This vignette describes the `SMC-Mallows`

functions of the
package. These use Sequential Monte Carlo (SMC) algorithms to provide
updated approximations to the posterior distribution of a single Mallows
model. We consider scenarios where we receive sequential information in
the form of complete rankings, partial rankings and updated rankings
from existing individuals who have previously provided a (partial)
ranking. We use an alternative data augmentation method, called the
pseudolikelihood approach, when we are using the footrule and Spearman
distance functions instead of using an independent sampler.
`SMC-Mallow`

uses functions similar to their base MCMC
counterparts in the `BayesMallows`

package, to visualise and
analyse the posterior distributions.

For an in-depth treatment of the implemented methodology, see Stein (2023).

Function Name | Description |
---|---|

`smc_mallows_new_users` |
Runs the SMC algorithm for case where we observe full rankings as new observational data. |

`smc_mallows_new_item_rank` |
Runs the SMC algorithm for case where we observe updated partial rankings as from existing users. |

`plot` |
Plots posterior density of \(\boldsymbol{\rho}\) or \(\alpha\) for a selection of items. |

`compute_consensus` |
Computes the CP estimate or MAP estimate of the latent ranks. |

`compute_posterior_intervals` |
Computes the Bayesian posterior intervals for \(\alpha\) and \(\boldsymbol{\rho}\). |

We provide a summary on the Bayesian Mallows model and the proposed
Sequential Monte Carlo framework which updates the parameter estimates
of the posterior each time we receive new observations for a fixed
computational cost. More information on the Bayesian Mallows model can
found in Vitelli et al. (2018) and Liu
et al. (2019), and a vignette on the
`BayesMallows`

R package can be found in Sørensen et al. (2020). A general discussion on SMC
can be found in Del Moral, Doucet, and Jasra (2006) and Doucet and Johansen (2009).

We have set of \(m\) distinct items, labelled \(\mathcal{A} = \{A_1, \dots, A_m \}\), and we are asked to rank these items in order of personal preference with respect to some attribute. This information can be expressed as a ranking \(\boldsymbol{R} = \{ R_1 , \dots , R_m\}\), which is a mapping \(\boldsymbol{R}:\mathcal{A} \rightarrow \mathcal{P}_m\), where \(\mathcal{P}_m\) is the space of \(m\)-dimensional permutations of \(\{1, \dots, m\}\). Each \(R_i \in \{1,\dots,m\}\) corresponds to the rank of an item \(A_i\). We fundamentally assume that the preference information we receive from a group of individuals is transitive, i.e., each individual does not contradict themselves when specifying their preferences. In other words, for any three distinct items \(\{A_i,A_j,A_k\}\) in a set, then if \(A_i \prec A_j\) and \(A_j \prec A_k\), then it must follow that \(A_i \prec A_k\). Sometimes, we are unable to provide full rankings, so instead we provide a ranking for a subset of the items in \(\mathcal{A}\). These are referred to as partial rankings. Partial rankings can occur either randomly or an individual can specify their top-\(k\) ranked items. In this scenario, we will need perform data augmentation in order to estimate the parameters of the Mallows model.

The Mallows model (Mallows 1957) is a probability distribution for ranking data. The probability of observing a ranking \(\boldsymbol{R}\) is defined as

\[p(\boldsymbol{R}) = p(\boldsymbol{R}|\boldsymbol{\rho},\alpha) =\frac{1}{Z_m(\alpha)} \exp \lefts\{ -\frac{\alpha}{m} { d(\boldsymbol{R}, \boldsymbol{\rho})} \right\},\]

where: \(\boldsymbol{\rho} \in
\mathcal{P}_m\) is the consensus ranking; \(\alpha > 0\) is the scale parameter
which represents the variability in rankings within the group of
individuals around the consensus ranking; and \(Z_m(\alpha)\) is the normalisation
constant. The distance function, \(d(\cdot,\cdot) : \mathcal{P}_m \times
\mathcal{P}_m \rightarrow [0,\infty)\), measures the
‘’closeness’’ of a ranking to the consensus ranking. The Mallows
literature discusses the use of a right-invariant distance function,
which means that the distance between two items is unaffected by
relabelling of items (Diaconis 1988). The distance metrics
that the `BayesMallows`

R package currently uses are:
footrule, Spearman, Cayley, Kendall and Hamming. This also means that
the normalisation constant is independent of the consensus ranking.

