A wireless passive sensor network is distinguished from ordinary wireless sensor networks by radio frequency (RF) sources that radiate RF waves and supply energy to sensor nodes. Against theoretical expectations, such a network suffers from a scarcity of energy, which compels to employ a contending-type medium access control (MAC) scheme. Consequently, a collision may take place among the packets sent by some sensor nodes. Also, the interference power brought by the colliding packets may hinder a sink node from identifying any packet. Apparently, the interference power highly influences the performance of a wireless passive sensor network. Thus, it is essential to appropriately describe the interference power to evaluate the performance. In this paper, we characterize the interference power, occurring in a wireless passive sensor network, as a sum of random variables governed by log-normal distributions, which are mutually independent but are not necessarily identical. Then, we prove the asymptotic normality of the interference power and, in pursuit of tractability rather than accuracy, we propose a normal approximation to the interference power brought by a finite number of sensor nodes. To investigate the accuracy of the proposed normal approximation, we take the Kolmogorov-Smirnov distance and present an exact expression of the Berry-Esséen bound associated with the interference power. Also, to explore the acceptability of the normal approximation, we employ three well-known statistical tests for normality. Comparative studies of Berry-Esséen bounds reveal that the normal approximation to the interference power is moderately accurate. Furthermore, hypothesis testing for normality shows that the normal approximation is acceptable even in the case that the interference power is incurred by a relatively small number of packets (or, equivalently, sensor nodes) in a wireless passive sensor network.

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## 1. Introduction

A wireless passive sensor network is a kind of wireless sensor network, which
consists of sink nodes, sensor nodes, and notably radio frequency (RF) sources ^{[1]}. In a wireless passive sensor network, an RF source radiates RF waves and supplies
energy to sensor nodes. In turn, a sensor node receives the RF waves and obtains energy.
Consuming the energy, the sensor node senses the environment, gathers information
and produces a packet. Then, the sensor node transmits the packet to a sink node.
Theoretically, an RF source is able to continuously supply abundant energy to sensor
nodes. In practice, however, an RF source is only permitted to transmit a limited
amount of power. Furthermore, the RF waves radiated by an RF source experiences a
heavy path loss en route to a sensor node. Also, the received RF waves are not efficiently
converted to direct currents at the sensor node. Against theoretical expectations,
a sensor node suffers from a scarcity of energy in a wireless passive sensor network
^{[2,}^{3]}.

A typical wireless passive sensor network consists of a single sink node and a
coexisting single RF source. Due to a scarcity of energy, the wireless passive sensor
network provides only limited coverage, where a relatively small number of sensor
nodes are geographically scattered around the sink node. In such a configuration,
two or more sensor nodes can simultaneously attempt to deliver their packets to the
sink node ^{[2,}^{3]}. Thus, a medium access control (MAC) scheme is needed for the sensor nodes to succeed
in packet delivery. For some reasons, e.g., scarcity of energy and difficulty in signaling,
the wireless passive sensor network opts to employ a contending-type MAC scheme, for
example, framed and slotted ALOHA ^{[4]}.

In a wireless passive sensor network governed by a contending-type MAC scheme,
two or more sensor nodes may send their packets to a sink node simultaneously. Then,
the packets collide ^{[5]} and the sink node may not be able to identify any of them. In a wireless passive
sensor network, sensor nodes are not usually equidistant from a sink node. Thus, when
some packets collide with each other, the sink node receives them with mutually different
power levels. If the power level of a packet is sufficiently higher than the interference
power, i.e., the power level of interfering packets, the sink node may be able to
identify the packet. Such a phenomenon, called capture ^{[5]}, makes a significant impact on the performance of a wireless passive sensor network.

As explained above, it is necessary to characterize the interference power to
evaluate the performance of a wireless passive sensor network. Then, the characteristics
of the channel from a sensor node to a sink node should be specified a priori since
a packet experiences path loss while traveling from a sensor node to a sink node.
There have been a number of reports on the channel characteristics in wireless sensor
networks ^{[6}-^{10]}. They considered a variety of environments; agricultural greenhouses, underground
pipes, and rice fields. However, most of them commonly assumed a log-distance path
loss model ^{[11]} and focused on calibrating the parameters involved in the log-distance path loss
model. The log-distance path loss model contains a term that expresses the effect
of shadowing, which is represented by a random variable governed by a log-normal distribution.
Then, the interference power in a wireless sensor network can also be represented
by a sum of log-normal random variables.

The exact distribution function of a sum of log-normal random variables is not
in a tractable form so that it is not easy to calculate the signal-to-interference
ratio (SIR), for example. Naturally, many efforts have been made to approximate a
sum of log-normal random variables. Such efforts are largely classified into two categories.
One is the category of approximations that focus on enhancing the accuracy ^{[12}-^{21]} and the other is the category of approximations which are more concerned with tractability
^{[22}-^{24]}. In ^{[22]}, Fenton considered a log-normal random variable, which has the same mean and variance
as a sum of log-normal random variables, and proposed it to approximate the sum. In
^{[23]}, Marlow showed that a sum of log-normal random variables asymptotically has a normal
distribution by using the Lindeberg-Feller central limit theorem ^{[25]}. Then, he suggested a normal approximation to a sum of log-normal random variables.
In ^{[24]}, Bkassiny considered multiple-input multiple-output free space optical systems and
introduced an Erlang approximation to a sum of log-normal random variables. As far
as we know, however, there has been no work on approximating the interference power,
which can be modeled as a sum of log-normal random variables, in the unique environment
of wireless passive sensor networks.

In this paper, we consider a wireless passive sensor network in which a number
of sensor nodes deliver their packets to a sink node according to a contending-type
MAC scheme. In the network, two or more sensor nodes may simultaneously send their
packets to a sink node, which incurs a collision among the packets. Then, the delivery
of a packet may be hampered by the other packets involved in the collision since the
power of the packet can be overwhelmed by the interference power, i.e., the power
brought by the interfering packets. The interference power highly influences the performance
of the wireless passive sensor network. Thus, it is of necessity to appropriately
model the interference power to evaluate the performance of the wireless passive sensor
network. First, we assume that a packet suffers a log-distance path loss ^{[11]} while traveling from a sensor node to a sink node. Then, we characterize the interference
power as a sum of log-normal random variables. Secondly, we prove the asymptotic normality
of the interference power by using the Lyapunov central limit theorem ^{[25]}. Based on the property of asymptotic normality, pursuing tractability rather than
accuracy, we propose a normal approximation to the interference power brought by a
finite number of sensor nodes. Thirdly, as a measure for the accuracy of normal approximation,
we take the Kolmogorov-Smirnov distance ^{[26]} between the exact and approximate distribution functions of the interference power.
Then, we develop an exact expression of the Berry-Esséen bound ^{[27}-^{29]}, which is in fact an upper bound on the Kolmogorov-Smirnov distance, associated with
the interference power. Fourthly, we employ three kinds of statistical tests for normality;
the Kolmogorov-Smirnov test ^{[30]}, the Cramér-von Mises test ^{[31,}^{32]} and the Anderson-Darling test ^{[33]}. Then, we evaluate the acceptability of the normal approximation to the interference
power.

