Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

International Journal of Energy Economics and Policy
Vol. 1, No. 1, 2011, pp. 1-17
www.econjournals.com

Energy Consumption and Economic Growth in Vietnam: Threshold
Cointegration and Causality Analysis

Phung Thanh Binh
Environmental Economics Unit, Faculty of Development Economics,

University of Economics, Ho Chi Minh City, Vietnam. Email: ptbinh@ueh.edu.vn.

ABSTRACT: This study investigates the energy consumption-growth nexus in Vietnam. The causal
relationship between the logarithm of per capita energy consumption (LPCEC) and the logarithm of per capita
GDP (LPCGDP) during the 1976-2010 period is examined using the threshold cointegration and vector error
correction models for Granger causality tests. The estimation results indicate that the LPCEC and LPCGDP for
Vietnam are cointegrated and that there is a strong uni-directional causality running from LPCGDP to LPCEC,
but not vice versa. It is also found that the effect of LPCGDP on LPCEC in Vietnam is time-varying (i.e.
significantly different between before and after the structural breakpoint, 1992). The research results strongly
support the neoclassical perspective that energy consumption is not a limiting factor to economic growth in
Vietnam. Accordingly, an important policy implication resulting from this analysis is that government can
pursue the conservation energy policies that aim at curtailing energy use for environmental friendly development
purposes without creating severe effects on economic growth. The energy should be efficiently allocated into
more productive sectors of the economy.

JEL classification: Q4
Key words: Energy consumption, Economic growth, Cointegration, Granger causality, Vietnam

1. Introduction

The high volatility of energy prices and increased greenhouse gas emissions have recently
attracted attention from academics and policy-makers in terms of designing the energy conservation
policy. This kind of policy can, however, not is arbitrarily enacted without considering the causal
relationship between energy consumption and economic growth. In certain context, policy-makers
must understand whether economic growth boosts energy consumption or whether energy
consumption causes economic growth. And this issue has not been reached to a consensus among
energy economists.

The causal relationship between energy consumption and economic growth has been extensively
investigated since the seminal paper of Kraft and Draft in 1978. Different studies have done in
different countries, time periods, and proxy variables using different econometric methodologies
(Ozturk, 2010). However, evidences from empirical researches are still mixed and controversial in
terms of the direction of causality and the intensity of impact on energy policy. Understanding the
linkage between these two variables is extremely significant because energy policy implications
mostly depend upon what kind of causal relationship exists. Bartleet and Gounder (2010) state that it’s
more important to know whether energy consumption causes economic growth than the case where
either economic growth promotes energy consumption or no causal relationship exists between them.
The underlying reason of this justification is that it’s really difficult for policy-makers to enact energy
conservation policies if a country is known as energy-dependent. In the presence of such a
relationship, any structural policies that aim at reducing energy consumption might possibly slow
economic growth (Tsani, 2010).

Theoretically, an appropriate energy policy choice depends on the actual direction of the causal
relationship between energy consumption and economic growth. Ozturk (2010) and Ozturk &
Acaravci (2010) sum up four possible hypotheses about energy-growth nexus. Firstly, no causality
between these variables is referred to as ‘neutrality hypothesis’. In other words, energy is assumed to
be neutral to growth. If this is not a case, conservative or expansive policies on energy consumption
could adversely affect economic growth. Supporters of this view emphasize the role of substitution

mailto:ptbinh@ueh.edu.vn

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

and technological progress. According to Belloumi (2009), the main reason for the neutral impact of
energy on economic growth is that the cost of energy is negligible, so it is not likely to have a
significant impact on economic growth. It has also been argued that the possible impact of energy
consumption on growth will depend on the structure of the economy and the level of economic growth
of the country concerned. As the economy grows, its production structure is likely to shift towards
service sectors, which are not much dependent on energy (Solow, 1974; and Cheng, 1995). Secondly,
uni-directional causality from economic growth to energy consumption supports the ‘conservation
hypothesis’. This implies that a country might implement the energy conservation policy without
having any adverse effect on economic growth. Thirdly, uni-directional causality from energy
consumption to economic growth is commonly considered as ‘energy-led growth hypothesis’. Within
this situation, policy makers should pay special attention on restrictions of energy use because this
action may, to which extent, impede economic growth. Proponents of this hypothesis believe that
energy is a critical input of production and plays as a complement to the basic factors of land, labour
and capital. If this is a case, energy is said to be a limiting factor of economic growth (Stern, 1993; and
Cleveland et al., 2000). Finally, bi-directional causality between energy consumption and economic
growth is known as ‘feedback hypothesis’. This provides an insight that energy consumption and
economic growth are jointly determined and affected together.

Chen et al. (2007) explains the mixed findings from previous studies are due to differences in not
only data set, econometric approaches, but also countries’ characteristics. For this reason, it’s very
dangerous to design future energy policy of one country based on experiences of others. Accordingly,
a country-specific causality study between energy consumption and economic growth must be done to
provide deep insights into design of energy policies. Therefore, is it correct to believe in the statement
of ‘energy for economic development’ stated by the Vietnam energy industry? What is the evidence
for the national energy development strategy that still provides special favour for energy sectors?

A great number of causality studies have been conducted for many countries around the world;
however, none have focused on the causal relationship between energy consumption and economic
growth for Vietnam. Hence, it’s now necessary to examine this causal relationship in Vietnam. This
research aims at answering the following questions: (1) Does long-term equilibrium exist between
energy consumption and economic growth in Vietnam? (2) Is the energy consumption quickly
adjusted to reach the long-run equilibrium if an instantaneous shock to its consumption occurs? (3)
Which of the above hypotheses is acceptable for the case of Vietnam?

The remainder of this paper is organized as follows: Section 2 briefly reviews the literature on the
energy-growth nexus. Section 3 presents the data and methodology used. Section 4 discusses the
empirical results. We’ll then provide the policy implications and give conclusions in Section 5.

