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An Analysis of Bitcoin’s Price Dynamics

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This paper aims to enhance the understanding of which factors affect the price development of Bitcoin in order for investors to make sound investment decisions. Previous literature has covered only a small extent of the highly volatile period during the last months of 2017 and the beginning of 2018. To examine the potential price drivers, we use the Autoregressive Distributed Lag and Generalized Autoregressive Conditional Heteroscedasticity approach. Our study identifies the technological factor Hashrate as irrelevant for modeling Bitcoin price dynamics. This irrelevance is due to the underlying code that makes the supply of Bitcoins deterministic, and it stands in contrast to previous literature that has included Hashrate as a crucial independent variable. Moreover, the empirical findings indicate that the price of Bitcoin is affected by returns on the S&P 500 and Google searches, showing consistency with results from previous literature. In contrast to previous literature, we find the CBOE volatility index (VIX), oil, gold, and Bitcoin transaction volume to be insignificant.
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Journal of
Risk and Financial
Management
Article
An Analysis of Bitcoin’s Price Dynamics
Frode Kjærland 1, 2, * , Aras Khazal 1, Erlend A. Krogstad 1, Frans B. G. Nordstrøm 1
and Are Oust 1
1NTNU Business School, Norwegian University of Science and Technology, 7491 Trondheim, Norway;
aras.kj@ntnu.no (A.K.); erlekrog@gmail.com (E.A.K.); fransbgn@gmail.com (F.B.G.N.);
are.oust@ntnu.no (A.O.)
2Nord University Business School, Nord University, 8049 Bodø, Norway
*Correspondence: frode.kjarland@ntnu.no
Received: 20 September 2018; Accepted: 11 October 2018; Published: 15 October 2018
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Abstract:
This paper aims to enhance the understanding of which factors affect the price development
of Bitcoin in order for investors to make sound investment decisions. Previous literature has covered
only a small extent of the highly volatile period during the last months of 2017 and the beginning
of 2018. To examine the potential price drivers, we use the Autoregressive Distributed Lag and
Generalized Autoregressive Conditional Heteroscedasticity approach. Our study identifies the
technological factor Hashrate as irrelevant for modeling Bitcoin price dynamics. This irrelevance is
due to the underlying code that makes the supply of Bitcoins deterministic, and it stands in contrast
to previous literature that has included Hashrate as a crucial independent variable. Moreover,
the empirical findings indicate that the price of Bitcoin is affected by returns on the S&P 500 and
Google searches, showing consistency with results from previous literature. In contrast to previous
literature, we find the CBOE volatility index (VIX), oil, gold, and Bitcoin transaction volume to
be insignificant.
Keywords: Bitcoin; cryptocurrency; Hashrate
JEL Classification: C10; G15
1. Introduction
The purpose of this study is to identify the factors that have an impact on the price of Bitcoin.
The market value of Bitcoin has grown tremendously in 2017. As the market values of cryptocurrencies
grow, it is reasonable to assume that they will start having an effect on certain economies. By estimating
the price drivers during the period ranging from 2013 to 2018, this paper will assist investors in
making sound investment decisions and aid in the understanding of what drives this phenomenon’s
price fluctuations.
Cryptocurrencies are decentralized digital currencies that use encryption to verify transactions.
In 2008, Nakamoto (2008) released his paper describing Bitcoin. In January of the following year,
Nakamoto released the software that launched the Bitcoin network. As of 2018, Bitcoin is the most
commonly known and used cryptocurrency. Since its founding in 2009, the price of Bitcoin has risen
from USD 0.07 to an all-time high of USD 20,089 on 17 December 2017 (Quandl.com). At this point in
time, its market capitalization was approximately USD 336.4 Billion.
From January 2017 through December, Bitcoin increased by 1270%, and the total cryptocurrency
trading volume passed USD 5 billion a day. Interest from the mainstream media, regulators, and the
public and financial markets accelerated so much that some call this period Bitcoin’s “IPO moment”
(Forbes.com 2017). During 2017, Bitcoin garnered more focus from institutional money, hedge funds,
and public funds. Its success culminated with the approval and introduction of Bitcoin derivatives.
J. Risk Financial Manag. 2018,11, 63; doi:10.3390/jrfm11040063 www.mdpi.com/journal/jrfm
J. Risk Financial Manag. 2018,11, 63 2 of 18
Due to the exponential rise in attention, we have included two sub-periods to test if the factors have
been the same before and after 2017.
We believe that it is important to understand the underlying factors affecting the price of
such a highly volatile financial phenomenon. Just as the price of Bitcoins has had an exponential
rise in the past year, the academic literature on Bitcoin and cryptocurrencies has experienced a
similar increase. Previous literature has used macro-economic, technological, and publicity factors in
Bitcoin models (Aalborg et al. 2018;Bouoiyour and Selmi 2016;Ciaian et al. 2016;Garcia et al. 2014;
Kristoufek 2013
,
2015
;
Kjærland et al. 2018
). However, few academic studies include data that reflect
the price fluctuations that Bitcoin experienced in 2017 and 2018. This paper addresses this gap in the
literature by assessing what variables drive the price of Bitcoin. As Kristoufek (2015) noted, “because of
the dynamic nature of Bitcoin and its rapid price fluctuations, it is logical that the drivers behind the price will
vary over time.” Therefore, we have chosen to analyze the drivers yet again.
To estimate the short- and long-term effects of potential price drivers on Bitcoin, an Autoregressive
Distributed Lag (ARDL) and Generalized Autoregressive Conditional Heteroscedasticity (GARCH)
model is estimated. We find Hashrate to be an irrelevant variable due to the deterministic feature of
the Bitcoin supply. The supply of Bitcoins are not dependent on price, as a normal good, but instead
the supply of Bitcoins are given by the Bitcoin code and solely dependent on time. Consistent with the
previous literature, we find that the S&P 500, Google searches and last week’s return on Bitcoin to be
significant explanatory variables, while gold, oil, CBOE volatility index (VIX), and Bitcoin transaction
volume are found to be insignificant in the estimation period.
This paper is organized as follows. Section 2contains a literature review, Section 3includes a
description of the data and econometric methods, Section 4presents the results, and Section 5includes
a discussion of the results and provides the conclusions.
2. Background and Literature Review
2.1. Introduction to Cryptocurrencies and Bitcoin
Several studies focus on the key concepts of cryptocurrencies and particularly
Bitcoin (Becker et al. 2013
;Brandvold et al. 2015;Dwyer 2015;Nica et al. 2017;Segendorf 2014).
According to
Dwyer (2015)
, the major innovation in Bitcoin is its decentralized technology. Instead of
storing transactions on a single or set of servers, the database is distributed across a network of
participating computers (Böhme et al. 2015). This database is what is called a Blockchain. Blocks are
added to the chain in the process of mining Bitcoins. The process of mining revolves around solving
complex computational puzzles, and the incentive for miners to participate are transaction fees and
Bitcoin rewards. To solve these puzzles, miners need computational power, which is measured
by the Hashrate. The Hashrate is the speed at which a computer can complete an operation in
the Bitcoin code, while the mining difficulty refers to the level of complexity in the computational
puzzles and is directly correlated with the Hashrate. As the Hashrate, either increases or decreases,
the underlying Bitcoin algorithm adjusts the mining difficulty so that the supply of Bitcoins follows
a predetermined path.