The main contributions of this paper are as follows.

▪ First, we describe the power brought by interfering packets in a wireless passive sensor network employing a contending-type MAC scheme.

▪ Secondly, by using the Lyapunov central limit theorem, we prove the asymptotic normality of the interference power, i.e., the property that the interference power converges in distribution to a normal random variable as the number of interfering packets goes to infinity.

▪ Thirdly, we consider Kolmogorov-Smirnov distance between the distribution function of the interference power and the proposed normal distribution function. Then, we derive the exact Berry-Esséen bound associated with the interference power in a closed form. In addition, we develop an exact expression of the Berry-Esséen bound associated with the sum of normal, uniform, shifted exponential, or Pareto random variables which are mutually independent but are not necessarily distributed identically.

▪ Fourthly, we adopt three well-known statistical tests for normality; the Kolmogorov-Smirnov test, the Cramér-von Mises test and the Anderson-Darling test. Then, we investigate the acceptability of the proposed normal approximation to the interference power occurring in a wireless passive sensor network.

The rest of the paper is organized as follows. In Section 2, we characterize the interference power in a wireless sensor network ruled by a contending-type MAC scheme. In Section 3, we prove the asymptotic normality of the interference power and we also propose a normal approximation to the interference power. In Section 4, we derive the Berry-Esséen bound associated with the interference power and carry out a comparative evaluation of the accuracy of the normal approximation. Section 5 is devoted to statistical testing for the normality of the interference power, which provides evidence that the normal approximation to the interference power is acceptable. In Section 6, we consider log-normal and Erlang approximations besides the normal approximation. Then, we compare the approximate distribution functions with the empirical distribution function of the interference power. In Section 7, we suppose that a specific contending-type MAC scheme is employed by a wireless passive sensor network. Then, we investigate the accuracy of the throughput value calculated by using the normal approximation.

## 2. Interference Power

Consider a typical wireless passive sensor network which consists of a single sink node, a single RF source and $M$ sensor nodes, denoted by $s_{1},\cdots ,s_{M}$. As illustrated in Fig. 1, the RF source is positioned near the sink node while the sensor nodes are scattered around the sink node. Let $D_{m}$ denote the distance between the sink node and sensor node $s_{m}$ for $m\in \left\{1,\cdots ,M\right\}$. In practice, distance $D_{m}$ is bounded from above as well as from below, i.e., there exist $D_{U}$ and $D_{L}$ such that

for all $m\in \left\{1,\cdots ,M\right\}$.

Theoretically, a sink node is able to supply abundant energy to sensor nodes in
a wireless passive sensor network. Against theoretical expectations, however, a sensor
node suffers from a scarcity of energy in practice. As a result, a sensor node is
not able to directly power itself with the received energy but has to temporarily
store it at a capacitor, for example. Inevitably, a sensor node needs a time interval
for charging its capacitor prior to transmitting a packet. For wireless passive sensor
networks, a MAC scheme is thus devised in consideration of such a scarcity of energy
^{[2,}^{3]}. In the wireless passive sensor network, we suppose that time is divided into frames
and each frame is partitioned again into a charging interval and a transmitting interval.
(See Fig. 2) During a charging interval, the RF source radiates RF waves. Then, a sensor node
receives the RF waves, converts them to direct current (DC) and charges its internal
capacitor. As a transmitting interval begins, a sensor node senses the environment,
gathers information and encapsulates the data into a packet. During the transmitting
interval, the sensor node then transmits the packet to the sink node according to
a contending-type MAC scheme, e.g., framed and slotted ALOHA ^{[4]}.

Assume that all the sensor nodes are equipped with same energy harvester, capacitor
charger and packet transmitter. Suppose that every sensor node transmits a packet
with same power. Let $U$ denote the amount of power transmitted by a sensor node.
Assume that each charging interval is long enough so that even the farthest sensor
node from the sink node is able to charge its capacitor as sufficiently as it can
construct a packet and then transmit the packet with power $U$. Even if two sensor
nodes transmit their packets with same power, the sink node does not necessarily receive
these packets with same power since the packets experience different path losses ^{[5]}. There have been several research works on the channel characteristics in wireless
sensor networks ^{[6}-^{10]}. They were conducted in a wide variety of environments; agricultural greenhouses,
rice fields and suburbs. However, many of them focused on calibrating the parameters
involved in the log-distance path loss model ^{[11]}. Also, a wireless passive sensor network has some unique characteristics related
to the fading that happens in the signal from a sensor node to the sink node; a sensor
node transmits a relatively narrowband signal to the sink node, a sensor node has
almost no mobility and the distance between the sink node and a sensor node is relatively
short. Thus, frequency-selective fading and Doppler shift rarely take place in a wireless
passive sensor network. Furthermore, the multipath fading is not as significant as
the shadow fading in a wireless passive sensor network.

Let $V_{m}$ denote the power with which the sink node receives a packet from sensor node $s_{m}$ for $m\in \left\{1,\cdots ,M\right\}$. Then, based on the argument above, we set that received power $V_{m}$ conforms to the log-distance path loss model. Then, received power $V_{m}$ is related to transmitted power $U$ as follows.

in decibels and

in watts, where $K$ is the path loss constant, $D_{0}$ is the reference distance, $\gamma $ is the path loss exponent, and $Z_{m}$ is a random variable governed by a normal distribution with mean $0$ and variance $\omega ^{2}$. Also, assume that random variables $Z_{1},\cdots ,Z_{M}$ are mutually independent. Set

for $m\in \left\{1,\cdots ,M\right\}$. Then, $V_{m}$ can be rewritten by

for $m\in \left\{1,\cdots ,M\right\}$. Note that $Z_{m}$ has the normal distribution with mean $0$ and variance $\omega ^{2}$ so that $\alpha Z_{m}+\beta _{m}$ also has the normal distribution with mean $\beta _{m}$ and variance $\alpha ^{2}\omega ^{2}$. Thus, $V_{m}$ is governed by the log-normal distribution whose density

for $x\in \left(0,\infty \right)$. Also, the mean and variance of $V_{m}$ are as follows.