2. Literature Review

The relationship between energy consumption and economic growth has been theoretically
investigated through two main different approaches. In the neoclassical growth models, energy is
simply considered as an intermediate input of production (Tsani, 2010). According to Bartleet and
Gounder (2010), proponents of this view think that there are some mechanisms by which economic
growth could remain in spite of a limited source of energy resources. The underlying explanation for
this is built upon the possibility of technological change and substitution of other physical inputs for
energy to use existing energy resources efficiently, and to generate renewable energy resources that
are not subject to binding supply constraints (Solow, 1974, 1997; Stiglitz, 1997). Accordingly, energy
is merely one of the non-essential inputs in production process. In other words, the advocates of this
theory support the ‘neutrality hypothesis’ and ‘conservation hypothesis’. These hypotheses imply that
energy supply restrictions might not have any harmful effect on economic growth. Thus, the
government can simultaneously adopt the energy conservation and economic growth policies (Bartleet
and Gounder, 2010).

On the other hand, the ecological economic theory states that energy consumption is a limiting
factor to economic growth, especially in modern economies. Ecological economists judge that
technological progress and other physical inputs could not possibly substitute the vital role of energy
in production process (Stern, 1993, 2000). They even consider energy as the prime source of value
because other factors of production such as labour and capital cannot perform without energy
(Belloumi, 2009). The promoters of this perspective protect the so-called ‘growth hypothesis’, and

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

hence, advise that any shock to energy supply will ultimately have a negative impact on economic
growth. As a result, they are against the energy conservation policies.

Lots of empirical studies on energy consumption and economic growth nexus using different data
set from different countries have so far provided various and contradictive results. The idea of
causality between energy consumption and economic growth was first introduced in the seminal paper
of Kraft and Kraft (1978) with the application of a standard version of Granger causality (standard
Granger) test, which provided proof to support a uni-directional long-run relationship running from
gross domestic product (GDP) to energy consumption for the USA over the 1947-74 period. This
study suggests that government could pursue the energy conservation policies. On the other hand, by
employing the Sims causality technique, Akarca and Long (1980) showed no evidence of causality
between energy consumption and GDP, so they criticized drastically the Kraft and Kraft’s result in
terms of the temporal sample instability. Since then, many academics have zealously joined the
debate, but they have never reached the consensus (Belloumi, 2009). In the same manner, Yu and
Hwang (1984) took up the Sims causality test with annual data and found no causality between energy
consumption and GDP in the USA for the 1947-79 period. When using quarterly data with the same
testing method, conversely, these authors discovered a uni-directional causality running from gross
national product (GNP) to energy consumption in the USA for the 1973-81 period (Belloumi, 2009).

Yu and Choi (1985) employed the standard Granger causality test for the 1954-1976 period to
explore the causal linkages between GNP and various types of energy consumption for a set of
countries. Their empirical studies indicated that uni-directional causality running from economic
growth to energy consumption for Korea, uni-directional causality running from energy consumption
to income for the Philippines, while no causality existed in the USA, Poland and UK. Erol and Yu
(1987) employed both Sims and Granger causality tests and found unidirectional causality from
energy consumption to income for West Germany while bi-directional causality for Italy, and no
evidence of causality for UK, Canada and France. Besides, they also uncovered the uni-directional
causality running from energy consumption to economic growth for Japan over the 1950-1982 period.
On the contrary, when the sample was restricted to the 1950-1973 period, this causal relationship was
no longer significant. Hwang and Gum (1992) used the cointegration and error correction model, and
the bi-directional causal relationship between energy consumption and economic growth was observed
in Taiwan for the period 1955-1993.

By using the cointegration and error-correction version of Granger causality test (ECM), Cheng
(1995) realized the presence of uni-directional causality running from economic growth to energy
consumption in India. In addition, Masih and Masih (1996, 1997) found the existence of cointegration
between energy and GDP in India, Pakistan and Indonesia, but non-cointegration in Malaysia,
Singapore and the Philippines. With the same data set, these authors applied the vector error correction
model (VECM), and recognised a uni-directional causality running from energy consumption to
income in India, a uni-directional causality running from economic growth to energy consumption in
Indonesia, and bi-directional causality in Pakistan. This study also made use of the standard Granger
causality test for the non-cointegrated countries (including Malaysia, Singapore and the Philippines),
but did not find any Granger causality.

Glasure and Lee (1997) examined the causality between energy consumption and GDP for South
Korea and Singapore and reported different results from different methodologies used. The standard
Granger causality tests revealed no causal relationship for South Korea and a uni-directional causal
relationship running from energy consumption to GDP for Singapore, while the ECM model gave
signal of bi-directional causality for both countries. Cheng and Lai (1997) applied Hsiao’s version of
Granger causality to investigate the link between energy consumption and GDP for Taiwan for 1955–
1993 period. This study showed that causality runs from GDP to energy consumption without
feedback in Taiwan. Yang (2000) re-examined the causality between energy consumption and GDP
for Taiwan using updated data for the 1954–1997 period. The finding of this paper totally denies the
findings of Cheng and Lai (1997) of unidirectional causality from GDP to energy consumption. They
found evidences of bi-directional causality between energy consumption and GDP.

Asafu-Adjaye (2000) tested the causal relationship between energy use and income in four Asian
countries (including India, Indonesia, Thailand and the Philippines) using the ECM models. The test
results indicated a uni-directional causality running from energy to income in India and Indonesia, and
a bi-directional causality in Thailand and the Philippines. Aqeel and Butt (2001) used the ECM models

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

to investigate the causal relationship between energy consumption and economic growth as well as
between energy consumption and employment for Pakistan. The results inferred that economic growth
caused total energy consumption. Soytas and Sari (2003) studied causality between energy
consumption and GDP for the G7 countries and for the top 10 emerging economies. Their research
results found a bi-directional causality for Argentina, uni-directional causality running from GDP to
energy consumption in Italy and Korea, and uni-directional causality running from energy
consumption to GDP in Turkey, France, Germany and Japan.