1
New coins are generated approximately every 10 min independent of the
current price, meaning that the Bitcoin supply is inelastic and time-dependent, as shown in Figure 1.
Since the supply is solely dependent on time, we choose to classify the Bitcoin supply as deterministic.
1
Bitcoin rewards are currently at 12.5 coins per block, but the protocol requires that the reward is halved every 210,000 mined
blocks. Mining 210,000 blocks takes approximately four years. Given the current level of network processing power, the next
halving will take place around early June 2020, bringing the mining reward down to 6.25 coins.
J. Risk Financial Manag. 2018,11, 63 3 of 18
J. Risk Financial Manag. 2018, 6, x FOR PEER REVIEW 3 of 18
Figure 1. Bitcoin deterministic supply.
2.2. Literature Review
Several authors have attempted to describe Bitcoin as a currency, stock, or asset. Yermack (2013)
argues that Bitcoin appears to behave more similar to a speculative store of value rather than a
currency. Dwyer (2015), on the other hand, describes Bitcoin as an electronic currency that can be
used to trade and store in a personal balance sheet. Dwyer’s argument is supported by Polasik et al.
(2015), who adds that Bitcoin can operate as a medium of exchange alongside other payment
technologies.
An increasing number of researchers have focused on the existence of a fundamental value of
Bitcoin, and some have studied whether or not it is a bubble. Garcia et al. (2014) finds that Bitcoin is
a financial bubble because of the difference between the exchange rate and fundamental value of
Bitcoin. He argues for a fundamental value given the cost of mining. Similarly, Hayes (2015, 2018)
proposed a specific cost of production model for valuating Bitcoin. Additionally, Cheah and Fry
(2015) conclude that Bitcoin is a speculative bubble and that the fundamental value of Bitcoin is zero.
Unlike earlier studies, Corbet et al. (2017) found that there is no clear evidence of a bubble in Bitcoin.
While these authors discuss if Bitcoin is a bubble or not, Bouri et al. (2017b) found that Bitcoin could
be used as an effective diversifier and, in some periods, also display safe-haven and hedge properties.
Some studies have been dedicated to determining the factors that drive the price of Bitcoin.
Bouoiyour and Selmi (2015) argue that long-term fundamentals are likely to be major contributors to
Bitcoin price variations. Among others, they also found technical factors to be a positive driver of
Bitcoin prices (Bouoiyour and Selmi 2015; Ciaian et al. 2016; Garcia et al. 2014; Georgoula et al. 2015;
Hayes 2015; Kristoufek 2015). Specifically, Georgoula et al. (2015) and Hayes (2015) found the
technical factor Hashrate to be a significant positive price driver. Bouoiyour and Selmi (2016), Garcia
et al. (2014), Kristoufek (2015), Kjærland et al. (2018), and Sovbetov (2018) have all used Hashrate as
a variable in their respective models.
Other scholars also argue for the significance of fundamental factors such as exchange-trade,
equity market indices, currency exchange rates, commodity prices, and transaction volume (Balcilar
et al. 2017; Bouri et al. 2018a; Bouoiyour and Selmi 2016; Bouoiyour et al. 2016; Ciaian et al. 2016;
Dyhrberg 2016; Kristoufek 2013; Yermack 2013). In contrast to Bouoiyour and Selmi (2015), Polasik
et al. (2015) states that an increase in the transaction volume will lead to higher prices and that global
economic factors do not seem to be an important driver. Ciaian et al. (2016) also found that supply
and demand factors have strong impacts on price and that standard economic currency models can
partly explain price fluctuations.
Kristoufek (2013, 2015) analyzed the frequency of online searches on Bitcoin, found them to be
a good proxy for interest and popularity, and discovered that the relationship between the price of
Bitcoin and online popularity is bidirectional. Ciaian et al. (2016) also found a positive relationship
between Wikipedia searches and Bitcoin. Others argue along the same lines as Kristoufek in that it is
primarily popularity and investor attractiveness that drive price movements (Bouoiyour et al. 2016,
Mai et al. 2015).
0
3,000,000
6,000,000
9,000,000
12,000,000
15,000,000
18,000,000
21,000,000
24,000,000
2009 2016 2023 2030 2037 2044 2051 2058
Figure 1. Bitcoin deterministic supply.
2.2. Literature Review
Several authors have attempted to describe Bitcoin as a currency, stock, or asset. Yermack (2013)
argues that Bitcoin appears to behave more similar to a speculative store of value rather than a currency.
Dwyer (2015), on the other hand, describes Bitcoin as an electronic currency that can be used to trade
and store in a personal balance sheet. Dwyer ’s argument is supported by Polasik et al. (2015), who adds
that Bitcoin can operate as a medium of exchange alongside other payment technologies.
An increasing number of researchers have focused on the existence of a fundamental value of
Bitcoin, and some have studied whether or not it is a bubble. Garcia et al. (2014) finds that Bitcoin is a
financial bubble because of the difference between the exchange rate and fundamental value of Bitcoin.
He argues for a fundamental value given the cost of mining. Similarly, Hayes (2015,2018) proposed a
specific cost of production model for valuating Bitcoin. Additionally, Cheah and Fry (2015) conclude
that Bitcoin is a speculative bubble and that the fundamental value of Bitcoin is zero. Unlike earlier
studies, Corbet et al. (2017) found that there is no clear evidence of a bubble in Bitcoin. While these
authors discuss if Bitcoin is a bubble or not, Bouri et al. (2017b) found that Bitcoin could be used as an
effective diversifier and, in some periods, also display safe-haven and hedge properties.
Some studies have been dedicated to determining the factors that drive the price of Bitcoin.
Bouoiyour and Selmi (2015) argue that long-term fundamentals are likely to be major contributors
to Bitcoin price variations. Among others, they also found technical factors to be a positive driver of
Bitcoin prices (Bouoiyour and Selmi 2015;Ciaian et al. 2016;Garcia et al. 2014;
Georgoula et al. 2015
;
Hayes 2015
;
Kristoufek 2015
). Specifically, Georgoula et al. (2015) and Hayes (2015) found the technical
factor Hashrate to be a significant positive price driver. Bouoiyour and Selmi (2016), Garcia et al. (2014),
Kristoufek (2015), Kjærland et al. (2018), and Sovbetov (2018) have all used Hashrate as a variable in
their respective models.
Other scholars also argue for the significance of fundamental factors such as exchange-trade,
equity market indices, currency exchange rates, commodity prices, and transaction volume
(
Balcilar et al. 2017
;Bouri et al. 2018a;Bouoiyour and Selmi 2016;Bouoiyour et al. 2016;
Ciaian et al. 2016
;
Dyhrberg 2016
;Kristoufek 2013;Yermack 2013). In contrast to Bouoiyour and
Selmi (2015),
Polasik et al. (2015)
states that an increase in the transaction volume will lead to higher
prices and that global economic factors do not seem to be an important driver. Ciaian et al. (2016)
also found that supply and demand factors have strong impacts on price and that standard economic
currency models can partly explain price fluctuations.