For

In conclusion, the interference power is expressed by the sum of log-normal random variables which are mutually independent but are not necessarily identical.

## 3. Asymptotic Normality of and Normal Approximation to Interference Power

In Section 2, the interference power $X_{n}$ is expressed by a sum of log-normal random variables. Unfortunately, a sum of log-normal random variables has neither distribution function nor density in a tractable form. To easily calculate the ratio of signal to interference and noise (SINR), for example, we suggest a normal approximation to the interference power. Also, we show the validity of the normal approximation by proving that interference power $X_{n}$ converges to a normal random variable in distribution.

As shown in (8), interference power $X_{n}$ is the sum of mutually independent random variables,
denoted by $V_{{\pi _{1}}},\cdots ,V_{{\pi _{M}}}$. However, they are not necessarily
distributed identically. Nonetheless, the Lyapunov central limit theorem ^{[25]} says that interference power $X_{n}$ converges to a normal random variable as far
as the Lyapunov condition ^{[25]} is satisfied. The following lemma shows that random variables $V_{{\pi _{1}}},V_{{\pi
_{2}}},\cdots $ satisfy the Lyapunov condition. Also, the theorem below confirms that
interference power $X_{n}$ consequently converges to a normal random variable in distribution
according to the Lyapunov central limit theorem.

$\textbf{Lemma 1}$ There exists $\delta >0$ such that

##### (9)

$\frac{\sum _{k=1}^{n}E\left(\left| V_{{\pi _{k}}}-E\left(V_{{\pi _{k}}}\right)\right| ^{2+\delta }\right)}{\left[\sqrt{\sum _{k=1}^{n}Var\left(V_{{\pi _{k}}}\right)}\right]^{2+\delta }}\rightarrow 0$as $n\rightarrow \infty $.

Proof: A proof is given in the appendix.

$\textbf{Theorem 1}$ Set

##### (10)

$\overset{˜}{X}_{n}=\frac{1}{\sqrt{\sum _{k=1}^{n}Var\left(V_{{\pi _{k}}}\right)}}\sum _{k=1}^{n}\left[V_{{\pi _{k}}}-E\left(V_{{\pi _{k}}}\right)\right].$Let $\overset{˜}{F}_{n}$ denote the distribution function of $\overset{˜}{X}_{n}$. Set $W$ to be a random variable having the standardized normal distribution. Let $\Phi$ denote the distribution function of $W$, i.e.,

for $y\in \mathrm{\mathbb{R}}$. Then,

as $n\rightarrow \infty $, i.e.,

for every $x\in \mathrm{\mathbb{R}}$.

Proof: A proof is given in the appendix.

From Theorem 1, interference power $X_{n}$ can be approximated by a random variable $Y_{n}$, which is

##### (14)

$Y_{n}=\sqrt{\sum _{k=1}^{n}Var\left(V_{{\pi _{k}}}\right)}W+\sum _{k=1}^{n}E\left(V_{{\pi _{k}}}\right)$for $n\in \mathrm{\mathbb{N}}$.

## 4. Berry-Esséen Bound Associated with Interference Power

This section is devoted to investigating the similarity between the interference
power $X_{n}$ defined in (8) and the normal random variable $Y_{n}$ defined in (14). A way of investigating the accuracy of an approximate distribution function of the
interference power is to examine the distance or metric between the exact and approximate
distribution functions. Many kinds of distances have been proposed, e.g., the total
variation distance ^{[34]}, the Wasserstein metric ^{[34]}, the Lévy metric ^{[34]}, and the Kolmogorov-Smirnov distance ^{[27]}. Among them, the Kolmogorov-Smirnov distance is known to have a considerable advantage
of computational simplicity ^{[35]}. Also, a useful bounding technique is available for the Kolmogorov-Smirnov distance
^{[34]}. Thus, we first adopt the Kolmogorov-Smirnov distance to measure the accuracy of
the normal approximation. Next, we derive an exact expression of the Berry-Esséen
bound ^{[27}-^{29]}, which is an upper bound on the Kolmogorov-Smirnov distance, associated with the
interference power. Finally, we carry out a comparative study of the Berry-Esséen
bound associated with the interference power.

### 4.1 Derivation of Berry-Esséen Bound Associated with Interference Power

Recall the normalized interference power, denoted by $\overset{˜}{X}_{n}$ in (10). Note that it has mean $0$ and variance $1$. Remind that $W$ denotes a random variable having the standardized normal distribution. Then, the Kolmogorov-Smirnov distance between $\overset{˜}{X}_{n}$ and $W$ is defined by}

##### (15)

$d_{KS}\left(\overset{˜}{X}_{n},W\right)=\sup \left\{\left| \overset{˜}{F}_{n}\left(x\right)-\Phi \left(x\right)\right| \colon x\in \mathrm{\mathbb{R}}\right\}$where $\overset{˜}{F}_{n}$ and $\Phi$ are distribution functions of $\overset{˜}{X}_{n}$ and $W$, respectively. Apparently, a small value of $d_{KS}\left(\overset{˜}{X}_{n},W\right)$ indicates strong similarity between $\overset{˜}{X}_{n}$ and $W$. Define

##### (16)

$b_{BE}\left(X_{n}\right)=C_{0}\frac{\sum _{k=1}^{n}E\left(|V_{{\pi _{k}}}-E\left(V_{{\pi _{k}}}\right)|^{3}\right)}{\left[\sum _{k=1}^{n}Var\left(V_{{\pi _{k}}}\right)\right]^{\frac{3}{2}}}$where $C_{0}$ is a positive constant. Then, $b_{BE}\left(X_{n}\right)$, which is the Berry-Esséen bound associated with interference power $X_{n}$, is an upper bound on the Kolmogorov-Smirnov distance between $\overset{˜}{X}_{n}$ and $W$, i.e.,

for all $n\in \mathrm{\mathbb{N}}$. Thus, it can be said that a small value of $b_{BE}\left(X_{n}\right)$ tends to indicate a short Kolmogorov-Smirnov distance and hence strong similarity between $\overset{˜}{X}_{n}$ and $W$. The following theorem exactly calculates the Berry-Esséen bound associated with interference power $X_{n}$.