Paul and Bhattacharya (2004) re-examined the direction of causality between energy consumption
and economic growth in India by employing the ECM model for the 1950–96 period. As a result, they
realized that a bi-directional causality existed between energy consumption and economic growth.
Besides, they also applied the Johansen cointegration testing approach and figured out the same
direction of causality between energy consumption and economic growth. Altinaya and Karagol
(2004) detected causality between the GDP and energy consumption in Turkey employing the Hsiao’s
version of Granger causality for the 1950–2000 period, characterized by structural break. The main
conclusion of this study is that there is no evidence of causality between energy consumption and GDP
in Turkey based on the detrended series.

Lee (2005) investigated the cointegration and the causality relationship between energy
consumption and GDP in 18 developing countries, using data for the 1975–2001 period and
employing panel unit root tests, heterogeneous panel cointegration and panel ECM models. The
empirical results supported a long-run cointegration relationship between energy consumption and
GDP after allowing for the heterogeneous country effects. The evidence illustrated that there were
only long-run and short-run causalities running from energy consumption to GDP. This result
suggested that energy conservation policies might, to which extent, harm economic growth in
developing countries. Wolde-Rufael (2005) ran a cointegration and a modified version of the Granger
causality test to investigate the long-run and causal relationship between per capita GDP and per
capita energy use for 19 African countries for the 1971–2001 period. Their results offered further
evidence of the long-run relationship for eight out of the nineteen countries and causality for twelve
out of nineteen countries.

Mehrara (2007) examined the causal relationship between the per capita energy consumption and
the per capita GDP in a group of eleven oil-exporting countries (including Iran, Kuwait, Saudi Arabia,
United Arab Emirates, Bahrain, Oman, Algeria, Nigeria, Mexico, Venezuela and Ecuador) by using
panel unit root tests and panel cointegration tests. The test results found a uni-directional causality
running from economic growth to energy consumption for these oil-exporting countries. Interestingly,
the results recommened that energy conservation policies through reforming energy prices could not
have any adverse effect on economic growth.

Chiou-Wei et al. (2008) conducted both linear and nonlinear Granger causality tests to examine
the causal relationship between energy consumption and economic growth for a panel of Asian newly
industrialized countries as well as the USA for the 1954–2006 period. Their study again supported a
neutrality hypothesis for the USA, Thailand, and South Korea. Moreover, they unearthed the existence
of a uni-directional causality running from economic growth to energy consumption for Philippines
and Singapore while energy consumption might have negative effects on economic growth for
Taiwan, Hong Kong, Malaysia and Indonesia. Chontanawat et al. (2008) tested for causality between
energy and GDP using a consistent data set and Granger test for thirty OECD countries and seventy
non-OECD countries. They discovered that causality running from energy to GDP appeared to be
more prevalent in the developed OECD countries.

Tsani (2010) studied the causal relationship between aggregated and disaggregated levels of
energy consumption and economic growth in Greece for the 1960–2006 period by using the
methodology proposed by Toda and Yamamoto (1995). At the aggregated levels of energy
consumption, the empirical findings suggested the presence of a uni-directional causality running from
total energy consumption to real GDP. At the disaggregated levels, the results indicated a bi-
directional causal relationship between industrial and residential energy consumption to real GDP, and
no causality between transport energy consumption and real GDP. The energy policy implication from
these findings focused on the demand side and energy efficiency improvements in order to put less
impact on economic growth.

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

Esso (2010) investigated the cointegration and causal relationship between energy consumption
and economic growth in seven Sub-Saharan countries over the 1970–2007 period. This study used the
Gregory and Hansen (1996b) threshold cointegration approach and the Toda and Yamamoto (1995)
version of Granger causality test. The test results revealed that energy consumption was cointegrated
with economic growth in Cameroon, Cote d'Ivoire, Ghana, Nigeria and South Africa in the presence of
a structural break. Moreover, threshold cointegration test and ECM models suggested that economic
growth had a significant positive long-run impact on energy consumption in these countries before
1988 while this effect was negative after the breakpoint, 1988, for Ghana and South Africa.
Furthermore, Granger-causality tests suggested a bi-directional causality between energy consumption
and real GDP in Cote d'Ivoire and a uni-directional causality running from real GDP to energy
consumption in the case of Congo and Ghana.

By using the recently developed autoregressive distributed lag (ARDL) bounds testing approach
of cointegration and dynamic vector error correction (VECM) model for four Eastern European
countries, Ozturk and Acaravci (2010) figured out that there is weak evidence about the long-run and
causal relationships between energy consumption (or electric power consumption) and economic
growth. Specifically, they found that evidence of cointegration and bi-directional strong Granger
causality between these variables is only found in Hungary for the 1980-2006 period. This study
contributes not only the proof of causality, but the methodology used. The authors explained clearly
the ARDL bounds testing approach used in the energy-growth linkage. In addition, Ozturk et al.
(2010) employed the panel cointegration and causality analysis to investigate the differences in energy
consumption and economic growth relationship among three groups of 51 countries classified as low
income countries, lower middle income countries, and upper middle income countries for the 1971-
2005 period. The test results indicate that there exists cointegration between energy consumption and
real GDP for all three income groups. From the panel causality tests, they conclude that there is a
long-run Granger causality running from GDP to energy consumption for low income countries and
bi-directional Granger causality between these variables for the other groups. Furthermore, these
authors also provide evidence that there is no strong relation between energy consumption and
economic growth in these countries.