Vitelli et al. (2018) extended the Mallows model to incorporate a Bayesian framework for inference. A uniform prior is elicited for the consensus ranking \(\pi(\boldsymbol{\rho}) = (m!)^{-1} 1_{\mathcal{P}_m} (\boldsymbol{\rho})\) in the space of \(\mathcal{P}_m\), and an exponential prior for \(\alpha\), with density \(\pi(\alpha|\lambda) = \lambda \exp \{ -\lambda \alpha \} 1_{[0,\infty)}(\alpha)\). Given \(M\) observed complete rankings and the prior distributions \(\pi(\boldsymbol{\rho})\) and \(\pi(\alpha)\), assuming prior independence of these variables, we have the following posterior density, known as the Bayesian Mallows model,

\[ \pi(\boldsymbol\rho, \alpha | \boldsymbol{R}_1, \dots, \boldsymbol{R}_M) \propto \frac{\pi(\boldsymbol\rho)\pi(\alpha) }{[Z(\alpha)]^M} \exp \left\{ - \frac{\alpha}{m} \sum_{j=1}^{M} d(\boldsymbol{R}_j, \boldsymbol\rho ) \right\}.\]

Any posterior estimates of interest, such as the marginal posterior for \(\boldsymbol{\rho}\), are obtained through the use of Metropolis-Hastings based Markov Chain Monte Carlo (MCMC) algorithm. Full details of the algorithm can be found in Vitelli et al. (2018). In each iteration of the algorithm, a new consensus ranking \(\boldsymbol{\rho}'\) is proposed to update \(\boldsymbol{\rho}\) according to a distribution which is centered around the current rank \(\boldsymbol{\rho}\). The proposal step for \(\boldsymbol{\rho}\) is done using the leap-and-shift proposal algorithm of Vitelli et al. (2018) and a new value \(\alpha'\) is sampled from the log-normal distribution to update the current value of \(\alpha\).

Inference on the Bayesian Mallows model can sample the posterior
distribution of the unknown consensus ranking and scale parameter using
a variety of observed data including: full rankings, incomplete rankings
(e.g. top-\(k\) rankings and ranks
missing at random), and implicit data such as pairwise comparisons. For
example, in the case of partial rankings, we can create augmented full
ranking \(\tilde{R}_1, \dots,
\tilde{R}_M\) by using an independent sampler for each assessor
containing the set of rankings not already chosen. The MCMC algorithm
alternates between sampling a new value of \(\boldsymbol{\rho}\) and \(\alpha\) given the current \(\tilde{R}_1, \dots, \tilde{R}_M\) and
sampling \(\tilde{R}_1, \dots,
\tilde{R}_M\) given the current values of \(\boldsymbol{\rho}\) and \(\alpha\). The existing methods are
discussed in Vitelli et al. (2018) and are provided in the
`BayesMallows`

R package (Sørensen et al. 2020).

Sequential Monte Carlo (SMC) methods are a class of sampling algorithms which are used to estimate a sequence of target distributions given a stream of observations over discrete time. Each target distribution is approximated by a collection of random samples, termed particles, at each time step and evolve according to importance sampling and resampling steps. The literature on SMC methods is vast and diverse, but we are interested in using SMC as an alternative to MCMC methods (Chopin 2002). A nice summary of the different variants of SMC is given in Del Moral, Doucet, and Jasra (2006).