$\textbf{Theorem 2}$ The Berry-Esséen bound associated with $X_{n}$##### (18)

$$ \begin{aligned} b_{B E}\left(X_{n}\right)=& \frac{C_{0} \sum_{k=1}^{n} e^{3 \beta_{\pi_{k}}}}{e^{\alpha^{3} \omega^{3}}\left[e^{\alpha^{2} \omega^{2}}-1\right]^{\frac{3}{2}}\left[\sum_{k=1}^{n} e^{2 \beta_{\pi_{k}}}\right]^{\frac{3}{2}}} \\ & \times\left[e^{\frac{9 \alpha^{2} \omega^{2}}{2}}\left[2 \Phi\left(\frac{5 \alpha \omega}{2}\right)-1\right]\right.\\ &-3 e^{\frac{5 \alpha^{2} \omega^{2}}{2}}\left[2 \Phi\left(\frac{3 \alpha \omega}{2}\right)-1\right] \\ &+4 e^{\frac{3 \alpha^{2} \omega^{2}}{2}}\left[2 \Phi\left(\frac{\alpha \omega}{2}\right)-1\right] \end{aligned} $$Proof: A proof is given in the appendix.

### 4.2 Evaluation of Berry-Esséen Bound Associated with Interference Power

For a comparative evaluation of the Berry-Esséen bound associated with the interference power, we introduce a random variable $S_{n}$, which is a sum of random variables as follows.}

where random variables $T_{1},\cdots ,T_{n}$ are not necessarily distributed identically but are mutually independent. For all $k\in \left\{1,\cdots ,n\right\}$, we then consider the four cases below. In Case 1, $T_{k}$ has the normal distribution with mean $\mu _{k}$ and variance $\sigma _{k}^{2}$, i.e., the density of $T_{k}$

##### (20)

$g_{k}\left(x\right)=\frac{1}{\sqrt{2\pi \sigma _{k}^{2}}}e^{-\frac{1}{2\sigma _{k}^{2}}{\left(x-\mu _{k}\right)^{2}}}$for $x\in \mathrm{\mathbb{R}}$. In Case 2, $T_{k}$ has the uniform distribution in $\left[a_{k},b_{k}\right]$, i.e., the density of $T_{k}$

for $x\in \left(a_{k},b_{k}\right)$. In Case 3, $T_{k}$ has the exponential distribution with scale parameter $\delta _{k}$ and rate parameter $\lambda _{k}$, i.e., the density of $T_{k}$

for $x\in \left(\delta _{k},\infty \right)$. In Case 4, $T_{k}$ has the Pareto distribution with scale parameter $\delta _{k}$ and shape parameter $\nu _{k}$, i.e., the density of $T_{k}$

for $x\in \left(\delta _{k},\infty \right)$. The theorem below shows the Berry-Esséen bounds associated with $S_{n}$ in these four cases.

$\textbf{Theorem 3}$ In the four cases above, the Berry-Esséen bounds associated with $S_{n}$ are as follows.If $T_{k}$ has the normal distribution with mean $\mu _{k}$ and variance $\sigma _{k}^{2}$,

##### (24)

$b_{BE}\left(S_{n}\right)=\frac{2\sqrt{2}C_{0}\sum _{k=1}^{n}\sigma _{k}^{3}}{\sqrt{\pi }\left[\sum _{k=1}^{n}\sigma _{k}^{2}\right]^{\frac{3}{2}}}.$If $T_{k}$ has the uniform distribution in $\left[a_{k},b_{k}\right]$,

##### (25)

$b_{BE}\left(S_{n}\right)=\frac{3\sqrt{3}C_{0}\sum _{k=1}^{n}\left(b_{k}-a_{k}\right)^{3}}{4\left[\sum _{k=1}^{n}\left(b_{k}-a_{k}\right)^{2}\right]^{\frac{3}{2}}}.$If $T_{k}$ has the exponential distribution with scale parameter $\delta _{k}$ and rate parameter $\lambda _{k}$,

##### (26)

$b_{BE}\left(S_{n}\right)=\frac{2\left(6e^{-1}-1\right)C_{0}\sum _{k=1}^{n}\frac{1}{\lambda _{k}^{3}}}{\left[\sum _{k=1}^{n}\frac{1}{\lambda _{k}^{2}}\right]^{\frac{3}{2}}}.$If $T_{k}$ has the Pareto distribution with scale parameter $\delta _{k}$ and shape parameter $\nu _{k}$,

##### (27)

$$ \begin{aligned} b_{B E}\left(S_{n}\right)=& \frac{C_{0}}{\left[\sum_{k=1}^{n} \frac{\delta_{k}^{2} v_{k}}{\left(v_{k}-1\right)\left(v_{k}-2\right)}\right]^{\frac{3}{2}}} \\ & \times \sum_{k=1}^{n}\left[\frac{12 \delta_{k}^{3}}{\left(v_{k}-1\right)\left(v_{k}-2\right)\left(v_{k}-3\right)}\left(1-\frac{1}{v_{k}}\right)^{v_{k}-3}\right.\\ &\left.-\frac{2 \delta_{k}^{3} v_{k}\left(v_{k}+1\right)}{\left(v_{k}-1\right)^{3}\left(v_{k}-2\right)\left(v_{k}-3\right)}\right] \end{aligned} $$Proof: A proof is given in the appendix.

Using the formulae in theorems 2 and 3, we carry out a numerical comparison of $b_{BE}\left(X_{n}\right)$ with $b_{BE}\left(S_{n}\right)$. In the comparison, we suppose that the wireless passive sensor network is configured as a honeycomb. As illustrated in Fig. 3, the coverage of the network consists of 19 disks. The boundary of each disk then circumscribes one of the non-overlapping 19 hexagons. A sink node lies on the disk which is at the center of the coverage. On the other hand, a sensor node resides in each of the 18 disks except the central disk. Besides the network configuration, various physical parameters are involved in the numerical comparison. Their values are summarized in Table 1.

With respect to the number of interfering packets, Fig. 4 shows the average Berry-Esséen bound, which is yielded under some assumptions below.
Since the wireless passive sensor network consists of 18 sensor nodes and each sensor
node transmits at most one packet during a transmitting interval, the number of interfering
packets can be a number in $\left\{1,\cdots ,17\right\}$. Suppose that the number
of interfering packets is $n\in \left\{1,\cdots ,17\right\}$. Then, the sensor nodes
that sent such interfering packets form an $\textit{n}$-combination of $\left\{s_{1},\cdots
,s_{18}\right\}$. Note that the number of $\textit{n}$-combinations is equal to $\left(\begin{array}{l}18\\n\end{array}\right)$.
In Fig. 4, we first assume that the set of the sensor nodes whose packets create interference
is an $\textit{n}$-combination of $\left\{s_{1},\cdots ,s_{18}\right\}$ equally likely,
i.e., with probability $\frac{1}{\left(\begin{array}{l}18\\n\end{array}\right)}$.
Then, we exhibit the expected Berry-Esséen bound with respect to the set of interfering
sensor nodes. In this figure, $S_{n}$ is a sum of normal, uniform, exponential, or
Pareto random variables, denoted by $T_{1},\cdots ,T_{n}$ as in (19). Secondly, we then assume that $T_{k}$ has the same mean and variance as $V_{{\pi
_{k}}}$ has for every $k\in \left\{1,\cdots ,n\right\}$. The exact value of positive
constant $C_{0}$ appearing in the Berry-Esséen bound has not been found up to now.
However, a number of upper bounds on the constant have been reported. Among them,
the smallest upper bound is known to be 0.56 ^{[36]}. Thirdly, we assume that constant $C_{0}$ is 0.56 in Fig. 4.

