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\begin{document}

\title{A framework for Bayesian Posterior Simulation Methods in clinical
practice}



\author[1]{Razieh Bidhendi Yarandi}%
\author[1]{mohammad ali Mansournia}%
\author[1]{hojjat Zeraati}%
\author[1]{Kazem Mohammad}%
\affil[1]{Tehran University of Medical Sciences}%


\vspace{-1em}



  
  \date{June 14, 2024}


\begingroup
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\selectlanguage{english}
\begin{abstract}
Purpose: Bayesian inference has become very popular science. It offers
some pragmatic approaches to account for uncertainty in
inference-decision making. Various estimation methods have been
introduced to implement Bayesian methods but although these algorithms
are powerful they are not always easy to grasp. This paper aims to
provide an intuitive framework of four key Bayesian computational
methods for researchers in clinical studies. We will not cover daunting
mathematical discussion of these approaches, but rather offer a
non-quantitative description of these algorithms and provide some
illuminating examples. Materials and methods: Bayesian computational
methods namely, i) Importance sampling, ii) Rejection sampling, iii)
Markov-chain Monte Carlo, iv) Data augmentation were introduced. Results
and conclusions: A load of literature published on Bayesian inference
has proved its popularity among researches while its concept is not
straightforward for amateur learners. We showed that alternative
approaches which are intuitively appealing and easy-to-understand work
well in case of low-dimensional problems and appropriate Prior
information such as weighted prior, otherwise MCMC is a Trouble-free
tool.%
\end{abstract}%



\sloppy


\textbf{INTRODUCTION}

The growth in the application of Bayesian analysis in the sciences has
created a need to present its complex concepts in a more understandable
language for non-statisticians. To many amateur users, their
computationally-intensive simulation approaches has the appearance of a
``black box''. This article aims to describe essential approaches used
in Bayesian methods for posterior simulation in an intuitive manner.
Four Bayesian computational methods are presented: \emph{Importance
Sampling (IS)} , \emph{Rejection sampling (RS),} \emph{Markov Chain
Monte Carlo (MCMC)} and \emph{Data Augmentation (DA)} . In this study,
we aim to provide a comprehensive, yet easy to follow explanation of
these techniques to make them practical for researchers. Some
illuminating examples are presented to illustrate the algorithms and the
concepts they embody. R software code and some other information are
available in the Supplementary file as well.

\textbf{METHODS}

Historically, one of the first methods of simulation by computer was
Monte Carlo integration (MCI). Its computational algorithm relies on
repeated random sampling from distributions of interest to obtain
numerical results. For example, MCI is used to calculate the expectation
of a distribution function when its estimation through integration is
impossible. Its plain logic is to generate samples from a distribution
function\(\ \)and approximate the expectation numerically by
calculating the average of generated samples. In other words, empirical
experimental summation is substituted for analytical (or possibly
numerical) integration. This technique is conceptually simple but not
always very efficient if the target probability function is not
well-behaved. Attempts to deal with this problem led to a gradual
evolution first to IS and next in RS; they can be regarded as improved
versions of MCI. However, in some more challenging cases they have also
proven inadequate to the task. Over the past 30 years or so, MCMC
methods have revolutionized statistical computing and permitted ever
more complex problems to be handled. Finally, we have included DA which
follows a different approach to calculate posterior estimates.

\textbf{Importance Sampling: A clever substitution of sampling region}

MCI as a computational method was first initiated to solve the
integration problem in estimating expectations. Later it was applied to
the simulation of Bayesian posterior (BP) distributions. It has a
transparent algorithm: Generate random samples from a distribution
function say ``Target'', then numerically calculate the integral by
summing the values. The expectation obtained by MCI can be referred to
as the \emph{Empirical Average} (Supplementary Formula-1). For example,
imagine we want to estimate the expectation of the
function\(h\left(x\right)=\sin\left(x\right)\sqrt{\left|\cos\left(x\right)\right|}\), where the random variable (X) follows a
Normal distribution with mean=0 and SD=5 (target). Firstly, generate
10000 random samples from target distribution, then obtain the values of
h(x) from each generated sample, and then calculate numerically the mean
and variance of this generated sample (Supplementary R Code-1). This
random sampling method is not cost-effective when the target
distribution is diffuse, because a large sample size is required to
obtain acceptable precision.

