. As a
consequence, exclusion of some quality characteristics may not be as serious as would
otherwise be the case.
Semisoft coking coal equations
Two sets of regression results for the semisoft coking coal equations are given in table 6
corresponding to the estimation periods JFY 1992 to 1994 and JFY 1995 to 1997. The
adjusted R squared is reduced from 0.97 in the first period to 0.84 in the second period.
All RESET and LM tests for both equations are accepted at the 5 per cent significant level.
For the first period, four quality characteristics are found to have a significant impact on
semisoft coking coal prices with the correct sign, including excess moisture, log fluidity,
crucible swelling number and sulphur. Coke strength after reaction was not included in
the semisoft coking coal equations since this information was not published for this coal
category.
In the second period, the coefficients of three quality characteristics — excess moisture,
volatile matter (for coal with a volatile level higher than 30) and crucible swelling number
— are found to be significantly different from zero. Log fluidity and sulphur are not
significant determinants of semisoft coking coal prices in the second period. Similar to
the hard coking coal equations, volatile matter is found to be a significant determinant of
prices of semisoft coking coal (with high volatile matter levels) in the second period, but
not the first.
The interpretation of the estimated coefficients for the quality variables is summarised as
follows.
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• For each one percentage point increase in excess moisture, the semisoft coking coal
price (in JFY 1995 US dollars) falls by an estimated US$0.61 a tonne in the first period
and US$0.92 a tonne in the second period.
• For each one unit increase in the crucible swelling number, the semisoft coking coal
price is estimated to increase by US$0.27 a tonne in the first period and US$0.57 a
tonne in the second period.
• For each one unit increase in log fluidity, the semisoft coking coal price is estimated to
increase by US$0.35 a tonne in the first period.
• For each one percentage point increase in sulphur, the semisoft coking coal price is
estimated to fall by US$1.31 a tonne in the first period.
• For each one percentage point increase in volatile matter above 30, the semisoft coking
coal price is estimated to fall by US$0.40 a tonne in the second period.
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Table 6: Regression results for semisoft coking coal equations