Kristoufek (2013,2015) analyzed the frequency of online searches on Bitcoin, found them to be
a good proxy for interest and popularity, and discovered that the relationship between the price of
Bitcoin and online popularity is bidirectional. Ciaian et al. (2016) also found a positive relationship
between Wikipedia searches and Bitcoin. Others argue along the same lines as Kristoufek in that it is
primarily popularity and investor attractiveness that drive price movements (Bouoiyour et al. 2016).
J. Risk Financial Manag. 2018,11, 63 4 of 18
2.3. Theoretical Foundation
2.3.1. Stock Price Theories and Momentum Theory
Santoni (1987) considers two theories that potentially explain stock prices: the Efficient Market
Hypothesis and the Greater Fool theory. The efficient market hypothesis tells us that all relevant
information is contained in current stock prices and that prices only change when investors receive
new information about fundamentals (Fama 1976). If this theory holds, past price changes contain
no useful information about future price changes. The Greater Fool theory says that investors regard
fundamental information as irrelevant. An investor buys shares on the belief that some bigger fool will
buy them from him at a higher price in the future. This scheme is all about speculation and anticipation
of continuing price increases due only to the fact that it has increased in the past.
Momentum in the financial market is an empirically observed trend for rising asset prices
to rise further and that decreasing asset prices lead to further decreases. Momentum theory
shows that stocks with strong past performance will outperform stocks that have a weak past
performance
(Jegadeesh and Titman 1993,2001)
. This theory relies on short-term movements rather
than fundamentals. In financial theory, the cause of momentum is known to be cognitive bias and
investors behaving irrationally.2
2.3.2. Volatility
Global financial turmoil impacts economies, assets, and currencies around the world.
Financial turmoil also affects the market participants and their investment decisions. During periods
of crisis, investors are more inclined to redistribute their investments to assets that are considered
to be safe-havens, including currencies. A currency is considered a safe-haven asset if international
investors invest in it to minimize losses during periods of financial turmoil. Because of its impact on
the development of currency exchange rates, financial turmoil, measured in volatility, is important to
include in an exchange rate model.
While there is evidence of negative shocks to equities generating more volatility than positive
shocks (Glosten et al. 1993), Baur and McDermott (2012) found that the volatility of gold returns
reacts inversely to negative shocks. According to Baur and McDermott (2012), this volatility relation
is due to the safe-haven properties of gold. Investors interpret rising gold prices as an increase in
macroeconomic uncertainty. Rising uncertainty increases the volatility of gold prices. However,
a study by
Bouri et al. (2016)
find no evidence of an asymmetric return-volatility relation in the
Bitcoin market–which in contrast support a safe haven property of Bitcoin. On the other side,
Kjærland et al. (2018) have the opposite finding.
3. Research Design
3.1. Data
The dependent variable to be explained by the models is the exchange rate between Bitcoin and
the US dollar. The original data are daily spot rates for BTC/USD for the period between 1 January
2013 and 20 February 2018.
To avoid potential issues related to autocorrelation, the daily data are modified into weekly
averages. As Bitcoin is traded every day of the week, we filter the data so that only common
observations are used. Days when some of the variables have missing values have been removed.
The data are gathered from various sources on 21 February 2018. The dependent variable and
independent explanatory variables are summarized in Table 1. These are chosen based on previous
literature and what we believe affects the price of Bitcoin.
2Cognitive biases are errors in thinking that affect the decisions and judgments that people make.
J. Risk Financial Manag. 2018,11, 63 5 of 18
Table 1. Variable Overview.
Variable Description Source
BTC exchange rate between Bitcoin and the US Dollar Quandl
Hashrate the estimated number of giga hashes per second the Bitcoin
network is performing Quandl
Volume total output volume of Bitcoin Quandl
S&P 500 S&P 500 is an index of the 500 largest US listed Corporations Thomson Reuters Eikon
Gold Goldman Sachs Commodity Index Gold Thomson Reuters Eikon
Oil WTI Crude Oil Spot Price in USD per barrel Thomson Reuters Eikon
VIX implicit volatility of options on the S&P 500, a measure of the
expected market volatility the next 30 days Thomson Reuters Eikon
Google normalized weekly statistics on the search term “Bitcoin”,
corrected for trends Google Trend
In accordance with the previous literature, we have included the S&P 500 and CBOE Volatility
Index (VIX). The S&P 500 is a good indicator of how financial markets are doing, and the VIX is
intended to provide an instantaneous measure regarding how much the market believes that the S&P
500 will fluctuate in the next 30 days. By including these two variables, we consider both the numerical
returns and risks in the financial markets. Furthermore, we have included the prices of WTI Oil and
Gold in our model. Both are considered to be important global commodities whose prices have impacts
on almost all economies around the world. These variables are all weekly observations obtained from
Thomson Reuters.
To test if publicity and attention given to Bitcoin has an impact on price changes, we include
Google Trends. Google search data show normalized weekly statistics that are corrected for trends on
searches mentioning the term “Bitcoin” (Google Trends Help)
3
. We also test for traditional supply
and demand effects by including Bitcoin transaction volume as a variable in this study. Finally,
the technological factor Hashrate is included. Volume and Hashrate are weekly data obtained from
Quandl.com.
3.2. Descriptive Statistics
Table 2shows the descriptive statistics of the variables. Figures 29display the changes in selected
variables over the estimated period.
Table 2. Descriptive statistics.
Variable Obs. Mean Std. Dev. Min. Max.
BTC 267 1372.9 2836.016 13.47221 17,612.51
Hashrate 267 2,132,903 3,936,302 20.80583
2.26
×
10
7
Volume 267 238,664.5 85,023.81 73,429.4 558,364.4
S&P 500 267 2053.9 296.3 1462.5 2844.4
Gold 267 1270.8 114.3 1063.0 1685.2
Oil 267 66.7 24.7 28.5 108.9
VIX 267 14.4 3.6 9.3 31.5
Google 267 7.3 13.4 1 100
3
Google does not differentiate between the upper- and lowercase letters, meaning that searches made on “Bitcoin” or “bitcoin”
are considered the same.
J. Risk Financial Manag. 2018,11, 63 6 of 18
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Figure 2. Bitcoin Market Price (USD), Quandl.
Figure 3. Bitcoin Hashrate, Quandl.
Figure 4. Bitcoin Transaction Volume, Quandl.
Figure 5. S&P 500 Index, Thomson Reuters.
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Figure 2. Bitcoin Market Price (USD), Quandl.
Figure 3. Bitcoin Hashrate, Quandl.
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Figure 2. Bitcoin Market Price (USD), Quandl.
Figure 3. Bitcoin Hashrate, Quandl.
Figure 4. Bitcoin Transaction Volume, Quandl.
Figure 5. S&P 500 Index, Thomson Reuters.
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Figure 4. Bitcoin Transaction Volume, Quandl.
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Figure 2. Bitcoin Market Price (USD), Quandl.
Figure 3. Bitcoin Hashrate, Quandl.
Figure 4. Bitcoin Transaction Volume, Quandl.
Figure 5. S&P 500 Index, Thomson Reuters.
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Figure 5. S&P 500 Index, Thomson Reuters.
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Figure 6. Gold Index (USD), Thomson Reuters.
Figure 7. Crude Oil-WTI Spot, Thomson Reuters.