MCI was improved by IS, a variance reduction technique which was first
presented in statistical physics (1; 2). IS relaxes the procedure of
treating all parts of the distribution equally, concentrating instead on
those where estimation was critical. In this respect, an alternative
function, say ``Proposal'', close to the target, is suggested by making
an educated guess. Contrary to MCI on which samples are treated evenly,
a ``weight'' which shows the importance of a sample is allocated to each
generated sample through an importance function. Actually, for each
sample, one calculates the likelihood of getting that sample from the
target distribution proportion to the likelihood of sampling it from the
proposal distribution. After the sampling process is finished, the
obtained relative likelihoods were normalized in a way that they sum to
one. In this way, each point has its own likelihood of occurrence as a
discrete probability distribution. The expectation obtained by IS
called\emph{Weighted Average} (Supplementary Formula-2)\emph{.}

Consider a Normal (0, 0.05) and a t-student (dF=1) as the two different
proposal distributions for Normal (0, 5), and estimate importance
weights through the importance functions\(\frac{target:\ N(0,5)}{proposal:\ N\left(0,0.05\right)}\)
and\(\ \frac{N(0,5)}{t(1)}\) for each generated point. After that, normalize
the weights as\(\frac{\frac{N(0,5)}{N\left(0,0.05\right)}}{\sum\frac{N(0,5)}{N\left(0,0.05\right)}}\)and \(\ \frac{\frac{N(0,5)}{t(1)}}{\sum\frac{N(0,5)}{t(1)}}\) . Therefore, we
have a discrete distribution function for which its properties such as
the mean and variance are easily estimable. Estimated means for our
generated samples were approximately 0 as obtained from MCI but their
variances are considerably lower than the MCI approach (Supplementary R
Code-2). Using alternative distributions can improve the variance of the
samples, although a wide proposal distribution leads to worse estimates
in terms of the variance and inefficient due to a large sampling number
(Figure1). Choosing an appropriate proposal distribution that looked
similar to the target would be ideal though difficult to find at times.
Unbiased estimates of parameters are obtained for large samples by IS.
It also works well when the importance function is not very variable.
Indeed, an appropriate proposal distribution leads to lower variances
and higher accuracy of approximation. Robert and Casella provided an
example illustrating the use of Normal (0, 1) as a proposal to resemble
sampling from a Cauchy C(0,1) target distribution caused infinite
variance of importance weights (3). It leads to attach high importance
to few points and provides inefficient estimates in terms of variance.
By substituting heavy tails distributions like t-student rather than
Normal, reasonable fitness is guaranteed.

\textbf{Weighted Prior as an application of IS in BP simulation}

In this section, we show how IS plays a role in BP simulation. As Bayes
rule, posterior distribution is proportional to the likelihood of
observed data and prior distribution of parameter (Supplementary
Formula-3). It emphasizes that likelihood can play the role of
importance function for priors or for a prior. Using this approach, the
expectation of a posterior distribution function can be calculated. In
addition, to paint a clearer picture of the posterior density and its
properties- such as percentiles-it suffices to generate weighted samples
from the prior.

This follows a straightforward algorithm: 1) Generate a random sample
from a prior distribution. 2) Consider the values of the likelihood of
the data in this generated sample as the weight and continue this
process, 3) Normalize the weights. Consequently, each point has its own
probability of occurrence by the likelihood of data. This na\selectlanguage{ngerman}ïve sampling
method which we referred to as Weighted Prior (WP) is equivalent to IS.
As such IS method provides an alternative way to exact simulation from
posterior (Supplementary Formula-4). It is worth mentioning that, bias
estimated through this approach is of an order of\(\frac{\mathbf{1}}{\mathbf{n}}\) so
in large samples it provides unbiased estimates. It should be considered
that, the contribution of generated samples to the estimation of the
final posterior depends on how much it is supported by data. This means
that, prior distributions should not be that much far off from the
likelihood if so, most generated points receive very small importance
weights. We depicted a case when this method fails (Figure 2).