In SMC, the aim is to approximate a sequence of target distributions \(\pi_t(\boldsymbol{\theta})\) with parameters \(\boldsymbol{\theta}\) that we wish to estimate given a set of observed data \(D_t\) which has accumulated up to time \(t\). So we can write a target distribution \(\pi_t\) as a posterior distribution

\[\pi_t(\boldsymbol{\theta}) = \pi_t(\boldsymbol{\theta} | D_t) \propto \pi_0(\boldsymbol{\theta})p_t(D_t| \boldsymbol{\theta}).\]

The SMC algorithm begins by generating \(N\) particles using the prior distributions for each parameter and assigning each particle an equal weight. In each time step \(t=1,\dots,T\), we assume that an additional \(p\) independent observations \(y_{1:p}\) become available from the target distribution. We reweight the particles in \(\pi(\boldsymbol{\theta}| D_{t-1})\) from time \(t-1\) to \(t\) such that they are weighted with respect to \(\pi(\boldsymbol{\theta}| D_{t})\),

\[ w^{(i)}_t = \frac{\pi(\boldsymbol{\theta}^{(i)}_{t-1} | D_{t})}{\pi(\boldsymbol{\theta}^{(i)}_{t-1} | D_{t-1})} \propto \frac{p(D_t | \boldsymbol{\theta}^{(i)}_{t-1})}{p(D_{t-1} | \boldsymbol{\theta}^{(i)}_{t-1})} = p( y_{1:p}| D_{t-1}, \boldsymbol{\theta}_{t-1}^{(i)} ), \ i = 1,\dots,N.\]

Next, we normalise the particle weights and resample the particles with replacement which replicates the heavier weighted particles and discard those with negligible weights. This results in a set of equally weighted particles \(\{ \boldsymbol{\theta}_t^{(i)}, w_t^{(i)} = \frac{1}{N} \}_{i=1}^N\) that represent a sample of the posterior distribution. A summary of the possible methods for resampling is given in Doucet and Johansen (2009).

In the final stage, we move the particles using an MCMC kernel within SMC after resampling to give back the diversity of particle values (Berzuini and Gilks 2001). This particular methodology in SMC is often referred to as the Resample-Move framework of Berzuini and Gilks (2001) and Berzuini and Gilks (2003). We can apply MCMC kernel many times as we would like since the particles are still moving within the stationary distribution \(\pi_t\).

The `SMC-Mallows`

functions contain algorithms to perform
the Resample-Move SMC framework of Berzuini and
Gilks (2001) using a
single Mallows model. Each algorithm begins by drawing \(N\) particles using the priors for \(\boldsymbol{\rho}\) and \(\alpha\) or by using specified initial
values. Each particle is also assigned equal weight so at the start of
the SMC algorithm we have \(\{\boldsymbol{\theta}^{(i)}_0 =
(\boldsymbol{\rho}_0^{(i)}, \alpha_0^{(i)}), w^{(i)}
\}_{i=1}^{N}\) as the set of particles. Next, we observe some
ranking data, \(D_t\), and we calculate
the updated weights of the particles with respect to the new
observations and their contribution to the current estimated posterior
distribution before reweighting and multinomial resampling. Finally, we
perturb the particles using the Metropolis-Hastings based MCMC kernel
and we use the proposal distributions described in Vitelli et al. (2018) for sampling values of \(\boldsymbol{\rho}\) and \(\alpha\).

For this case, we assume that we observe a collection of complete rankings from new assessors over a sequence of discrete time steps, \(t=1,\dots, T\), such that up to a time \(t\), we will have observed \(|M_t|\) complete rankings. The particles are reweighted such that they are representative of the underlying distribution of the \(|M_t|\) complete rankings. The new weights for each particle are calculated as