In summary, a general judgment is that the results are still mixed: that is, while some studies find
causality running from economic growth to energy consumption, others figure out causality running
from energy consumption to economic growth and even some studies suggest no causality and/or bi-
directional causality between these two variables. The differences among these studies lie on the
specific country characteristics, sample periods, research methodologies, and proxy variables.

3. Data and Methodology
3.1 Data Collection and Unit Root Tests

This paper uses the time series data of per capita GDP (PCGDP) and per capita energy
consumption (PCEC) for the 1976-2010 period in Vietnam. Data are obtained from three sources: (i)
the World Development Indicators (2011); (ii) International Financial Statistics (2011); and the
Vietnam’s General Statistics Office. In this study, per capita energy consumption is expressed in terms
of kg oil equivalent and per capita GDP is expressed in constant 2000 US$. The choice of the starting
period was constrained by the availability of data and the historical milestone as well. The Vietnam
War ended in 1975, and the country was united in 1976. The trends of PCGDP and PCEC for Vietnam
are graphically depicted in Figure 1. It’s noted that all variables are transformed into natural
logarithms in order to reduce heteroskedasticity and obtain the growth rate of the relevant variables by
their differenced logarithms (Orturk and Acaravci, 2010).

It seems that there might be a structural break in these series around the year 1991. By using the
Quandt-Andrews breakpoint test, we recognize that the PCGDP is broken at the year 1992, while the
PCEC at the year 1993. These breakpoints are depicted in Figure 2 and 3. This fact might, to which
extent, imply that PCGDP causes PCEC. The final answer will, however, be clear after performing
Granger causality tests presented in the empirical results. These figures also indicate that both series
have increasingly grown since these breakpoints.

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

200

300

400

500

600

700

800

900

1980 1985 1990 1995 2000 2005 2010

PCEC PCGDP

Figure 1: Vietnam's per capita energy consumption and per capita GDP

300

350

400

450

500

550

600

650

700

750

1980 1985 1990 1995 2000 2005 2010

PCEC PCECF

Figure 2: Fitted trend of per capita energy consumption
1993

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

200

300

400

500

600

700

800

900

1980 1985 1990 1995 2000 2005 2010

PCGDP PCGDPF

Figure 3: Fitted trend of per capita GDP
1992

In order to establish the order of integration of the variables concerned, this study first employs the

conventional unit root tests widely known as the augmented Dickey-Fuller (ADF) and Phillips-Perron
(PP) unit root tests. Generally, a variable is said to be integrated of order d, written by I(d), if it turns
out to be stationary after differencing d times. The variable is integrated of order greater than or equal
to 1 is non-stationary. According to Asteriou and Hall (2007), most economic variables are
cointegrated of order 1.

In testing for the existence of a unit root of the time series Yt (H0: δ = 0), we can select one out of
the following three possible forms of the ADF test (Yt can be either LPCECt or LPCGDPt):

∑
=

−− ++=
p

1i
titi1tt uYYY ∆βδ∆ (1)

∑
=

−− +++=
p

1i
titi1t0t uYYY ∆βδα∆ (2)

∑
=

−− ++++=
p

1i
titi1t10t uYYTY ∆βδαα∆ (3)

The difference between the three regressions concerns the presence of the deterministic elements
α0 and α1T. For choosing the best one among the three equations, we will first plot the data (of each
series) and then observe the graph because it can, to which extent, indicate the presence or not of the
deterministic trend regressors.

Esso (2010) states a break in the deterministic trend affects the outcome of unit root tests because
many previous studies have found that the conventional unit root tests fail to reject the unit root
hypothesis for series that are actually trend stationary with a structural break. Perron (1989) showed
that a Dickey and Fuller (1979) type test for unit root is not consistent if the alternative is that of a
stationary noise component with a break in the slope of the deterministic trend. His main point is that
the existence of exogenous shock which has a permanent effect will lead to a non-rejection of the unit
root hypothesis even though it is true. Perron (1989) proposed a unit root test allowing for a structural
break with three alternative models: crash model (i.e., shift in the intercept), changing growth model
(i.e., change in the slope) and the change both in the intercept and the slope. Several studies have
found that the conventional unit root tests fail to reject the unit root hypothesis for the series that are
actually trend stationary with a structural break. On the other hand, the Perron (1989) test has been

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

generally criticized for treating the time of break as exogenous (i.e., the time of break is known a
priori). (Christiano, 1992; and Altinay and Karagol, 2004).

Zivot and Andrews (1992) further developed the Perron unit root tests that consider the breakpoint
(τb) as endogenous. To test for a unit root against the alternative of trend stationarity process with a
structural break both in slope and intercept, the following regressions are used:

∑
=

−− +++++=
p

1i
titi1tbtt uYYT)(DUY ∆ϕαβτθµ (4)

∑
=

−− +++++=
p

1i
titi1tbtt uYYT)(DTY ∆ϕαβτγµ (5)

∑
=

−− ++++++=
p

1i
titi1tbtbtt uYY)(DTT)(DUY ∆ϕατγβτθµ (6)

where DUt and DTt are dummy variables for a mean shift and a trend shift respectively; DUt(τb) = 1 if
t > τb and 0 otherwise, and DTt(τb) = t- τb if t > τb and 0 otherwise. In other words, DUt is a sustained
dummy variable that captures a shift in the intercept, and DTt represents a shift in the trend occurring
at time τb. The breakpoint τb can be found by using the Quandt-Andrews breakpoint test. The optimal
lag length p is also determined by using the general to specific approach so as to minimize the AIC or
SIC. The Zivot and Andrews (1992) unit root test suggests that we reject the null hypothesis of a unit
root if computed α̂t is less than the left-tail critical t value.