\section*{Rejection Sampling: Hit the
Target!}

{\label{rejection-sampling-hit-the-target}}

Contrary to IS which has no restriction to the choice of proposal
distribution, RS is pickier. Imagine that our objective is to generate
samples from a function. The vital condition is that proposal
distributions should cover the target. Consider\(\ g(\theta)\) as a
proposal function from which samples generated, (Supplementary 1.
Rectangle), \(f(\theta)\) as target and K as a known constant
(Supplementary Figure 1. Zigzag triangle). The algorithm is
straightforward, generate sample\(\selectlanguage{greek}\text{\ θ}\selectlanguage{english}_{0}\ \)from a proposal and
calculate the likelihood of target and proposal functions at this point.
If the likelihood ratio of target to proposal was greater than a random
sample generated from uniform (0, 1), we accept\(\theta_{0}\mathbf{\ }\)as a
sample from target otherwise, reject it and continue the algorithm. In a
Bayesian context, this idea could extend to estimating the posterior
distribution by considering K as the Maximum likelihood
of\selectlanguage{ngerman}  \selectlanguage{greek}θ \selectlanguage{english}and prior as proposal, the probability of each
generated sample is compared to the value of maximum
likelihood\(\text{\ L}_{\text{Max\ }}\) so the proposal resembles posterior
distribution, (Supplementary Formula-5) and (Supplementary figure 2).
This method fails in high dimensional space parameters due to a decrease
in the acceptance rate. In addition, when
\(Kg(\theta)\)\(\ \)has considerable distance compared
to \(f\left(\theta\right)\), acceptance rate would decrease drastically as
well. Likewise, in Bayesian context when \(L_{\text{Max}}\ \)is not
practically possible to calculate, this method fails. Armitage presented
an example in which by this approach posterior estimation of parameters
of a linear regression model had been estimated via various priors as
the proposals(4), that using a diffuse prior caused high rejection rate
and proved the approach futile.

\textbf{Markov Chain Monte Carlo method (MCMC): Transition from
uncertainty to stationary state.}

The emergence of MCMC approach in the 1990s led to a rapid evolution of
methods to simulate posterior distributions. It can be regarded as MCI
through Markov chains. Contrary to other methods that have a static
mechanism, MCMC follows a dynamic mechanism whereby samples are
generated via a gentle transition through a target distribution function
by considering a proposal, eventually converging on a stationary
distribution. There are two popular algorithms: Metropolis-Hastings
firstly introduced by Metropolis in 1953 and its special case the Gibbs
sampler, introduced by Geman and Geman in 1984. Recent developments in
this era have provided huge literature, Armitage provided a neat
catalogue of the references and summaries (Armitage, Berry, and
Matthews, 2001; Armitage et al., 2005). Let us review some technical
jargon in MCMC. A chain with the property of being Markov is applied to
generate samples which in consequence are dependent. A transition matrix
illustrates the probability of movement from one state in the chain to
another. It is worth mentioning that the Markov chain should have some
properties that guarantee to produce samples from a stationary
distribution. By stationary, we meant that, if a sample is generated
from a distribution, the next is from the same distribution, as well.
Firstly, it should be \emph{Irreducible} meaning that the Markov chain
can get to any state from any state within a finite number of
iterations, \emph{Persistent or recurrent} means returning to the state
at least once and \emph{Non-Null} means finite mean number of
transitions. These three conditions had led to a property referred
to\emph{ergodicity} by which one can ensure the generation of samples
from a stationary distribution. Therefore, the variance of the generated
samples is estimable otherwise the chain may behave badly and be
effectively useless. It also proves the consistency of estimates.
Failure of this method occurs when there are convergence issues. For
instance, in the case of a non-persistent chain, convergence to a
stationary distribution never occurs. Being \emph{Symmetric} is another
property which influences the acceptance probability of sampling. It
means that\(g\left(\theta^{\prime\prime}\middle|\theta^{{}^{\prime}}\right)=g\left(\theta^{{}^{\prime}}\middle|\theta^{\prime\prime}\right)\), (Supplementary figure 3).