Table 1. Values of parameters.

In Fig. 4, we observe that the interference power incurs a higher Berry-Esséen bound than a sum of normal or uniform random variables. Note that a sum of normal random variables is also normally distributed. On the other hand, the interference power produces a lower Berry-Esséen bound than a sum of exponential or Pareto random variables. Such an observation reveals the tendency that the interference power is more similar to a normal random variable than a sum of exponential or Pareto random variables, wheares it is less similar than a sum of normal or uniform random variables. In addition, we notice that the average Berry-Esséen bound monotonically decreases as the number of interfering packets increases. This notice confirms that the interference power is apt to converge to a normal random variable as more packets create interference.

## 5. Test for Normality of Interference Power

In this section, we statistically test the normality of the interference power.
Specifically, we employ three kinds of well-known tests; the Kolmogorov-Smirnov test
^{[30]}, the Cramér-von Mises test ^{[31,}^{32]} and the Anderson-Darling test ^{[33]}.}

Recall that $\overset{˜}{X}_{n}$ and $W$ represent the normalized interference power and a standardized normal random variable, respectively. Also, note that $\overset{˜}{F}_{n}$ and $\Phi$ denote the distribution functions of $\overset{˜}{X}_{n}$ and $W$, respectively. To test the normality of the interference power, we construct null and alternative hypotheses, denoted by $\mathrm{H}_{0}$ and $\mathrm{H}_{1}$, as follows.

##### (28)

$\mathrm{H}_{0}\colon \overset{˜}{F}_{n}=\phi$ $\mathrm{H}_{1}\colon \overset{˜}{F}_{n}\neq \phi$
Then, we test the null hypothesis $\mathrm{H}_{0}$. For such a test, we adopt
the Kolmogorov-Smirnov test, the Cramér-von Mises test and the Anderson-Darling test,
which are commonly based on an empirical distribution function ^{[30]}. Suppose that we obtain $J$ samples of normalized interference power $\overset{˜}{X}_{n}$.
Let $\overset{˜}{X}_{n,1},\cdots ,\overset{˜}{X}_{n,J}$ denote the samples. Also,
let $\overset{˜}{X}_{n,\left(1\right)},\cdots ,\overset{˜}{X}_{n,\left(J\right)}$
denote the order statistics of $\overset{˜}{X}_{n,1},\cdots ,\overset{˜}{X}_{n,J}$
such that $\overset{˜}{X}_{n,\left(1\right)}\leq \cdots \leq \overset{˜}{X}_{n,\left(J\right)}$
almost surely. Then, an empirical distribution function of $\overset{˜}{X}_{n}$ is
defined by

##### (29)

$\overset{˜}{H}_{n}\left(x\right)=\frac{1}{J}\sum _{j=1}^{J}I_{\left\{\overset{˜}{X}_{n,\left(j\right)}\leq x\right\}}$for $x\in \mathrm{\mathbb{R}}$. Note that empirical distribution function $\overset{˜}{H}_{n}$ is an estimator of true distribution function $\overset{˜}{F}_{n}$.

First, the Kolmogorov-Smirnov test uses the supremum distance between $\overset{˜}{H}_{n}$ and $\Phi$. Set

for $j\in \left\{1,\cdots ,J\right\}$. Then, its test statistic $KS_{J}$ is expressed by

where

##### (32)

$$ \begin{aligned} K S_{J}^{+} &=\max \left\{\frac{j}{J}-U_{(j)}: j \in\{1, \cdots, J\}\right\} \\ K S_{J}^{-} &=\max \left\{U_{(j)}-\frac{j-1}{J}: j \in\{1, \cdots, J\}\right\} . \end{aligned} $$A smaller value of test statistic $KS_{J}$ indicates a higher similarity between $\overset{˜}{F}_{n}$ and $\Phi$. If the value of $KS_{J}$ is higher than a prescribed critical value, the Kolmogorov-Smirnov test rejects null hypothesis $\mathrm{H}_{0}$.

Secondly, the Cramér-von Mises test uses a quadratic distance between $\overset{˜}{H}_{n}$ and $\Phi$. Its test statistic $CM_{J}$ is expressed by

A smaller value of test statistic $CM_{J}$ indicates a higher similarity between $\overset{˜}{F}_{n}$ and $\Phi$. If the value of $CM_{J}$ is higher than a prescribed critical value, the Cramér-von Mises test rejects null hypothesis $\mathrm{H}_{0}$.

Thirdly, the Anderson-Darling test uses a weighted quadratic distance between $\overset{˜}{H}_{n}$ and $\Phi$. Its test statistic $AD_{J}$ is expressed by

##### (34)

$$ \begin{aligned} A D_{J}=&-J-\frac{1}{J} \sum_{j=1}^{J}(2 j-1) \log \left(U_{(j)}\right) \\ &-\frac{1}{J} \sum_{j=1}^{J}(2 J-2 j+1) \log \left(1-U_{(j)}\right) \end{aligned} $$A smaller value of test statistic $AD_{J}$ indicates a higher similarity between $\overset{˜}{F}_{n}$ and $\Phi$. If the value of $AD_{J}$ is higher than a prescribed critical value, the Anderson-Darling test rejects the null hypothesis $\mathrm{H}_{0}$.

Table 2. Interfering sensor nodes.

Figs. 5 and 6 demonstrate empirical distribution functions constructed from the samples of the normalized interference power. Meanwhile, Figs. 7-9 show the results of Kolmogorov-Smirnov, Cramér-von Mises and Anderson-Darling tests for the normality of interference power. In these figures, we suppose a wireless passive sensor network, which consists of a single sink node and 18 sensor nodes, as illustrated in Fig. 3. We also assume that the parameters involved in the path loss model take the values appearing in Table 1. In addition, we suppose that the interference is created by 2, 5, 8, 11, 14, or 17 packets, which are sent by the sensor nodes listed in Table 2. For each number of interfering packets, we generate 1000 samples of the normalized interference power and then perform the three tests for normality.