3.2 Cointegration Analysis

Once the order of integration of each variable is established, we then evaluate whether the
variables under consideration is cointegrated. According to Engle and Granger (1987), a linear
combination of two or more nonstationary series (with the same order of integration) may be
stationary. If such a stationary linear combination exists, the series are considered to be cointegrated
and long-run equilibrium relationships exist. Thanks to the existence of cointegration, although the
series are individually nonstationary, they cannot drift farther away form each other arbitrarily.
Cointegration implies that causality exists between the two variables, but it does not indicate the
direction of the causal relationship. The presence of cointegration among the variables rules out the
possibility of ‘spurious’ regression (Belloumi, 2010). There are various approaches to test for
cointegration, say, Engle and Granger approach, Johansen approach, ARDL bounds testing approach
(by Pesaran et al., 2001), and Gregory and Hansen approach.

According to Belloumi (2010), the bivariate approach of Engle and Granger is very restrictive
because it can be applied only if there is one cointegrating relation. And the most commonly used
method is the Johansen cointegration test based on the autoregressive representation discussed by
Johansen (1988) and Johansen and Juselius (1990). This test determines the number of cointegrating
equations for any normalization used. It provides two different likelihood ratio tests; one is based on
the trace statistic and the other on the maximum eigenvalue.

Esso (2010) states that the cointegration framework of Engle and Granger (1987), and Johansen
(1988) has its limitations especially when dealing with economic data containing the structural breaks.
In this case, we tend to reject the hypothesis of cointegration, albeit one with stable cointegrating
parameters. The reason is that the residuals from cointegrating regressions capture unaccounted breaks
and thus typically exhibit nonstationary behaviour. Therefore, it’s necessarily to employ the non-linear
techniques for testing cointegration if the economic data has structural breaks. One of the widely used
methods is the Gregory and Hansen (1996a,b) threshold cointegration test. And the test equation is
expressed as below:

tbtt2t1bt21bt21t u)(DUXX)(DUTT)(DUY +⋅++⋅+++= ταατββτµµ (7)

tbtt2t1bt21bt21t u)(DUYY)(DUTT)(DUX +⋅++⋅+++= ταατββτµµ (8)
where µ1 and µ2 represent the intercept before the shift and the change in the intercept at the time of
the shift; β1 and β2 are respectively the trend slope before the shift, the change in the trend slope at the
time of the shift; α1 is the cointegrating slope coefficient before the regime shift, α2 denotes the
change in the cointegrating slope coefficient at the time of the regime shift; Yt and Xt denote LPCECt
and LPCGDPt. The standard methods to test the null hypothesis of no cointegration are residual-based.

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

The equations (7) and (8) are estimated by OLS method, and the unit root tests are applied to the
regression errors (Gregory and Hansen, 1996a).

3.3 Granger Causality Analysis

Let’s denote LPCECt and LPCGDPt for the natural logarithms of the corresponding energy
consumption and real GDP per capita respectively; and suppose that LPCECt and LPCGDPt are both
integrated of order 1, the VAR model developed by Granger (1969) can be defined as:

1j
jtj

1i
itit uLPCGDPLPCECLPCEC +++= ∑∑

=
−

=
− ∆γ∆βα∆ (9)

1j
jtj

1i
itit uLPCECLPCGDPLPCGDP +++= ∑∑

=
−

=
− ∆δ∆θϕ∆ (10)

This study use the Akaike’s information criterion (AIC) and Schwarz’s Bayesian criterion (SBC)
to determine the optimal lag length of ∆LPCECt and ∆LPCGDPt. The equations (9) and (10) are first
estimated by OLS method, and then we apply the normal F Wald test for the joint significance of the
coefficients on the lagged terms in the unrestricted models. Specifically, the following null hypotheses
are necessarily tested:

nconsumptioenergy causeGranger not doesgrowth economicor 0 :H )A(

1j
j0 ∑

=
=γ

growth economic causeGranger not doesn consumptioenergy or 0 :H )B(

1j
j0 ∑

=
=δ

It is possible to have that (a) energy consumption causes economic growth (reject B, but do not

reject A), (b) economic growth causes energy consumption (reject A, but do not reject B), (c) there is a
bi-directional feedback between energy consumption and economic growth (reject A and B), and (c)
energy consumption and economic growth are independent (do not reject A and B).

According to Mehrara (2007), the most popular method for Granger causality tests, is based on the
VECM if variables are cointegrated. The VECM can avoid shortcomings of the VAR based models in
distinguishing between a long- and a short-run relationship among the variables. Theoretically,
cointegration implies the existence of causality between variables, but it does not indicate the direction
of the causal relationship. The VECM is estimated by using the following VAR model:

1j
jtj

1i
iti1t,11t uLPCGDPLPCECECTLPCEC ++++= ∑∑

=
−

=
−− ∆γ∆βπα∆ (11)

1j
jtj

1i
iti1t,22t uLPCECLPCGDPECTLPCGDP ++++= ∑∑

=
−

=
−− ∆δ∆θπϕ∆ (12)

where the error correction term (ECTt-1) is derived from the long-run cointegration relationship and
measures the magnitude of the past disequilibrium. The coefficients, π of the ECTt-1 represent the
deviation of the dependent variables from the long-run equilibrium.

Within this VECM model, we can examine whether the relationship between energy consumption
and economic growth is weak Granger causality, long-run Granger causality, or strong Granger
causality. The weak Granger causality exists if we can find the short-run relationship between energy
consumption and economic growth which is based on the normal F Wald test for the joint significance
of the coefficients on the lagged terms in the unrestricted models (equation (11) and (12)) in the same
manner as the null hypotheses (A) and (B). The long-run causality can be tested by looking at the
significance of the speed of adjustment, which is the coefficient of the error correction term. This is
easily based on the t statistic. Specifically, we must test the following null hypotheses:

run-long in thecausality -nonGranger or 0 :H )C( 10 =π
run-long in thecausality -nonGranger or 0 :H )D( 20 =π

According to Belloumi (2010), the significance of ‘π’ indicates that the long-run equilibrium
relationship is directly driving the dependent variable. If π, say, in equation (11) is zero, then it can be
implied that the change in energy consumption does not respond to deviation in the long-run
equilibrium for the t-1 period.