Figure 8. CBOE Volatility Index, Thomson Reuters.
Figure 9. Google Search “Bitcoin,” Google Trends.
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Figure 6. Gold Index (USD), Thomson Reuters.
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Figure 6. Gold Index (USD), Thomson Reuters.
Figure 7. Crude Oil-WTI Spot, Thomson Reuters.
Figure 8. CBOE Volatility Index, Thomson Reuters.
Figure 9. Google Search “Bitcoin,” Google Trends.
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Figure 7. Crude Oil-WTI Spot, Thomson Reuters.
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Figure 6. Gold Index (USD), Thomson Reuters.
Figure 7. Crude Oil-WTI Spot, Thomson Reuters.
Figure 8. CBOE Volatility Index, Thomson Reuters.
Figure 9. Google Search “Bitcoin,” Google Trends.
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Figure 8. CBOE Volatility Index, Thomson Reuters.
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Figure 6. Gold Index (USD), Thomson Reuters.
Figure 7. Crude Oil-WTI Spot, Thomson Reuters.
Figure 8. CBOE Volatility Index, Thomson Reuters.
Figure 9. Google Search “Bitcoin,” Google Trends.
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Figure 9. Google Search “Bitcoin,” Google Trends.
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3.3. Econometric Method
3.3.1. Autoregressive Distributed Lag Model
According to Im et al. (2003), the ARDL technique is used to estimate short- and long-term
relationships between a group of variables by including lags for both the dependent and independent
variables. The ARDL model is estimated using ordinary least squares (OLS), where the only difference
is the inclusion of lags. As long as the OLS assumptions are fulfilled, the ARDL approach will yield
consistent estimates. This procedure is also followed by Ciaian et al. (2016) and Bouri et al. (2018b).
To find the appropriate lag length for each of the underlying variables in the ARDL model,
we used the modified Akaike information criteria (AIC), since this criterion is known for having a
theoretical advantage over other information criteria (Enders 2009). The model with the lowest AIC
and highest R-squared is considered the best. We put in dummy variables for the minimum and
maximum observations, in order to tackle the outliers. Using these dummy variables in the regression
enhances the reliability of the model (Hansen 2001).
To test for stationarity in a single time series, we use an augmented Dickey–Fuller (ADF) test.
If the ADF test shows signs of non-stationarity, the variables can be transformed into first differences,
and the test is reapplied. To address possible structural breaks in the time series, we combine the ADF
test with a Zivot–Andrews (ZA) test. If structural breaks are identified, the ZA test can be used since it
takes structural breaks into account (Vogelsang and Perron 1998).
3.3.2. The Generalized Autoregressive Conditional Heteroscedasticity Model
Regarding the GARCH, in order to control for homoscedasticity, we test the unconditional
variance of the regression. Breaking this assumption means that the Gauss–Markov theorem does not
hold and that the OLS estimators are not BLUE. Even though the unconditional variance is stable,
the conditional variance may not be constant over time. Engle (1982) developed the Autoregressive
Conditional Heteroskedasticity (ARCH) model that recognizes the difference between unconditional
and conditional variance and lets the conditional variance change over time as a function of previous
periods’ error terms. This technique has the ability to capture the effect of volatility clustering, but it
requires a model with a relatively long lag structure, which makes estimation difficult. To make this
task easier, Bollerslev (1986) proposed the GARCH model that enables a reduction in the number of
parameters by imposing nonlinear restrictions. The GARCH model can predict unconditional variance
and requires fewer parameters. In a GARCH model, the most recent observations have greater impacts
on the predicted volatility.
3.4. Model Estimation
By using OLS, we present three ARDL models. The testing of the models has also been done
over different in-sample periods, from 2013, Week 1, to 2016, Week 52, and from 2017, Week 1, to 2018,
Week 7. These periods have been chosen to assess the potential changes in what variables affects the
price of Bitcoins. The extreme price development in 2017 is also the background for this choice.
Model 1:
lnBTCt=α+β1lnBTCt1+β2lnVolumet+β3lnSP500t+
n
p=1
β4lnOiltp
+
n
p=1
β5lnGoldtp+β6lnVIXt+
n
p=2
β7lnGoogletp+Trend +εt.
(1)
Model 2:
lnBTCt=α+β1lnBTCt1+β2lnVolumet+β3lnSP500t+
n
p=2
β5lnGoogletp
+Trend +εt.
(2)
J. Risk Financial Manag. 2018,11, 63 9 of 18
Model 3:
lnBTCt=α+β1lnBTCt1+β2lnHashratet+β3lnVolumet+β4lnSP500t
+
n
p=1
β5lnOiltp+
n
p=1
β6lnGoldtp+β7lnVIXt
+
n
p=2
β8lnGoogletp+Trend +εt.
(3)
A number of post-estimation tests were performed to consider if all the assumptions of OLS are
fulfilled. The data set contains of 267 observations. OLS prerequisites were handled by the logarithmic
transformation of the data. The post-estimation test results for both the ARDL- and GARCH model
can be found in Tables A3A5.4
4. Empirical Results
4.1. Main Model (Model 1)
Tables 3and 4present the results of the ARDL and GARCH models. The first model is our main
model that includes all variables, while the second model is a reduced version of Model 1 that includes
only the significant variables of Model 1 for both the ARDL and GARCH. Table 5presents the third
regression model that includes the variable Hashrate. In the following sections, we will present the
results from the main period 2013, Week 1, to 2018, Week 7.
Table 3. Results of ARDL & GARCH models (Model 1).
ARDL GARCH
Time Period (1) (2) (3) (1) (2) (3)
lnBTCt10.19
(2.23) **
0.222
(2.14) **
0.206
(1.04)
0.225
(5.43) ***
0.329
(6.44) ***
0.293
(4.98) ***
lnVolumet
0.042
(1.41)
0.027
(0.79)
0.134
(2.33) **
0.046
(2.61) ***
0.022
(1.26)
0.15
(0.62)
lnSP500t1.772
(2.16) **
2.55
(2.69) ***
1.707
(1.04)
1.038
(1.59)
1.272
(1.85) *
1.318
(1.90) *
lnOilt
0.072
(0.50)
0.075
(0.47)
0.141
(0.40)
0.001
(0.00)
0.021
(0.17)
0.005
(0.04)
lnOilt10.142
(0.95)
0.147
(0.87)
0.341
(0.77)
0.023
(0.18)
0.027
(0.22)
0.005
(0.04)
lnGoldt0.552
(1.06)
0.546
(0.92)
1.135
(1.08)
0.013
(0.06)
0.006
(0.003)
0.063
(0.27)
lnGoldt1
0.415
(0.62)
0384
(0.49)
0.337
(0.35)
0.049
(0.18)
0.068
(0.24)
0.048
(0.17)
lnV I Xt0.029
(0.34)
0.126
(1.34)
0.279
(1.92) *
0.008
(0.12)
0.039
(0.56)
0.186
(0.93)
lnGooglet0.109
(3.60) ***
0.102
(2.84) ***
0.140
(3.16) ***
0.045
(2.85) ***
0.030
(1.61)
0.022
(1.18)
lnGooglet10.105
(3.46) ***
0.093
(2.58) **
0.176
(4.04) ***
0.088
(4.27) ***
0.081
(4.34) ***
0.076
(3.86) ***
4
The following post-estimation tests have been conducted for OLS-assumptions: Ramsey RESET test, Durbin–Watson,
Variance Inflation Factors (VIF), and Adjusted Dickey–Fuller. For GARCH: Ljung Box Q-statistics.