The Metropolis-Hastings algorithm to produce a chain of samples by
iterative mechanism is defined as the following steps;

\begin{enumerate}
\tightlist
\item
  Generate a candidate \(\theta^{\left({}^{\prime}\right)}\ \)from proposal
  distribution\(\text{\ g}\left(.\middle|\theta^{\left(i\right)}\right)\)on which \(\theta^{\left(i\right)}\) is an
  initial point. The next step is to see whether it is an acceptable
  sample with a high probability of occurrences and accept it
  as\(\theta^{\left(i+1\right)}\) or not.
\item
  An acceptance rule akin to rejection sampling should be considered
  here. In this way, at first acceptance probability of the candidate
  should be estimated. It is the minimum of one and the value of the
  proportion of multiplication of proposal and target distribution in
  candidate sample condition on previous point (initial) and vice versa.
  When the proposal is symmetric, it reduces to the proportion of the
  values of target on the candidate and initial points (proposal
  eliminated from nominator and denominator (Supplementary Formula-6).
\item
  Now proceed to decide whether to accept the candidate as the next
  sample or not. So, generate a random number \emph{u} from uniform (0,
  1) distribution.
\item
  If \emph{u} was greater than acceptance probability, choose the
  candidate as the next sample, otherwise put initial as the next sample
  and continue the process.
\end{enumerate}

\textbf{Example: Accepted candidate}

Step 1:\({r(\beta}_{\text{new}},\ \beta_{t-1})=\frac{Posterior(\beta_{\text{new}})}{Posterior(\beta_{t-1})}=\frac{Beta(1,1,0.4)\times Binomial(10,4,0.4)}{Beta(1,1,0.5)\times Binomial(10,4,0.5)}=1.19\)

Step 2: Acceptance probability\(\selectlanguage{greek}\text{α\ }\selectlanguage{english}{(\beta}_{\text{new}},\ \beta_{t-1})=min(1,\ r{(\beta}_{\text{new}},\ \beta_{t-1}))=min(1,\ 1.19)=1\)

Step3: Draw a random number, \emph{u} , from a Uniform (0, 1), here
u=0.345

Step4: If \emph{u} is less than the acceptance probability, the proposed
value of \(\beta_{\text{new}}\) will be accepted. Otherwise, we
reject\(\beta_{\text{new}}\) and keep\(,\ \beta_{t-1}\). Here we accept it.

\textbf{Example: Rejected candidate}

Step 1:\(\ {r(\beta}_{\text{new}},\ \beta_{t-1})=\frac{Posterior(\beta_{\text{new}})}{Posterior(\beta_{t-1})}=\frac{Beta(1,1,0.2)\times Binomial(10,4,0.2)}{Beta(1,1,0.3)\times Binomial(10,4,0.3)}=0.43\)

Step 2: Acceptance probability\(\selectlanguage{greek}\text{α\ }\selectlanguage{english}{(\beta}_{\text{new}},\ \beta_{t-1})=min(1,\ r{(\beta}_{\text{new}},\ \beta_{t-1}))=min(1,\ 0.43)=0.43\)

Step3: Draw a random number, \emph{u} , from a Uniform (0, 1),
here\emph{u} =0.675

Step4: Since u \textgreater{} r, we reject it with the probability 43\%.