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

The strong Granger causality between energy consumption and economic growth, which is based
on the normal F Wald test for joint significance of both the coefficient associated with the ECTt-1 and
the coefficients on the lagged terms in the unrestricted models (equation (11) and (12)) as follows:

nconsumptioenergy causestrongly not doesgrowth economicor 0 :H )C(
q

1j
j10 ∑

=
== γπ

growth economic causesstrongly not doesn consumptioenergy or 0 :H )D(
q

1j
j20 ∑

=
== δπ

Above models could include dummy variables in order to take into account the existence of the
possible structural breaks during the study sample. In addition, we sometimes include the trend
variable if there is the existence of deterministic trend in the relationship between energy consumption
and economic growth. These inclusions depend on the actual data property.

4. Empirical Results
4.1 Unit Root Tests

The high coefficient of correlation between two variables (0.98) does not imply cointegration.
According to Granger (1988), the condition for cointegration is that each of the variables should be
integrated of the same order (more than zero) or that both series should contain a deterministic trend
(Belloumi, 2010). Table 1 reports the results of the standard unit root tests (ADF and PP) on the
integration properties of the LPCEC and LPCGDP variables for Vietnam. Because the actual values of
these series exhibit trends, so all unit root test regressions include constant and trend terms. The test
results suggest that none of the series are stationary at any levels and only the LPCEC series is
stationary in its first difference. On the other hand, the test results seem not much different between
LPCGDP and its first difference. The reason for this situation is that because the series are basically
characterized by structural breaks, so the ADF and PP tests may be suspect. Therefore, we have to
employ the Zivot and Andrews (1992) unit root tests.

Table 1: Unit root test results using ADF and PP

Variable
ADF PP

Level First difference Level First difference
LPCEC -0.58 -5.25* -0.60 -5.25*

LPCGDP -2.48 -1.98 -2.51 -2.04
1% Critical value -4.26 -4.26 -4.25 -4.25
5% Critical value -3.55 -3.55 -3.55 -3.55
10% Critical value -3.20 -3.20 -3.20 -3.20

Note: * indicates significance at 1% level.

The null hypothesis of the Zivot and Andrews (1992) is α = 1, i.e., a unit root, against the
alternative that the series are trend stationary process with a structural break. The estimated
regressions that yield the minimum α̂t statistic with the optimum number of k regressors are presented
in Table 2. As can be seen from this table, there is sufficient evidence against the unit root hypothesis
for the first differenced series. In other words, the LPCEC and LPCGDP variables are now said to be
individually I(1).

Table 2: Unit root test results using Zivot and Andrews (1992)

Variable
ZA (1992)

Level First difference
LPCEC -3.13 (0) -6.29***(0)

LPCGDP -3.01 (1) -3.95* (0)
Note: The numbers in parentheses are the lag order. The lag parameters are selected based on the AIC. *** and *
indicate significance at the 1% and 10% levels respectively.

4.2 Cointegration Tests
Applying the Engle and Granger cointegration approach discussed in Section 3.2, we obtain the

results as shown in Table 3. Residuals obtained from OLS regressions between LPCEC and LPCGDP
is then tested by using the ADF and PP tests. The test results indicate that there seems to be no

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

cointegrating relationship between the two variables. This approach, however, is very restrictive
because it cannot deal properly with economic data containing the structural breaks. Therefore, it’s
hard to have any conclusion about the cointegration between these variables from this simplified test.

Table 3: Unit root tests for residuals (Engle and Granger) using ADF and PP
from equation ADF PP
LPCEC = f(LPCGDP) -1.15 -1.31
1% Critical value -2.63 -2.63
5% Critical value -1.95 -1.95
10% Critical value -1.61 -1.61

We then employ the Johansen approach for testing cointegration between LPCEC and LPCGDP.
Before performing the Johansen cointegration tests, this study uses the AIC to determine the optimum
lag length. As a result, the optimum lag length is 6. In these tests, we denote the number of
cointegrating vectors by r0; the trace test is calculated under the null hypothesis H0: r0 ≤ r, and the
alternate hypothesis H1: r0 > r; and the maximum eigenvalue test is calculated under the null
hypothesis H0: r0 = r, and the alternate hypothesis H1: r0 > r. The test results are presented in Table 4
and Table 5. If the test statistic is greater than the critical value at a given level of significance, we
reject the null hypothesis, and vice versa.

Table 4: Johansen cointegration estimation results between PCEC and PCGDP
Rank test (trace)
Model 2 ( intercept (no trend) in CE, no intercept or trend in VAR)
Number of cointegration Eigenvalue Trace statistic 5% Critical value
None 0.554 27.13 20.27
At most 1 0.149 4.54 9.16
Model 3 (intercept in CE and VAR, no trend in CE and VAR)
None 0.542 21.89 15.87
At most 1 0.000 0.017 3.84
Model 4 (intercept in CE and VAR, linear trend in CE, no trend in VAR)
None 0.554 36.35 25.87
At most 1 0.388 13.73 12.52

These tests indicate clearly that all null hypotheses (r0 ≤ 0 for trace statistic, and r0 = 0 for
maximum eigenvalue statistic) are rejected at the 5% level of significance. Thus, we can reject the
hypothesis of no cointegrating equation between energy consumption and economic growth at the 5%
significance level. However, under the null hypothesis (r0 ≤ 1 for trace statistic, and r0 = 1 for
maximum eigenvalue statistic), both maximum eigenvalue statistic and trace statistic are below the 5%
level of significance. Accordingly, energy consumption and economic growth are said to be
cointegrated at 5% level of significance. As a result, there must be a long-run relationship between per
capita energy consumption and economic growth for Vietnam in the sample period.