J. Risk Financial Manag. 2018,11, 63 10 of 18
Table 3. Cont.
ARDL GARCH
Time Period (1) (2) (3) (1) (2) (3)
lnGooglet20.082
(2.18) **
0.088
(1.98) **
0.022
(0.39)
0.053
(2.76) ***
0.062
(3.35) ***
0.057
(3.00) ***
ARCH Effect 0.562
(3.52) ***
0.771
(3.42) ***
0.599
(3.62) ***
GARCH Effect 0.315
(2.42) **
0.214
(1.61)
0.374
(3.00) ***
Adjusted R20.29 0.23 0.54
Observations 264 205 56 264 205 56
Note: * p< 0.10, ** p< 0.05, *** p< 0.01. (1) = 2013w1–2018w7, (2) = 2013w1–2016w25, and (3) = 2017w1–2018w7.
Table 4. Results of ARDL & GARCH models (Model 2).
ARDL GARCH
Time Period (1) (2) (3) (1) (2) (3)
lnBTCt10.187
(2.05) **
0.226
(2.05) **
0.065
(0.46)
0.215
(5.24) ***
0.318
(6.05) ***
0.293
(5.01) ***
lnSP500t1.411
(3.45) ***
1.364
(2.99) ***
1.59
(1.62)
0.926
(2.76) ***
0.873
(2.62) ***
0.779
(2.27) **
lnGooglet0.105
(3.50) ***
0.099
(2.79) ***
0.100
(1.82) *
0.033
(2.43) **
0.023
(1.5)
0.09
(0.99)
lnGooglet10.097
(3.18) ***
0.089
(2.43) **
0.122
(2.82)
0.083
(4.66) ***
0.08
(4.68) ***
0.075
(3.91) ***
lnGooglet20.077
(2.01) **
0.084
(1.88) *
0.061
(1.06)
0.047
(2.63) ***
0.06
(3.35) ***
0.055
(2.84) ***
ARCH Effect 0.581
(3.58) ***
0.696
(3.59) ***
0.497
(3.59) ***
GARCH Effect 0.324
(2.64) ***
0.269
(2.05) **
0.426
(3.53) ***
Adjusted R20.29 0.23 0.50
Observations 264 205 56 264 205 56
Note: * p< 0.10, ** p< 0.05, *** p< 0.01. (1) = 2013w1–2018w7, (2) = 2013w1–2016w25, and (3) = 2017w1–2018w7.
Table 5. Results of ARDL and GARCH models including Hashrate (Model 3).
ARDL GARCH
Time Period (1) (2) (3) (1) (2) (3)
lnBTCt10.19
(2.25) **
0.222
(2.10) **
0.206
(1.66)
0.225
(5.28) ***
0.329
(6.28) ***
0.258
(3.89) ***
lnHashratet0.067
(0.96)
0.22
(0.26)
0.274
(2.51) ***
0.031
(0.50)
0.005
(0.08)
0.039
(0.57)
lnVolumet
0.041
(1.36)
0.027
(0.78)
0.139
(2.51) **
0.04
(2.64) ***
0.022
(1.27)
0.139
(0.34)
lnSP500t1.725
(2.11) **
2.532
(2.68) ***
1.952
(1.33)
1.023
(1.55)
1.274
(1.86) *
1.472
(1.98) **
lnOilt
0.069
(2.11) **
0.075
(0.47)
0.073
(0.22)
0.008
(0.06)
0.02
(0.16)
0.012
(0.10)
lnOilt10.137
(0.92)
0.145
(0.85)
0.324
(0.77)
0.019
(0.15)
0.028
(0.22)
0.067
(0.53)
lnGoldt0.53
(1.01)
0.537
(0.90)
0.918
(0.87)
0.03
(0.13)
0.004
(0.02)
0.061
(0.24)
J. Risk Financial Manag. 2018,11, 63 11 of 18
Table 5. Cont.
ARDL GARCH
Time Period (1) (2) (3) (1) (2) (3)
lnGoldt1
0.404
(0.60)
0.38
(0.49)
0.393
(0.41)
0.04
(0.15)
0.068
(0.23)
0.020
(0.08)
lnV I Xt0.023
(0.27)
0.123
(1.32)
0.294
(2.19) **
0.004
(0.07)
0.04
(0.57)
0.234
(1.17)
lnGooglet0.108
(3.58) ***
0.102
(2.84) ***
0.118
(2.53) **
0.046
(2.83) ***
0.03
(1.57)
0.021
(1.09)
lnGooglet10.104
(3.41) ***
0.093
(2.56) **
0.165
(3.96) ***
0.088
(4.68) ***
0.081
(4.32) ***
0.076
(3.41) ***
lnGooglet20.081
(2.17) **
0.088
(1.98) **
0.014
0.25
0.056
(2.87) ***
0.062
(3.35) ***
0.055
(2.87) ***
ARCH Effect 0.547
(3.64) ***
0.768
(3.41) ***
0.436
(3.72) ***
GARCH Effect 0.345
(2.76) ***
0.214
(1.6)
0.538
(5.43) ***
Adjusted R20.29 0.23 0.54
Observations 264 205 56 264 205 56
Note: * p< 0.10, ** p< 0.05, *** p< 0.01. (1) = 2013w1–2018w7, (2) = 2013w1–2016w25, and (3) = 2017w1–2018w7.
4.1.1. ARDL (1)
As shown in Table 3, the lag of Bitcoin seems to have a significant positive effect on the price of
Bitcoin at the 5% level. If last week’s return of Bitcoin is higher by 1%, it is estimated that the return of
Bitcoin this week will be higher by 0.19%.
The first difference of S&P 500 is significant at the 5% level and has a positive sign. When the S&P
500 increases by 1%, the price of Bitcoins increases by 1.77%. In contrast, VIX, Oil, Gold, and Volume
do not seem to have any significant impact on the price of Bitcoin in the estimated period.
The first difference in the Google Trends variable and its lag are significant at the 1% level.
The short-term effects show that, when Google trends increases by 1%, the Bitcoin price is expected
to increase by 0.11%. By including the lag of Google, the short-term effect that Google search has
on Bitcoin price is 0.22%. Additionally, by including the second lag of Google, which is significant
at the 5% level, the total short-term effect that Google search has on Bitcoin price is 0.30%. Lastly,
the long-term effect of Google trends on Bitcoin price is 0.37%.5
4.1.2. GARCH (1)
In the GARCH model, the lag of Bitcoin has an almost identical effect as in the ARDL model and
is significant at the 1% level. Google and its two lags are significant at the 1% level, which is almost the
same as in the ARDL model, although the coefficient for both the first difference and the two lags has
decreased. Furthermore, the S&P 500 is found to be insignificant, while it was found significant in the
ARDL model. Similar to the ARDL model, VIX, oil, and gold are insignificant. Volume is significant at
the 1% level, which is inconsistent with the ARDL model.