In each of Figs. 5 and 6, empirical distribution functions of the normalized interference power, denoted by $\overset{˜}{F}_{n}$ are compared with the distribution function of standardized normal random variable, denoted by $\Phi$. We suppose that the interference is created by 2 packets in Fig. 5, whereas it is induced by 17 packets in Fig. 6. In these figures, we observe that both empirical distribution functions are similar to the distribution function of the standardized normal random variable. In addition, we notice that $\overset{˜}{F}_{n}$ is more similar to $\Phi$ when 17 packets create interference than when 2 packets induce interference.

Fig. 7 shows the values of the Kolmogorov-Smirnov test statistic with respect to the number of interfering packets. In this figure, these values of the test statistic are compared with the critical values at the 0.01, 0.05, 0.1, and 0.2 levels of significance. In this figure, we observe that the value of the Kolmogorov-Smirnov test statistic is lower than the critical value at the 0.2 level of significance when the number of interfering packets is 8, 11, 14, or 17. Thus, null hypothesis $\mathrm{H}_{0}$, which proposes that $\overset{˜}{F}_{n}=\phi $, cannot be rejected at the significance level of 0.2. In other words, we cannot say that the interference power does not have a normal distribution. In this figure, we also notice that the value of the test statistic is lower than the critical value at the 0.05 level of significance when 2 packets induce interference. Thus, from such a notice, $\mathrm{H}_{0}$ cannot be rejected at the 0.05 level of significance.

Fig. 5. Empirical distribution function vs. normal distribution function (Number of interfering packets = 2).

Fig. 6. Empirical distribution function vs. normal distribution function (Number of interfering packets = 17).

Fig. 8 shows the values of the Cramér-von Mises test statistic with respect to the number of interfering packets. The tendency expressed by the test results in Fig. 8 is similar to the one exhibited in Fig. 7. When the number of interfering packets is 8, 11, 14, or 17, the value of the test statistic is lower than the critical value at the 0.2 level of significance. Thus, the null hypothesis $\mathrm{H}_{0}$ cannot be rejected at the significance level of 0.2. When the number of interfering packets is 2, the value of the test statistic is also lower than the critical value at the 0.05 level of significance so that $\mathrm{H}_{0}$ cannot be rejected.

Fig. 9 shows the values of the Anderson-Darling test statistic with respect to the number of interfering packets. In this figure, we observe that the value of the test statistic is lower than the critical value at the 0.05 level of significance when the number of interfering packets is 8, 11, 14, or 17. Thus, null hypothesis $\mathrm{H}_{0}$ cannot be rejected at the significance level of 0.05. However, when the number of interfering packets is 2, the value of the test statistic is higher than the critical value at the 0.01 level of significance. Thus, $\mathrm{H}_{0}$ can be rejected at the significance level of 0.01.

In conclusion, most of the test results confirm that the null hypothesis $\mathrm{H}_{0}$ cannot be rejected at the 0.05 level of significance, i.e., we cannot say that the interference power does not have a normal distribution. Also, the test results explicitly reveal the tendency that $\mathrm{H}_{0}$ cannot be rejected at higher level of significance as the interference is created by more packets.

## 6. Comparison of Approximate Distribution Functions of Interference Power

In this section, we consider log-normal and Erlang approximations in ^{[22]} and ^{[24]} in addition to the normal approximation. Then, we compare the approximate distribution
functions with the empirical distribution function of the interference power.

Figs. 10 and 11 compare 3 approximate distribution functions (normal, log-normal and Erlang distribution functions) with the empirical distribution function of the interference power when the number of interfering packets is 2 and 17, respectively. In these figures, we assume that the wireless passive sensor network is configured as illustrated in Fig. 3 and is characterized by the parameter values as listed in Table 1. Also, the interfering sensor nodes are selected according to Table 2. Figs. 10 and 11 demonstrate that each of the normal, log-normal and Erlang approximations yield a distribution function similar to the empirical distribution function of the interference power. Also, each approximation exhibits no meaningful difference between empirical and approximate distribution functions.

Fig. 10. Comparison of approximate distribution functions with empirical distribution function of interference power when packets of 2 sensor nodes interfere.

## 7. Investigation of Throughput Value Yielded by Normal Approximation

In this section, we focus on the throughput that the wireless passive sensor network is able to achieve and examine the accuracy of the throughput value yielded by the normal approximation.

Fig. 12 shows the exact and approximate values of the throughput with respect to the number
of slots per frame. In this figure, we suppose that the wireless passive sensor network
is configured as illustrated in Fig. 3 and is governed by the parameter values listed in Table 1. Also, we assume that the wireless passive sensor network employs the contending-type
MAC scheme addressed in ^{[37]}. The MAC scheme uses the same time structure as depicted in Fig. 2, where each transmitting interval consists of a number of slots and each charging
interval perdures for 40 slot times. In every frame, each sensor node equally likely
selects a slot among the slots that comprise the transmitting interval of the frame.
Then, the sensor node transmits a packet during the selected slot. In Fig. 12, both of the exact and approximate throughput curves are obtained by use of a simulation
method, where the approximate throughput curve is yielded by the normal approximation
to the interference power. Fig. 12 reveals that the approximate value of the throughput barely differs from the exact
value regardless of the number of slots per frame.

## 8. Conclusion

We considered a wireless passive sensor network in which sensor nodes deliver their packets to a sink node according to a contending-type MAC scheme. In the network, a packet collision inevitably takes place and the interference power brought by the colliding packets may hinder a sink node from identifying any packet. Apparently, the interference power has a strong influence on the performance of a wireless passive sensor network so that it is of necessity to appropriately describe the interference power for performance evaluation. In this paper, we characterized the interference power as a sum of random variables governed by log-normal distributions, which are mutually independent but are not necessarily identical. Then, we proved the asymptotic normality of the interference power and, in pursuit of tractability rather than accuracy, proposed a normal approximation to the interference power. First, as a measure for the accuracy of the proposed normal approximation, we chose the Kolmogorov-Smirnov distance and presented an exact expression of the Berry-Esséen bound associated with the interference power. Comparative studies of Berry-Esséen bounds implied that the normal approximation to the interference power is moderately accurate. Next, to explore the acceptability of the proposed normal approximation, we employed three well-known statistical tests for normality. Hypothesis testing revealed that the normal approximation is acceptable even in the case that the interference power is incurred by a relatively small number of sensor nodes in a wireless passive sensor network.