\textbf{MCMC by Metropolis-Hastings Algorithm: An Intuitive
illustration}

How samples are generated by MCMC, is the main question of interest.
Imagine we have a uniform distribution (0, 5) as proposal, (Dashed line)
and a target, (Zigzag pattern) (Supplementary figure 4). To show how to
estimate this transition matrix, we considered a discrete target
distribution as well as two numbers of states involved. Therefore, we
need a 2\selectlanguage{ngerman}×2 transition matrix exists whose elements are the probability
of movement. To explain clearly, four possible transitions for
generating samples exist. If the first sample generated from state 1,
what the probabilities of being in the next state 1 or 2 would be, and
so when the first in 2. According to the proposal, the probabilities of
being in state 1 and 2 are\(\frac{\mathbf{1}}{\mathbf{5}}\ \)and\(\ \frac{\mathbf{4}}{\mathbf{5}}\),
respectively. Therefore, estimated Transition Matrix is
P=\(\par
\begin{bmatrix}\frac{\mathbf{13}}{\mathbf{15}}&amp;\frac{\mathbf{1}}{\mathbf{5}}\\
\frac{\mathbf{2}}{\mathbf{15}}&amp;\frac{\mathbf{4}}{\mathbf{5}}\\
\end{bmatrix}\ \)(Table1). If the process of sampling is repeated n
times, based on stochastic process the n\textsuperscript{th}-step
transition matrix will be\(\text{\ P}^{n}=\frac{1}{\alpha+p}\par
\begin{bmatrix}\alpha&amp;\alpha\\
p&amp;p\\
\end{bmatrix}\par
\begin{bmatrix}\frac{\mathbf{3}}{\mathbf{5}}&amp;\frac{\mathbf{3}}{\mathbf{5}}\\
\frac{\mathbf{2}}{\mathbf{5}}&amp;\frac{\mathbf{2}}{\mathbf{5}}\\
\end{bmatrix}\). It is proven this sampling
mechanism converges to a stationary form which shows target distribution
(Supplementary Figure 4).

According to the stochastic process theorem, from this convergence time
onwards each sample generated is independent of previous states(5). From
the Bayesian point of view, considering proposal as prior distribution
and target as prior \selectlanguage{ngerman}× likelihood, this process of sampling is defined as
Metropolis-Hasting which eventually converges to a stationary posterior.

\textbf{Data Augmentation: Translation of Common Sense into Reality}

Methods introduced in the past followed the Monte Carlo approach for
computing posterior distribution. On the contrary, DA allows approximate
Bayesian analysis with a standard maximum likelihood function. Its
philosophy is to translate prior information as equivalent data and add
this external information to the observed study data then conventional
methods of frequentist can be applied. No specific tools are required to
compute posterior mean and variance; inverse-variance weighted averaging
is a rule of thumb for estimation (6). This technique provides an
effective remedy to treat bias estimation caused by data sparseness
(7-12). In fact, it considers prior information as a penalty for maximum
likelihood estimates and approximates posterior mode and variance.

\textbf{Learn Posterior Estimation by Heart: An illustration in
pharmacology}

DA is a remedial tool for sparse data issues to provide unbiased
estimation of parameters. To illustrate DA mechanism, we considered how
to estimate posterior properties via inverse-variance weighting, and
then show the influence of prior and likelihood components on estimating
posterior distributions. Finally, we depict how to construct data from
prior to observe its role as equivalent data augmented to actual data.

Result of a hypothetical data was reported Ln(RR)=Ln (6.3) =1.82,
Variance(Ln(RR)) = 0.84 and 95\% limits RR= ( 1.02 , 37.3) which was
subjected to sparse data due to wide 95\% limits . Suppose that, in a
meta-analysis for side effect of a drug study we found prior information
for RR with 95\% limits between \(\frac{\mathbf{1}}{\mathbf{3}}\) and 3. Mean and
variance of prior for Ln (RR) are estimated as Prior mean for ln
(RR)=\(\frac{(\text{Ln}\left(\frac{1}{3}\right)+\text{Ln}(3))}{2}=\)0, and Prior Variance for ln
(RR)=\(\ {(\frac{\left|\text{Ln}\left(\frac{1}{3}\right)-Ln\left(3\right)\right|}{2*1.96})}^{2}=0.10\). Inverse variances equaling \(\frac{1}{0.1}=10\)
and\(\frac{1}{0.84}=1.2\) illustrating prior information dominated data
information by nearly 8 times. Posterior mean and variance for Ln (RR)
could be estimated as the following weighted averaging rule of thumb;
Posterior mean for\(\ \ln\left(\text{RR}\right)=\frac{\frac{0}{0.10}+\frac{1.82}{0.84}}{\frac{1}{0.10}+\frac{1}{0.84}}=\ \ 0.19\)and Posterior variance
for\(\ln{\left(\text{RR}\right)\approx\frac{1}{\frac{1}{0.10}+\frac{1}{0.84}}}=0.09\). Posterior RR and its 95\%CI through DA provided
unbiased estimation of RR with more reasonable values of RR and narrower
95\%CI. In addition, the value of posterior mean which is closer to
prior means showed the influence of the prior as well. Various prior
ranges for Ln (RR), estimated posterior 95\% CI, as well as the
influence of prior and data, were illustrated in (Table 2). It was
depicted that, for prior (\(\frac{\mathbf{1}}{\mathbf{6}}\), 6) data and prior had the
same influence (equal weights) while for (\(\frac{\mathbf{1}}{\mathbf{10}}\), 10) it was
data dominated.