Table 5: Johansen cointegration estimation results between PCEC and PCGDP
Rank test (maximum eigenvalue)
Model 2 (intercept (no trend) in CE, no intercept or trend in VAR)
Number of cointegration Eigenvalue Max-Eigen statistic 5% Critical value
None 0.554 22.88 15.89
At most 1 0.149 4.54 9.16
Model 3 (intercept in CE and VAR, no trend in CE and VAR)
None 0.542 21.88 14.26
At most 1 0.000 0.017 3.84
Model 4( intercept in CE and VAR, linear trend in CE, no trend in VAR)
None 0.554 22.62 19.38
At most 1 0.387 13.73 12.52

As we believe that the sample series contain structural breaks, we apply the threshold
cointegration approach proposed by Gregory and Hansen (1996a,b) to make sure the above Johansen
cointegration tests. We first estimate the regression equations (7) and (8), then obtain the residuals. We

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

also plot line graph and correlogram of the residuals in a priori to recognize the stationarity of the
series. The results of the residual-based unit root tests presented in Table 6 indicate that there really
exists a cointegrating equation between energy consumption and economic growth.

Table 6: Unit root tests for residuals (threshold cointegration) using ADF and PP
from equation ADF PP
LPCEC = f(DUt, T, LPCGDP, LPCGDP*DUt) -4.24 -4.36
LPCGDP = f(DUt, T, T*DUt, LPCEC, LPCEC*DUt) -4.50 -3.47

1% Critical value -2.63 -2.63
5% Critical value -1.95 -1.95
10% Critical value -1.61 -1.61

4.3 Granger Causality Tests
It’s already known that cointegration implies the existence of Granger causality, but it does not
indicate the direction of the causality relationship. According to Altinay and Karagol (2004), taking
the first differences of the series with structural breaks to achieve stationarity may lead to spurious
results. In this case, they suggest that stationary series can be obtained by detrending the series taking
the estimated breakpoints into consideration through the following regression:

tLPCECtDTT)b(tDUtLPCEC
~

++++= αβτδµ (13)

tLPCGDPtDTT)b(tDUtLPCGDP
~

++++= αβτδµ (14)

where tLPCEC
~

and tLPCGDP
~

are detrended stationary series. This alternative could be appropriate
if we just consider the short-run relationship between two variables of interest. Therefore, this might
not be the best choice of this empirical study. In order to investigate the direction of causality between
energy consumption and economic growth for Vietnam, this section employs the VECM models
because the properties of these series are completely satisfactory.

In order to specify the proper VECM models, we should now consider two important things. First,
it’s necessary to understand whether the first differences of the two series exhibit deterministic trend.
In doing so, we simply have a look at the line graphs depicted in Figure 4. And the result is quite
straightforward. Accordingly, the most specified VECM models can be written as follows:

1j
jtj

1i
iti1t,1110t uLPCGDPLPCECECTTLPCEC +++++= ∑∑

=
−

=
−− ∆γ∆βπαα∆ (15)

1j
jtj

1i
iti1t,2210t uLPCECLPCGDPECTTLPCGDP +++++= ∑∑

=
−

=
−− ∆δ∆θπϕϕ∆ (16)

where ECT1,t-1 and ECT2,t-1 are the equilibrium error lagged one period, obtained from the threshold
cointegrating equations. Since all variables are stationary, so the OLS method is reliable for estimating
equations (15) and (16). The long-run effects of economic growth on energy consumption and vice
versa are provided by Table 7 (cointegrating equations), and estimates for the dynamic relationship
between these two variables are represented in Table 8 (VECM equations). Second, we have to specify
the optimum lag length of the VAR model based on the minimum Akaike information criterion. The
test result indicates that the optimum lag length of the VAR model is 1.

Table 7: Threshold cointegrating equations
Independent variable: LPCECt
Variable Intercept DUt T T·DUt LPCGDP LPCGDP·DUt
Coef. 7.624 -4.957 0.008 -0.334 0.871
t-Stat. (12.57) (-9.96) (4.50) (-2.92) (9.75)
R2 = 0.99 F-Stat. = 1563.64 DW = 1.41
Independent variable: LPCGDPt
Variable Intercept DUt T T·DUt LPCEC LPCEC·DUt
Coef. 13.996 -11.005 0.017 0.028 -1.482 1.770
t-Stat. (11.56) (-7.82) (15.78) (5.20) (-7.17) (7.06)
R2 = 0.99 F-Stat. = 6874.53 DW = 1.20

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

-.04

-.02

.00

.02

.04

.06

.08

.10

.12

1980 1985 1990 1995 2000 2005 2010

D(LPCEC) D(LPCGDP)

Figure 4: First differences of LPCEC and LPCGDP series

The estimation results of the error correction models in Table 8 indicate that the lagged error

correction terms have the negative signs as expected. While the coefficient of the lagged error
correction term in the LPCGDP equation is zero and insignificant, this is relatively high (-0.898, or
89.8%) and significant at the 1% level in the LPCEC equation. This significance implies that the
change in logarithm of per capita energy consumption (growth rate) does rapidly respond to any
deviation in the long-run equilibrium (or short-run disequilibrium) for the t-1 period. In other words,
the effect of an instantaneous shock to per capita energy consumption will be completely adjusted in
the long-run. In addition, the effect of economic growth on energy consumption seems to be time-
varying. The long-run effect on energy of growth is -0.334 before the date break, 1993, while it is (-
0.334+0.871) after the breakpoint. This means that economic growth has positively influenced energy
consumption after 1993. This provides further evidence to believe that economic liberalization and
reforms in 1989 which have helped Vietnam to achieve relatively rapid growth rates have resulted in a
corresponding rapid increase in demand for energy.