### REFERENCES

### Appendices

#### A.1 Proof of Lemma 1

In (9), we set $\delta =2$ and will show that

##### (A1)

$\frac{\sum _{k=1}^{n}E\left(\left[V_{{\pi _{k}}}-E\left(V_{{\pi _{k}}}\right)\right]^{4}\right)}{\left[\sum _{k=1}^{n}Var\left(V_{{\pi _{k}}}\right)\right]^{2}}\rightarrow 0$as $n\rightarrow \infty $. Note that the variance of $V_{{\pi _{k}}}$

##### (A2)

$Var\left(V_{{\pi _{k}}}\right)=\left[e^{{\alpha ^{2}}{\omega ^{2}}}-1\right]e^{{\alpha ^{2}}{\omega ^{2}}+2{\beta _{{\pi _{k}}}}}$and the kurtosis of $V_{{\pi _{k}}}$

##### (A3)

$$ \begin{array}{l} \operatorname{Kur}\left(V_{\pi_{k}}\right) \\ =E\left(\left[\frac{V_{\pi_{k}}-E\left(V_{\pi_{k}}\right)}{\sqrt{\operatorname{Var}\left(V_{\pi_{k}}\right)}}\right]^{4}\right) \\ =e^{4 \alpha^{2} \omega^{2}}+2 e^{3 \alpha^{2} \omega^{2}}+3 e^{2 \alpha^{2} \omega^{2}}-3 \end{array} $$for $k\in \left\{1,\cdots ,n\right\}$. From (A1), (A2) and (A3), we have

##### (A4)

$$ \begin{array}{l} \frac{\sum_{k=1}^{n} E\left(\left[V_{\pi_{k}}-E\left(V_{\pi_{k}}\right)\right]^{4}\right)}{\left[\sum_{k=1}^{n} \operatorname{Var}\left(V_{\pi_{k}}\right)\right]^{2}} \\ =\operatorname{Kur}\left(V_{\pi_{k}}\right) \frac{\sum_{k=1}^{n} e^{4 \beta_{\pi_{k}}}}{\left[\sum_{k=1}^{n} e^{2 \beta_{\pi_{k}}}\right]^{2}} \end{array} $$Since $\omega $ is finite, $Kur\left(V_{{\pi _{k}}}\right)$ is finite and positive as well. Thus, to prove (A1), it is enough to show that

##### (A5)

$\frac{\sum _{k=1}^{n}e^{4{\beta _{{\pi _{k}}}}}}{\left[\sum _{k=1}^{n}e^{2{\beta _{{\pi _{k}}}}}\right]^{2}}\rightarrow 0$as $n\rightarrow \infty $. Recall that

for $k\in \left\{1,\cdots ,n\right\}$. Then,

##### (A7)

$\frac{\sum _{k=1}^{n}e^{4{\beta _{{\pi _{k}}}}}}{\left[\sum _{k=1}^{n}e^{2{\beta _{{\pi _{k}}}}}\right]^{2}}=\frac{\sum _{k=1}^{n}\frac{1}{D_{\pi _{k}}^{2\gamma }}}{\left[\sum _{k=1}^{n}\frac{1}{D_{\pi _{k}}^{\gamma }}\right]^{2}}.$Note that $D_{{\pi _{k}}}$ is bounded from above and from below by $D_{U}$ and $D_{L}$, respectively, for all $k\in \left\{1,\cdots ,n\right\}$. Thus,

##### (A8)

$\frac{\sum _{k=1}^{n}\frac{1}{D_{\pi _{k}}^{2\gamma }}}{\left[\sum _{k=1}^{n}\frac{1}{D_{\pi _{k}}^{\gamma }}\right]^{2}}\leq \frac{\sum _{k=1}^{n}\frac{1}{D_{L}^{2\gamma }}}{\left[\sum _{k=1}^{n}\frac{1}{D_{U}^{\gamma }}\right]^{2}}=\frac{1}{n}\left(\frac{D_{U}}{D_{L}}\right)^{2\gamma }\rightarrow 0$as $n\rightarrow \infty $. Therefore, (A1) holds.

#### A.2 Proof of Theorem 1

$V_{{\pi _{1}}},V_{{\pi _{2}}},\cdots $ satisfy the Lyapunov condition from lemma
1. Thus, from the Lyapunov central limit theorem ^{[25]}, $\overset{˜}{X}_{n}$ converges to the standardized normal random variable in distribution
as $n\rightarrow \infty $.

#### A.3 Proof of Theorem 2

To calculate the Berry-Esséen bound associated with $X_{n}$, we need the third absolute central moment and the variance of $V_{{\pi _{k}}}$. First, the third absolute central moment of $V_{{\pi _{k}}}$ is directly calculated as follows.

##### (A9)

$$ \begin{array}{l} E\left(\left|V_{\pi_{k}}-E\left(V_{\pi_{k}}\right)\right|^{3}\right) \\ =\int_{0}^{E\left(V_{\pi k}\right)}\left[-x+E\left(V_{\pi_{k}}\right)\right]^{3} f_{\pi_{k}}(x) d x \\ +\int_{E\left(V_{\pi_{k}}\right)}^{\infty}\left[x-E\left(V_{\pi_{k}}\right)\right]^{3} f_{\pi_{k}}(x) d x \end{array} $$where $f_{{\pi _{k}}}$ is the density of $V_{{\pi _{k}}}$ given in (6) and

##### (A10)

$E\left(V_{{\pi _{k}}}\right)=e^{\frac{1}{2}{\alpha ^{2}}{\omega ^{2}}+{\beta _{{\pi _{k}}}}}.$Replacing $x$ with $e^{\alpha \omega y+{\beta _{{\pi _{k}}}}}$ in (A9), we have

##### (A11)

$$ \begin{aligned} =& \int_{-\infty}^{\frac{\alpha \omega}{2}}\left[-e^{\alpha \omega y+\beta_{\pi k}}+\frac{1}{2} \alpha^{2} \omega^{2}+\beta_{\pi_{k}}\right]^{3} d \Phi(y) \\ &+\int_{\frac{\alpha \omega}{2}}^{\infty}\left[e^{\alpha \omega y+\beta_{\pi_{k}}}-\frac{1}{2} \alpha^{2} \omega^{2}+\beta_{\pi_{k}}\right]^{3} d \Phi(y) \end{aligned} $$where $\Phi$ is the distribution function of the standardized normal random variables given in (11). Straightforwardly calculating (A11), we then have

##### (A12)

$$ \begin{array}{l} E\left(\left|V_{\pi_{k}}-E\left(V_{\pi_{k}}\right)\right|^{3}\right) \\ =e^{\frac{9 \alpha^{2} \omega^{2}}{2}+3 \beta_{\pi_{k}}}\left[2 \Phi\left(\frac{5 \alpha \omega}{2}\right)-1\right] \\ -3 e^{\frac{5 \alpha^{2} \omega^{2}}{2}+3 \beta_{\pi_{k}}}\left[2 \Phi\left(\frac{3 \alpha \omega}{2}\right)-1\right] \\ +4 e^{\frac{3 \alpha^{2} \omega^{2}}{2}+3 \beta_{\pi_{k}}}\left[2 \Phi\left(\frac{\alpha \omega}{2}\right)-1\right] . \end{array} $$Secondly, the variance of $V_{{\pi _{k}}}$ is given in (7). Using (A12) and (7), we finally obtain the Berry-Esséen bound given in (18).