Here, \emph{Compatibility} of prior and data is a great issue as well.
DA fails in case of incompatibility causes misleading results(13). For
our example,\(\frac{(Ln\left(6.3\right)-0)}{{(0.84+0.10)}^{\frac{1}{2}}}=1.9P_{\text{value}}=0.057\)showed that compatibility hypothesis is
not rejected.

\textbf{Summary of pros and cons of the approaches}

Supplementary table 2 presents advantages, failures and their remedies
of the approaches.

\textbf{DISCUSSION}

A load of literature published on Bayesian inference has proved its
popularity among researches while its concept is not straightforward for
amateur learners (14). The purpose of our paper was to provide a
comprehensive framework with illuminating examples to shed light upon
the concept of some of Bayesian mechanisms of sampling. We showed that
alternative approaches which are intuitively appealing and
easy-to-understand work well in case of low-dimensional problems and
appropriate Prior information such as weighted prior, otherwise MCMC is
a Trouble-free tool. Although its concept is not an intuitively
realizable advanced method of MCMC tackles most complex issues.
Different studies tried to cover Bayesian statistical approach as a need
specifically for many sciences (15-26). Also, DA method as an
alternative approach gives researchers more tangible sense in the role
of prior and data for inference making, the posterior calculation is
simple using this method. We tried to cover the sufficient methods of
Bayesian simulation approaches with some clear examples and provide an
introductory work of Bayesian foundation; R software codes are available
in the Supplementary as well.

\textbf{Declarations}

Ethics approval and consent to participate: Not applicable

Consent for publication: Not applicable

Availability of data and material: All data generated or analyzed during
the current study are

included in this published article.

Competing interests: The authors declare that they have no conflict of
interest.

Funding: This research received no specific grant from any funding
agency in the public,

commercial, or not-for-profit sectors.

Authors' contributions: Dr. RBY, Prof. KM, Prof. MAM and Prof. HZ had
significant contributions to the conception, design, acquisition,
analysis and interpretation of the information. Methodological concepts
were considered Dr. RBY, Prof. KM, Prof. MAM, Prof. HZ. All authors
worked on the drafting and agreed on final approval of the version to be
published. Also, the agreement to be accountable for all aspects of the
work in ensuring that questions related to the accuracy or integrity of
any part of the work are appropriately investigated and resolved.

\textbf{REFERENCES}

1. \textbf{Hammersley JM \& Morton KW}(1954) Poor Man's Monte Carlo.
\emph{Journal of the Royal Statistical Society. Series B} \textbf{16} ,
23-38.

2. \textbf{Rosenbluth MN \& Rosenbluth AW} (1955) Monte Carlo
Calculation of the Average Extension of Molecular Chains. \emph{The
Journal of Chemical Physics} \textbf{23} , 356-59.

3. \textbf{Robert C \& Casella G}(2010) \emph{Introducing Monte Carlo
Methods with R} : Springer-Verlag New York.

4. \textbf{Armitage P, Berry G \& Matthews JNS} (2001) \emph{Statistical
Methods in Medical Research} . 4th Edition ed: Wiley-Blackwell.

5. \textbf{T.J.Bailey N} (1966)\emph{Elements of stochastic processes
with applications to the Natural Sciences} : Wiley.

6. \textbf{Greenland S} (2006) Bayesian perspectives for epidemiological
research: I. Foundations and basic methods. \emph{International Journal
of Epidemiology} \textbf{35} , 765-75.