Table 8: Estimation results of the error correction models
Dependent variable: ∆LPCECt
Variable Intercept T ECT1,t-1 ∆LPCECt-1 ∆LPCGDPt-1
Coef. -0.898 0.372 0.396
t-Stat. (-3.955) (2.254) (3.471)

R2 = 0.55 DW = 2.05
Dependent variable: ∆LPCGDPt
Variable Intercept T ECT2,t-1 ∆PCGDPt-1 ∆PCECt-1
Coef. -0.081 0.991 0.000
t-Stat. (-0.056) (14.708) (0.000)
R2 = 0.793 DW = 2.13

From the above estimations, we can derive the following equations that express the long-run
relationship between energy consumption and economic growth (all coefficients are statistically
significant at 1% level):

Energy Consumption and Economic Growth in Vietnam: Threshold Cointegration and Causality Analysis

∆LPCECt = –0.898ECTt-1 + 0.372∆LPCECt-1 + 0.396∆LPCGDPt-1 + ut (17)

ECTt-1 = LPCECt-1 – 7.624 + 4.957DUt – 0.008T + 0.334LPCGDPt-1

– 0.871LPCGDPt-1·DUt (18)

LPCECt = – 7.624 + 4.957DUt + 0.0072T + 0.096LPCGDPt-1 +

0.782LPCGDPt-1·DUt - 0.396LPCGDPt-2 +0.474LPCECt-1 –

0.372LPCECt-2 + ut (19)

For 1976-1993 period:

LPCECt = – 7.624 + 0.0072T + 0.096LPCGDPt-1 – 0.396LPCGDPt-2 +

0.474LPCECt-1 – 0.372LPCECt-2 + ut (19a)

For 1994-2010 period:

LPCECt = – 2.667 + 0.0072T + 0.878LPCGDPt-1 – 0.396LPCGDPt-2 +

0.474LPCECt-1 – 0.372LPCECt-2 + ut (19b)

Two important observations can be made from equation (17). First, there is a short-run (week)
exogeneity as shown by the significant estimate of ∆LPCECt-1. Second, there is a long-run (strong)
exogeneity as shown by the significant estimate of the error correction term.

From the above equations we can also realize that the per capita energy consumption in a specific
year in the second period is strongly influenced by the per capita GDP and per capita energy
consumption of the previous year (positive sign). It seems that per capita GDP of the previous year has
a moderate effect on per capita energy consumption in the first period. The estimation results also
indicate that both per capita GDP and per capita energy consumption of two years ago have a negative
influence on the specific year energy consumption.

Table 9 presents the results for Granger causality tests from economic growth to energy
consumption, and vice versa. It is shown that the null hypothesis that energy consumption does not
Granger-cause economic growth in the short-run cannot be rejected. This frankly refuses justifications
of energy-led growth hypothesis in Vietnam. Incorporating with the cointegration analysis, we can
conclude that there is a strong Granger causality running from economic growth to energy
consumption. This fact shows that energy consumption is determined by the economic growth in
Vietnam. In other words, the conservation hypothesis is acceptable. Thus, energy conservation policy
will have little effect on economic growth.

Table 9: Granger causality tests
Dependent variable ∆LPCECt ∆LPCGDPt

Causality type Weak Strong Weak Strong

Test statistic
F-Stat.

12.05
(0.0016)

21.59
(0.000)

0.000
(0.999)

0.129
(0.879)

χ2-Stat.
12.05

(0.0005)
43.19

(0.000)
0.000

(0.999)
0.258

(0.878)

Excluded variables
∆LPCGDPt-1 ECT1,t-1,

∆LPCGDPt-1
∆LPCECt-1 ECT2,t-1,

∆LPCECt-1
Conclusion LPCGDP causes LPCEC LPCEC does not cause LPCGDP

Note: Numbers in (.) are p-values.

This research finding is consistent with the empirical study of Ozturk et al. (2010). We
approve with the way Ozturk et al. (2010) explain for this causality process. First, like other low
income countries, the recent economic growth in Vietnam has resulted in an expansion in commercial
and industrial sectors where energy is a fundamental input. Second, higher disposable income
increases demand for electronic devices for entertainment and comfort for households. This is also
consistent with the previous findings by Khanh Toan, P. (2009)1. He stated that energy might be

1 Khanh Toan, P., et al., (2009)

International Journal of Energy Economics and Policy, Vol. 1, No. 1, 2011, pp.1-17

mostly allocated into the residential uses (i.e. 39% residential, 36% industry, 20% transport in 2007;
and 60.3% residential, 27.3% industry, and 8.7% transport in 1990).

5. Conclusion

This article investigates the causal relationship between per capita energy consumption and per
capita GDP for Vietnam during the 1976-2010 period. In doing so, various cointegration testing
approaches are employed before estimating the vector error correction models. The empirical findings
suggest the existence of a uni-directional causality running from per capita GDP to per capita energy
consumption. In addition, it is also found that economic growth has a significant positive long-run
impact on energy consumption after the pointbreak, 1992. The research results strongly support the
neoclassical view that energy consumption is not a limiting factor for the Vietnam’s economic growth.
This in turn implies that the rise in energy prices can be a good opportunity for the economy to
promote substitution and technological innovation.

From a policy perspective, the results in this study are consistent with the conservation hypothesis.
Since a high level of economic growth leads to a high level of energy demand, but not vice versa, the
government can pursue the conservation energy policies that aim at curtailing energy use for
environmental friendly development purposes. We should gradually establish a competitive energy
market in order to allocate these resources into the most productive uses in the economy.

This study just focuses on the bivariate causality tests for the aggregated energy consumption and
economic growth, so it contains space for criticism. Further studies can be done by using either
multivariate models for total energy use or bivariate models for disaggregated energy consumption in
industrial, residential, and transport sectors.

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3.2 Cointegration Analysis