The ARCH effects are positive and significant at the 1% level, which indicates that a shock in the
variance two weeks ago will have an impact of approximately 56.2% on the volatility in the following
week. The GARCH effects are significant at the 5% level. This significance indicates that 31.5% of the
volatility last week has an impact on volatility this week. The sum of the ARCH and GARCH effects
is approximately 87.7%, which shows the persistence of all volatility and shocks last week, and the
impact it has on this week.
5The long-term effect of a variable is calculated in following way: ßt+ ßt1+ ... + ßtn/(1 ß1lnBTCt1).
J. Risk Financial Manag. 2018,11, 63 12 of 18
4.2. Reduced Model (Model 2)
4.2.1. ARDL (1)
As shown in Table 4, the relationship between the lag and price of Bitcoin is almost the same as in
Model 1 and is significant at the 5% level. If the price of Bitcoin last week increased by 1%, the effect is
an increase in price this week of 0.19%. Moreover, the S&P 500 seems to have a significant impact on
the price of Bitcoin, similar to Model 1. This variable is significant at the 1% level. When the S&P 500
increases by 1%, Bitcoin is estimated to increase by 1.41%.
Google trends has the same significance level as Model 1 and almost equal coefficients.
The short-term effect of Google searches is 0.11%, and the total short-term effect is 0.21%. The total
long-term effect is 0.34%.
4.2.2. GARCH (1)
The lag of Bitcoin has an almost identical effect as in the ARDL model and is significant at the 1%
level. The S&P 500 index is also found to be significant in the GARCH model, just as the ARDL model,
but with a slightly lower coefficient.
Google and its two lags are significant at the 5% and 1% levels, respectively, which is almost
consistent with the ARDL model. However, the coefficient for both the first difference and the lags
has decreased.
The ARCH effects are positive and significant at the 1% level and has an impact of approximately
58.1% on the volatility in the following week. The GARCH effects are positive and significant at the
1% level. About a third (32.4%) of the volatility last week has an impact on the volatility this week.
The sum of the ARCH and GARCH effects is approximately 90.5%.
4.3. Model Including Hashrate (Model 3)
The model presented in Table 5includes the variable Hashrate but is otherwise similar to Model 1.
The properties displayed by the variables and their results are also similar to the results of Model 1.
However, the first difference of Hashrate has a positive sign in all the estimated periods but is only
significant in the third period, from 2017, Week 1, to 2018, Week 7.
4.4. Model Assessment
The weekly log-transformed ARDL models have adjusted R-square values of 29% and 31% for
Models 1 and 2, respectively. The ADF test for stationarity indicates that all the variables’ residuals
are stationary.
6
Other diagnostic tests are run to examine the models’ goodness of fit, and they are
fulfilled.
7
Lastly, to check for misspecification of the models, a Ramsey RESET test was performed.
This test indicates that the models may be misspecified.
5. Discussion and Conclusions
5.1. Discussion
In Model 3, which includes the Hashrate, we observe that the Hashrate has a positive sign in
both the estimated period and in-sample periods. The positive sign is contrary to the law of supply
and demand, considering that increasing the processing power should in theory lead to an increased
supply, which would exert a downward pressure on prices. Due to the deterministic supply of
Bitcoins, adding more processing capacity to mining will not lead to an increase in output. However,
this variable is only significant in the period from Week 1 of 2017 to Week 7 of 2018, a period of
6For a complete overview of the ADF and ZA tests, see Tables A1 and A2.
7For a complete overview of diagnostic tests for all models, see Tables A3A6.
J. Risk Financial Manag. 2018,11, 63 13 of 18
exponential growth in both Bitcoin and Hashrate. Therefore, we believe that the causality between
Bitcoin and Hashrate is such that it is the Bitcoin price that drives Hashrate, not the other way around.
This outcome is consistent with economic theory since an increase in price will naturally result in
the increased profitability of mining. As profitability increases, new actors will enter the mining
business, and current miners will increase computational power to the point where excess profits are
zero. A price drop will naturally lead to computational power being pulled out of Bitcoin mining.
Thus, we consider it irrelevant to include Hashrate as an explanatory variable in a model describing
Bitcoin’s price drivers or in calculations of fundamental values of Bitcoin. This outcome is in contrast
to previous research that included Hashrate as a variable (Bouoiyour and Selmi 2015;Garcia et al. 2014;
Georgoula et al. 2015;Hayes 2015;Kristoufek 2013,2015;Kjærland et al. 2018;Sovbetov 2018).
The results from the reported regression models indicate that publicity measured in Google
Trends has a positive impact on the price of Bitcoin. According to our findings, when people’s curiosity
and attention to Bitcoin increase, the demand for Bitcoins also increases. This outcome is consistent
with Kristoufek (2013,2015) and Ciaian et al. (2016), who found that when Google searches on Bitcoin
increase, the price of Bitcoin also increases.
We find that the S&P 500 has a positive impact on the price of Bitcoin. This is also the independent
variable with the largest coefficient, so it exerts the most influence on the price of Bitcoin in this
regression. The interpretation may be as follows: when optimism in financial markets increases,
investors also display optimism in Bitcoins. Since risk measured in standard deviation is higher in
Bitcoin than that in the S&P 500, Bitcoin investors are likely risk-seeking investors. These findings are
also supported by Yermack (2013) and Dyhrberg (2016) studies in which stock markets have an impact
on the price of Bitcoin. Interestingly, Bouri et al. (2018c) find moderate integration between Bitcoin
and most of the asset classes studied, included MSCI World and gold.
Our results indicate a positive relationship between the Bitcoin price and its lag, which indicates
that the efficient market hypothesis seemingly does not hold. Past returns should be uncorrelated with
present returns, and an investment strategy based on past returns should not be profitable. However,
it is known that the efficient market hypothesis is widely disputed. Some behavioral economists
blame imperfections in financial markets on errors in human reasoning and information processing.
Since most investors probably have limited experience with Bitcoins, the context around it is confusing,
and there is too much new information to consider in too little time; investors must make quick
decisions whether or not to invest. Thus, it is reasonable to assume that investors are affected by the
momentum effect of rising prices and vice versa. Observing the price increase last week fuels demand
and creates a momentum in price. Combined with Momentum theory, one can think along the lines of
the Greater Fool theory in which as the price rapidly increases, investors see get-rich-quick potential
by buying now and selling to a greater fool next week.
The estimated regression shows that fear in financial markets, as measured in VIX, does not have
a significant impact on the price of Bitcoin. However, in the sub-period between 2017 and 2018, we find
a significant negative relationship between VIX and Bitcoin price. During this period, the results
indicate that increasing fear of financial turmoil reduces demand for Bitcoins. Since a currency is
considered a safe-haven if demand rises during periods of financial stress, the abovementioned results
indicate that Bitcoin does not inhibit safe-haven properties, which is inconsistent with the findings of
Bouri et al. (2017a,2017b) and to some extent, with Bouri et al. (2016).
Additionally, both oil and gold were found to be insignificant in the estimated regression period.
These findings are in contrast to Kristoufek (2015) and Ciaian et al. (2016), who found that gold and
oil have significant positive impacts on Bitcoin prices. This outcome indicates that Bitcoin does not
inhibit commodity properties. In addition, the volatility in the price of Bitcoin is unlike any of the two
commodities, making it difficult to compare. However, our findings are much in line with the recent
study of Bouri et al. (2018b), who find no effects of an aggregate commodity index and gold prices on
the price of Bitcoin.