#### A.4 Proof of Theorem 3

To calculate the Berry-Esséen bound associated with $S_{n}$, we need the third absolute central moment and the variance of $T_{k}$.

Suppose that $T_{k}$ is a normal random variable with density in (20). The third absolute central moment of $T_{k}$ is calculated by

##### (A13)

$E\left(\left| T_{k}-E\left(T_{k}\right)\right| ^{3}\right)=2\sqrt{\frac{2}{\pi }}\sigma _{k}^{3}.$Also, the variance of $T_{k}$ is $\sigma _{k}^{2}$. Using (A13), we then obtain the Berry-Esséen bound given in (24).

Suppose that $T_{k}$ is a uniform random variable with density in (21). The third absolute central moment of $T_{k}$ is yielded by integrating polynomials as follows.

##### (A14)

$$ \begin{aligned} & E\left(\left|T_{k}-E\left(T_{k}\right)\right|^{3}\right) \\ =& \int_{a}^{E\left(T_{k}\right)}\left[-x+E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ &+\int_{E\left(T_{k}\right)}^{b}\left[x-E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ =& \frac{1}{32}\left(b_{k}-a_{k}\right)^{3} \end{aligned} $$where $g_{k}$ is given in (21) and $E\left(T_{k}\right)=\frac{1}{2}\left(a_{k}+b_{k}\right)$. Also, the variance of $T_{k}$

Using (A14) and (A15), we then obtain the Berry-Esséen bound given in (25).

Suppose that $T_{k}$ is an exponential random variable with density in (22). The third absolute central moment of $T_{k}$ is yielded by integrating exponential functions as follows.

##### (A16)

$$ \begin{array}{l} E\left(\left|T_{k}-E\left(T_{k}\right)\right|^{3}\right) \\ =\int_{\delta_{k}}^{E\left(T_{k}\right)}\left[-x+E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ +\int_{E\left(T_{k}\right)}^{\infty}\left[x-E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ =\frac{1}{\lambda_{k}^{3}}\left(12 e^{-1}-2\right) \end{array} $$where $g_{k}$ is given in (22) and $E\left(T_{k}\right)=\delta _{k}+\frac{1}{\lambda _{k}}$. Also, the variance of $T_{k}$

Using (A16) and (A17), we then obtain the Berry-Esséen bound given in (26).

Suppose that $T_{k}$ is a Pareto random variable with density in (23). The third absolute central moment of $T_{k}$ is yielded by integrating polynomials as follows.

##### (A18)

$$ \begin{aligned} E &\left(\left|T_{k}-E\left(T_{k}\right)\right|^{3}\right) \\ =& \int_{\delta_{k}}^{E\left(T_{k}\right)}\left[-x+E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ &+\int_{E\left(T_{k}\right)}^{\infty}\left[x-E\left(T_{k}\right)\right]^{3} g_{k}(x) d x \\ =& \frac{12 \delta_{k}^{3}}{\left(v_{k}-1\right)\left(v_{k}-2\right)\left(v_{k}-3\right)}\left(1-\frac{1}{v_{k}}\right)^{v_{k}-3} \\ &-\frac{2 \delta_{k}^{3} v_{k}\left(v_{k}+1\right)}{\left(v_{k}-1\right)^{3}\left(v_{k}-2\right)\left(v_{k}-3\right)} \end{aligned} $$where $g_{k}$ is given in (23) and $E\left(T_{k}\right)=\delta _{k}\frac{\nu _{k}}{\nu _{k}-1}$. Also, the variance of $T_{k}$

##### (A19)

$Var\left(T_{k}\right)=\delta _{k}^{2}\frac{\nu _{k}}{\left(\nu _{k}-1\right)^{2}\left(\nu _{k}-2\right)}.$Using (A18) and (A19), we then obtain the Berry-Esséen bound given in (27).

## Author

Heewon Seo received the B.S. degree in electronics and computer engi-neering and the M.S. degree in computer science and engineering, and the Ph.D. degree in computer science and engineering from Dankook University, Korea, in 2010, 2012, and 2019, respectively. In 2019, he joined the faculty of Dankook University, Korea, where he is presently an invited professor in the department of computer engineering. His research interests include design and performance analysis of medium access control schemes for energy harvesting wireless sensor networks and RFID networks. He is a member of IEIE and IEEE.

Jin Kyung Park received the B.A. degree in economics and the M.S. degree in information and communi-cations, and the Ph.D. degree in electronics and computer engineering from Dankook University, Korea, in 1998, 2002, and 2009, respectively. From 2009 to 2010, Dr. Park was a post-doctorate researcher at the electrical engineering department of the University of California at Los Angeles, U.S.A. From 2017 to 2019, she was an invited professor at the department of applied computer engineering, Dankook University, Korea. In 2019, she joined the faculty of Ajou University, Korea, where she is presently a teaching assistant professor at the department of software and computer engineering. Her research interests include design and performance analysis of medium access control schemes for wireless networks, especially wireless MANs, wireless sensor networks, and RFID networks. She is a member of IEIE and IEEE.

Cheon Won Choi received the B.S. degree and the M.S. degree in electronics engineering from Seoul National University, Korea, and the Ph.D. degree in electrical engineering from the University of California at Los Angeles, U.S.A. in 1986, 1988, and 1996, respectively. From 1996 to 1997, Dr. Choi was an engineer at IRI Computer Communications Corporation. Since 1997, he has been on the faculty of Dankook University, Korea, where he is presently a professor at the department of computer engineering. From 2006 to 2008, he served as the chief information officer at Dankook University. In 2011, he was appointed as the chair of the research institute for convergence of information and telecommunications technologies at Dankook University and led the institute for four years. His research interests include medium access control and resource allocation in wireless networks, and application of queueing theory and game theory to performance analysis. He is a life member of IEIE, and a senior member of IEEE.