7. \textbf{Greenland S, Mansournia MA \& Altman DG} (2016) Sparse data
bias: a problem hiding in plain sight.\emph{BMJ} \textbf{352} .

8. \textbf{Greenland S \& Christensen R} (2001) Data augmentation priors
for Bayesian and semi-Bayes analyses of conditional-logistic and
proportional-hazards regression.\emph{Statistics in Medicine}
\textbf{20} , 2421-28.

9. \textbf{Bedrick EJ, Christensen R \& Johnson W} (1996) A New
Perspective on Priors for Generalized Linear Models. \emph{Journal of
the American Statistical Association}\textbf{91} , 1450-60.

10. \textbf{Bedrick EJ, Christensen R \& Johnson W} (1997) Bayesian
Binomial Regression: Predicting Survival at a Trauma Center. \emph{The
American Statistician} \textbf{51} , 211-18.

11. \textbf{Greenland S \& Mansournia MA} (2015) Penalization, bias
reduction, and default priors in logistic and related categorical and
survival regressions. \emph{Statistics in medicine} \textbf{34} ,
3133-43.

12. \textbf{Mansournia MA, Heinze G, Geroldinger A \& Greenland S}
(2017) Separation in Logistic Regression: Causes, Consequences, and
Control. \emph{American Journal of Epidemiology} \textbf{187} , 864-70.

13. \textbf{George EPB} (1980) Sampling and Bayes' Inference in
Scientific Modelling and Robustness.\emph{Journal of the Royal
Statistical Society. Series A (General)}\textbf{143} , 383-430.

14. \textbf{Albert J} (1997) Teaching Bayes' Rule: A Data-Oriented
Approach. \emph{The American Statistician}\textbf{51} , 247-53.

15. \textbf{Turner BM \& Van Zandt T}(2012) A tutorial on approximate
Bayesian computation. \emph{Journal of Mathematical Psychology}
\textbf{56} , 69-85.

16. \textbf{Etz A \& Vandekerckhove J} (2018) Introduction to Bayesian
Inference for Psychology.\emph{Psychonomic Bulletin \& Review}
\textbf{25} , 5-34.

17. \textbf{Matzke D, Boehm U \& Vandekerckhove J} (2018) Bayesian
inference for psychology, part III: Parameter estimation in nonstandard
models. \emph{Psychonomic Bulletin \& Review} \textbf{25} , 77-101.

18. \textbf{Wagenmakers E-J, Love J, Marsman M, et al.} (2018) Bayesian
inference for psychology. Part II: Example applications with JASP.
\emph{Psychonomic Bulletin \& Review}\textbf{25} , 58-76.

19. \textbf{Wagenmakers E-J, Marsman M, Jamil T, et al.} (2018) Bayesian
inference for psychology. Part I: Theoretical advantages and practical
ramifications. \emph{Psychonomic Bulletin \& Review} \textbf{25} ,
35-57.

20. \textbf{Zhang L, Pfister M \& Meibohm B} (2008) Concepts and
challenges in quantitative pharmacology and model-based drug
development. \emph{The AAPS journal} \textbf{10} , 552-59.

21. \textbf{Racine A, Grieve A, Fluhler H \& Smith A} (1986) Bayesian
methods in practice: experiences in the pharmaceutical industry.
\emph{Applied Statistics} , 93-150.

22. \textbf{Barrett JS, Fossler MJ, Cadieu KD \& Gastonguay MR} (2008)
Pharmacometrics: a multidisciplinary field to facilitate critical
thinking in drug development and translational research settings.
\emph{The Journal of Clinical Pharmacology} \textbf{48} , 632-49.

23. \textbf{Grieve AP} (2007) 25 years of Bayesian methods in the
pharmaceutical industry: a personal, statistical bummel.
\emph{Pharmaceutical Statistics} \textbf{6} , 261-81.

24. \textbf{Morgan D} (2018) Bayesian applications in pharmaceutical
statistics. \emph{Pharmaceutical Statistics} \textbf{17} , 298-300.