J. Risk Financial Manag. 2018,11, 63 14 of 18
Volume seems to have an insignificant impact on the price of Bitcoin in the estimated period,
reflecting Kristoufek (2013) findings, which state that the price of Bitcoin cannot be explained by
standard economic theory. However, in the GARCH model, volume seems to be a significant variable
with a negative sign. The reason may be our use of average daily prices or this outcome may be
explained by traditional economic theory regarding supply and demand. When volume increases
and demand is met, the price naturally drops, confirming the findings of Ciaian et al. (2016) and
Polasik et al. (2015) in which volume exerts an impact on the price of Bitcoin.
In the estimated GARCH model, we find that many of the included variables describe both the
return on Bitcoin and volatility. The results of the GARCH model show that the price of Bitcoin is
greatly affected by its own historical volatility. The results of the GARCH model are approximately the
same as those of the ARDL model, indicating that the ARDL model is robust. Similarly, we observe that
our model has approximately the same significant variables during an in-sample period, from Week 1
of 2013 to Week 52 of 2016, the period leading up to the volatile 2017. Although the in-sample period
from Week 1 of 2017 to Week 7 of 2018 exhibits different results, it is questionable whether these results
are reliable given the low number of observations, the high spike and subsequent fall in price during
the period.
5.2. Conclusions
Because of the increase in volatility and the dramatic price fluctuations in 2017, this paper
aims to help investors understand the price dynamics of Bitcoin. The results from the empirical
analysis provide compelling findings, and the estimated model has strong explanatory power with
a high degree of robustness. The primary contribution to Bitcoin research that this study provides
is the conclusion that the technological factor Hashrate should not be included in modeling price
dynamics or fundamental values since it does not affect Bitcoin supply. Based on our full and
reduced model, past price performance, optimism, and Google search volume all play significant
roles in explaining Bitcoin prices. When both optimism in financial markets and attention to Bitcoin
increase, investors’ willingness to allocate funds to more risky assets, such as Bitcoin, increases. Lastly,
we observe that price fluctuations in Bitcoin can be associated with investment theories such as The
Greater Fool and Momentum theory.
Appendix A
Table A1.
Results from Adjusted Dickey–Fuller test and Zivot–Andrews on log-transformed variables.
ADF-Test Zivot–Andrews
Variable Lag C, T t-Statistic Result Structural
Break Lag t-Statistic Result
BTC 4 C, T 2.095 I(1) 2016w26 2 3.983 I(1)
Hashrate
8 C, T 3.425 I(0) 2013w49 4 5.142 ** I(0)
Volume 10 C, T 1.991 I(1) 2014w38 1 6.059 *** I(0)
S&P 500
6 C, T 2.043 I(1) 2015w34 1 5.299 ** I(0)
Gold 2 C, T 2.974 I(1) 2016w4 2 4.748 I(1)
Oil 1 C, T 1.173 I(1) 2014w40 1 3.767 I(1)
VIX 15 C, T 2.119 I(1) 2015w34 0 6.133 *** I(0)
Google 1 C, T 2.481 I(1) 2016w25 0 4.709 I(1)
Note: ** p< 0.05, *** p< 0.01. All variables are in logarithmic form, C = Constant, T = trend, I(1) = unit root
(non-stationarity), and I(0) = no unit root (stationary). The Zivot–Andrews structural break is defined as the lowest
(most negative) t-statistic in the ADF test. Structural breaks are allowed for both the incline and the level of trend.
The Zivot–Andrews critical values are 1% (5.57), 5% (5.08), and 10% (4.82).
J. Risk Financial Manag. 2018,11, 63 15 of 18
Table A2.
Results from Adjusted Dickey–Fuller test and Zivot–Andrews on the first difference
log-transformed variables.
ADF-Test Zivot–Andrews
Variable Lag C, T t-Statistic Result Structural
Break Lag t-Statistic Result
BTC 2 C, T 6.899
***
I(0) 2013w50 2 6.670
***
I(0)
Hashrate
9 C, T 1.812 I(1) 2014w39 4 6.299
***
I(0)
Volume 15 C, T 5.105
***
I(0) 2014w6 1 11.382
***
I(0)
S&P 500
3 C, T 8.091
***
I(0) 2016w7 1 15.302
***
I(0)
Gold 4 C, T 7.399
***
I(0) 2014w12 2 12.356
***
I(0)
Oil 7 C, T 4.421
***
I(0) 2016w7 1 12.900
***
I(0)
VIX 15 C, T 5.267
***
I(0) 2017w17 0 14.305
***
I(0)
Google 1 C, T 12.17
***
I(0) 2013w48 0 18.273
***
I(0)
Note: *** p< 0.01. All variables are first difference on the logarithmic form; otherwise, see the note to Table A1.
Table A3. Model 1 assessment.
ARDL GARCH
Period (1) (2) (3) (1) (2) (3)
Outliers Yes Yes No Yes Yes No
Dummies Yes Yes No Yes Yes No
Observations 264 205 56 264 205 56
R20.32 0.27 0.65
Adjusted R20.29 0.23 0.54
AIC 483.83 359.30 117.63 545.39 432.77 548.31
Ramsey RESET, p-value 0.0000 0.0000 0.911
Durbin–Watson 2.07 2.13 2.01
Ljung-Box Q Stat 0.4265 0.5193 0.5088
ADF, residual value 0.0002 0.0017 0.0000 0.0000 0.0004 0.0000
Note: (1) = 2013w1–2018w7, (2) = 2013w1–2016w52, and (3) = 2017w1–2018w7.
Table A4. Model 2 assessment.
ARDL GARCH
Period (1) (2) (3) (1) (2) (3)
Outliers Yes Yes No Yes Yes No
Dummies Yes Yes No Yes Yes No
Observations 264 205 56 264 205 56
R20.31 0.25 0.56
Adjusted R20.29 0.23 0.50
AIC 489.41 366.10 117.35 552.80 442.78 554.30
Ramsey RESET, p-value 0.0000 0.0000 0.765
Durbin–Watson 2.08 2.14 1.85
Ljung-Box Q Stat 0.4155 0.5127 0.5041
ADF, residual value 0.0003 0.0024 0.0008 0.0000 0.0005 0.0000
Note: (1) = 2013w1–2018w7, (2) = 2013w1–2016w52, and (3) = 2017w1–2018w7.
J. Risk Financial Manag. 2018,11, 63 16 of 18
Table A5. Model 3 assessment.
ARDL GARCH
Period (1) (2) (3) (1) (2) (3)
Outliers Yes Yes No Yes Yes No
Dummies Yes Yes No Yes Yes No
Observations 264 205 56 264 205 56
R20.33 0.27 0.68
Adjusted R20.29 0.22 0.57
AIC 482.70 357.37 120.47 543.71 430.78 548.59
Ramsey RESET, p-value 0.0000 0.0000 0.817
Durbin–Watson 2.06 2.13 1.87
Ljung-Box Q Stat 0.4265 0.5187 0.5075
ADF, residual value 0.0001 0.0014 0.0004 0.0000 0.0005 0.0000
Note: (1) = 2013w1–2018w7, (2) = 2013w1–2016w52, and (3) = 2017w1–2018w7.
Table A6. Results of the variance inflation factors: test for autocorrelation.
Model 1 Model 2 Model 3
Variable VIF-Value Variable VIF-Value Variable VIF-Value
lnBTCt11.29 lnBTCt11.26 lnBTCt11.29
lnVolumet1.11 lnSP500t11.04 lnVolumet1.11
lnSP500t14lnGooglet1.09 lnSP500t14.01
lnOilt1.24 lnGooglet11.17 lnOilt1.24
lnGoldt1.15 lnGooglet21.16 lnGoldt1.16
lnV I Xt4.03 lnV IXt4.06
lnGooglet1.13 lnGooglet1.14
lnGooglet11.25 lnGooglet11.25
lnGooglet21.18 lnGooglet21.18
lnHashratet1.02
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2018 by the authors. Licensee MDPI, Basel, Switzerland. This article is an open access
article distributed under the terms and conditions of the Creative Commons Attribution
(CC BY) license (http://creativecommons.org/licenses/by/4.0/).
... However, these authors only consider 17 cryptocurrencies and they cannot fully confirm the hypothesis that the Bitcoin price has a stronger relationship with those cryptocurrencies in the long term. Some empirical studies (Ciaian et al., 2016;Goczek & Skliarov, 2019;Kjaerland et al., 2018) also observed a significant statistical relationship between cryptocurrencies's price and oil price without explaining the potential algorithmic reasons of this observation. 1 Despite the existence of studies on the link between financial uncertainty (financial markets' volatility) and crytpocurrencies dynamics, there is no research, to our knowledge, investigating the potential link between the latter and the economic policy uncertainty form an algorithmic perspective. ...
... 1 Despite the existence of studies on the link between financial uncertainty (financial markets' volatility) and crytpocurrencies dynamics, there is no research, to our knowledge, investigating the potential link between the latter and the economic policy uncertainty form an algorithmic perspective. Finally, few articles (Kjaerland et al., 2018) indicated the existence of a statistical interrelation between Bitcoin and S&P 500 but without comparing this observation with another cryptocurrency. Giudici et al. (2020) discussed the fundamental value of cryptocurrencies and the extent to which they should/ could be related to economic indicators whereas Ali et al. (2019) showed some links between the evolution of cryptocurrencies' price and the economic searches in search engines. ...
... Despite these debates, no scientific research has yet been done on this theme; this article being the first empirical studies on this matter. Indeed, the existing articles (Ciaian et al., 2016;Kjaerland et al., 2018;Goczek and Skliarov (2019) emphasizing a link between oil price and cryptocurrencies's one provided a statistical analysis of the dynamics of these factors without discussing the potential influence of the algorithms in the relationship-in doing so, these studies did not distinguish POW from POS-based cryptocurrencies. The integration of the algorithmic nature in the explanation of this statistical relationship is a contribution of our article that shows that POW-based cryptocurrencies have a stronger correlation with the oil prices. ...
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This article aims at investigating the extent to which the algorithmic nature (i.e., mining process) of cryptocurrencies might influence their dynamics and interaction with some major economic indicators. Our study observes that proof-of-stake based cryptocurrencies are less correlated with other crypto-assets offering more opportunities for diversifying portfolio strategy. We also observe a positive correlation between the proof-of-work based cryptocurrencies and the oil price. This article discusses these matters and suggests that the differences in cryptocurrencies' dynamics are more related to their service or purpose rather than their mining protocol. This claim contributes to the current debates on the intrinsic value of cryptocurrencies and it is illustrated with a discussion of the Stellar (XLM) and Ether (ETH) cases. Beyond our empirical results, our article suggests that, the liquidity and the returns dynamics of cryptocurrencies might be affected by two different aspects. Precisely, the former appears to be influenced by the economic service for which these cryptocurrencies are used, while cryptocurrencies' returns are more reactive to the way their cryptographic validation is operated. Our findings also suggest that an analysis through the economic service/purpose of cryptocurrencies is actually appropriate to understand their dynamics in relation to economic indicators. This perspective implicitly questions the monetary aspect often associated with cryptocurrencies and it calls for a more categorized research (by economic purpose) of cryptocurrencies whose potential intrinsic value would then be related to their economic purpose.
... Fantazzini & Kolodin (2020, p. 17), whose data run from August 1, 2015 to February 29, 2020, conclude, "there was neither evidence of Granger-causality nor cointegration in the first examined sample This outcome is consistent with economic theory since an increase in price will naturally result in the increased profitability of mining." (Kjaerland, et al., 2018). ...
... Exhibits 2 through 5 display, with appropriate scales for each period, the dollar price of a bitcoin versus the number of exahashes used to mine a bitcoin, which is a proxy for its cost. While both the price of a bitcoin and the number of exahashes to create one have increased over the years, a cursory examination of these exhibits demonstrates that there is little, if any, relationship between the two, and if anything, the hash rate follows the price dynamic rather than the reverse, as found by Kjaerland, et al. (2018), Fantazzini & Kolodin (2020), and Kristoufek (2020). In Period 1, the hash rate was slow to rise as the price of bitcoin rose to a peak. ...
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We examine the existence and dates of pricing bubbles in Bitcoin and Ethereum, two popular cryptocurrencies using the (Phillips et al., 2011) methodology. In contrast to previous papers, we examine the fundamental drivers of the price. Having derived ratios that are economically and computationally sensible, we use these variables to detect and datestamp bubbles. Our conclusion is that there are periods of clear bubble behaviour, with Bitcoin now almost certainly in a bubble phase.
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Prior studies on the price formation in the Bitcoin market consider the role of Bitcoin transactions at the conditional mean of the returns distribution. This study employs in contrast a non-parametric causality-inquantiles test to analyse the causal relation between trading volume and Bitcoin returns and volatility, over the whole of their respective conditional distributions. The nonparametric characteristics of our test control for misspecification due to nonlinearity and structural breaks, two features of our data that cover 19th December 2011 to 25th April 2016. The causality-in-quantiles test reveals that volume can predict returns – except in Bitcoin bear and bull market regimes. This result highlights the importance of modelling nonlinearity and accounting for the tail behaviour when analysing causal relationships between Bitcoin returns and trading volume. We show, however, that volume cannot help predict the volatility of Bitcoin returns at any point of the conditional distribution.
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We study the relationship between Bitcoin and commodities by assessing the ability of Bitcoin to act as a diversifier, hedge, or safe haven against daily movements in commodities in general, and energy commodities in particular. We focus on energy commodities because energy, in the form of electricity, is an essential input in the Bitcoin production. For the entire period, results show that Bitcoin is a strong hedge and a safe-haven against movements in both commodity indices. We further examine whether that ability is also present for non-energy commodities and our analysis show insignificant results when energy commodities are excluded from the general commodity index. We also account for the December 2013 Bitcoin price crash and our results reveal that Bitcoin hedge and safe-haven properties against commodities and energy commodities are only present in the pre-crash period, whereas in the post-crash period Bitcoin is no more than a diversifier. In addition to uncovering the time-varying role of Bitcoin, we highlight the dissimilarity in the dynamic correlations between the extreme downward and extreme upward movements.