25. \selectlanguage{polish}\textbf{Miočević M}\selectlanguage{english} (2019) A Tutorial in Bayesian Mediation Analysis
With Latent Variables.\emph{Methodology} .

26. \textbf{Natesan P} (2019) Fitting Bayesian Models for Single-Case
Experimental Designs.\emph{Methodology} .

\textbf{Table1.} Estimation of a Two- State Transition Matrix for a
discreet target and a uniform (0, 5) proposal distribution\selectlanguage{english}
\begin{longtable}[]{@{}lll@{}}
\toprule
\textbf{PS} & \textbf{NS} &
\textsuperscript{\textbf{@}}\textbf{Probability of
transition}\tabularnewline
\midrule
\endhead
1 & 1 & \(\frac{1}{5}\times 1+\frac{10}{15}\) \emph{(repeat the sampling in case of
rejecting)}\tabularnewline
1 & 2 & \(\frac{4}{5}\times\frac{1}{6}\)\tabularnewline
2 & 1 & \(\frac{1}{5}\times 1\)\tabularnewline
2 & 2 & \(\frac{4}{5}\times 1\)\tabularnewline
\bottomrule
\end{longtable}

Abbreviations: Previous State (PS), Next State (NS),

\textsuperscript{@}Probability of transition: probability of state *
(probability of target distribution in NS \selectlanguage{ngerman}÷ probability of target
distribution in PS)

\textbf{Table 2.} Posterior 95\% CI for the Range of Prior Information
via Data Augmentation Method\selectlanguage{english}
\begin{longtable}[]{@{}llll@{}}
\toprule
\textbf{Prior95\% CI for Ln (RR)} & \textbf{Posterior 95\% CI for Ln
(RR)} & \textbf{Prior Information weight:} \(\mathbf{(}\frac{\mathbf{1}}{\mathbf{\text{prior\ variance\ }}\left(\mathbf{\ln}\left(\mathbf{\text{RR}}\right)\right)}\mathbf{)}\) &
\textbf{Data information Weight:} \(\mathbf{(}\frac{\mathbf{1}}{\mathbf{\text{data\ variance\ }}\left(\mathbf{\ln}\left(\mathbf{\text{RR}}\right)\right)}\mathbf{)}\)\tabularnewline
\midrule
\endhead
\textbf{(\(\frac{\mathbf{1}}{\mathbf{2}}\), 2)} & (-0.42, 0.88) & 8.3 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{3}}\), 3)} & (-0.44, 1.43) & 3.2 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{4}}\), 4)} & (-0.41, 1.77) & 2.0 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{5}}\), 5)} & (-0.38, 2.01) & 1.5 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{6}}\), 6)} & (-0.36, 2.18) & 1.2 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{7}}\), 7)} & (-0.33, 2.30) & 1.0 &
1.2\tabularnewline
\textbf{(\(\frac{\mathbf{1}}{\mathbf{10}}\), 10)} & (-0.28, 2.56) & 0.7 &
1.2\tabularnewline
\bottomrule
\end{longtable}

\textbf{Hosted file}

\verb`image1.emf`  available at \url{https://authorea.com/users/354031/articles/477715-a-framework-for-bayesian-posterior-simulation-methods-in-clinical-practice}

\textbf{Figure1} . Comparison of target and proposals distribution\selectlanguage{english}
\begin{figure}[H]
\begin{center}
\includegraphics[width=0.70\columnwidth]{figures/image2/image2}

\end{center}
\end{figure}

\textbf{Figure2.} As an example, which prior is not supported by data,
imagine prior (lower) as Normal (?=10, SD=2) and likelihood (upper)
Normal (0,1) is far from each other, most of the drawn samples from
prior get very small weights. The probability of sampling
from\(-3\sigma\) and lower (the parts that get higher weights) is
nearly 1\% so out of 10000 we expected 100 non-zero weights. Posterior
mean estimated equals 1.5 while we expected 0.01 (Posterior
\textasciitilde{} Normal (0.01, 0.01). Therefore, this approach proved
inefficient in terms of accuracy of estimate and number of sampling.

\selectlanguage{english}
\FloatBarrier
\end{document}