\[\begin{align*} {w}^{(i)}_t(\boldsymbol{\theta}^{(i)}_{t-1}, \boldsymbol{\theta}^{(i)}_{t}) &= \frac{ (Z_m(\alpha^{(i)}_{t-1}))^{-|M_t|} \exp \left\{ - \frac{\alpha^{(i)}_{t-1}}{m} \sum_{j=1}^{|M_t|} d(\mathbf{R}^{(i)}_j, \boldsymbol{\rho}^{(i)}_{t-1} ) \right\} }{ (Z_m(\alpha^{(i)}_{t-1}))^{-|M_{t-1}|} \exp \left\{ - \frac{\alpha^{(i)}_{t-1}}{m} \sum_{j=1}^{|M_t|} d(\mathbf{R}^{(i)}_j, \boldsymbol{\rho}^{(i)}_{t-1} ) \right\} } \\ &= (Z_m(\alpha^{(i)}_{t-1}))^{-(|M_t|-|M_{t-1}|)}\exp \left\{ - \frac{\alpha^{(i)}_{t-1}}{m} \sum_{j= |M_{t-1}|+1}^{|M_t|} d(\mathbf{R}^{(i)}_j, \boldsymbol{\rho}^{(i)}_{t-1} ) \right\} , \end{align*}\]

where \(\alpha^{(i)}_{t-1}\) and \(\boldsymbol{\rho}^{(i)}_{t-1}, \ i=1,\dots,N\) are the current estimated parameter values of the Mallows posterior before we reweight.

We are interested in updating the parameter estimates of the Bayesian
Mallows model based on the existing ranking data and new observations.
We demonstrate the `SMC-Mallows`

functions using the
`sushi_rankings`

dataset (Kamishima 2003), which
contains 5000 rankings for 10 sushi dishes.

```
## shrimp sea eel tuna squid sea urchin salmon roe egg fatty tuna tuna roll
## [1,] 2 8 10 3 4 1 5 9 7
## [2,] 1 8 6 4 10 9 3 5 7
## [3,] 2 8 3 4 6 7 10 1 5
## [4,] 4 7 5 6 1 2 8 3 9
## [5,] 4 10 7 5 9 3 2 8 1
## [6,] 4 6 2 10 7 5 1 9 8
## cucumber roll
## [1,] 6
## [2,] 2
## [3,] 9
## [4,] 10
## [5,] 6
## [6,] 3
```

The function `smc_mallows_new_users`

with
`type = "complete"`

shows the simplest version of the SMC
algorithm in action. Many of the input variables, that the function
requires, are also used in some of the existing functions in the
`BayesMallows`

R package. However, we will discuss what the
new variables refer to in the algorithm. The variable `R_obs`

in this instance is a 2D dataset with the number of columns as
`n_items`

and the number of rows, \(M\), represents the number of individuals
in the dataset. We begin the algorithm with no observations and in each
artificial time step we introduce a batch number of observations, this
is controlled by the variable `num_new_obs`

to control the
size of the bath, and `Time`

to specify the number of updates
we will perform. There is a limit on the value for `Time`

,
and this depends on the number of individuals in the dataset and the
value of `num_new_obs`

. In the move stage we can apply an
MCMC kernel as many times as we would like using the variable
`mcmc_kernel_app`

to specify the value.

```
n_items <- ncol(sushi_rankings)
metric <- "footrule"
logz_list <- prepare_partition_function(metric = metric, n_items = n_items)
data <- sushi_rankings[1:100, ]
leap_size <- floor(n_items / 5)
N <- 1000
Time <- 20
smc_test <- smc_mallows_new_users(
R_obs = data, type = "complete", n_items = n_items,
metric = metric, leap_size = leap_size,
N = N, Time = Time,
logz_estimate = logz_list$logz_estimate,
cardinalities = logz_list$cardinalities,
mcmc_kernel_app = 5,
num_new_obs = 5,
alpha_prop_sd = 0.5,
lambda = 0.15,
alpha_max = 1e6
)
```

The example `smc_test`

returns a list variable of the
particles values for \(\boldsymbol{\rho}\) and \(\alpha\) at each time step, so we can
observe the posterior as it evolves. Specifically the list contains a
three dimensional matrix of size `N`

by `n_items`

by `Time+1`

, named `rho_samples`

, and an
`N`

by `Time+1`

matrix called
`alpha_samples`

. These matrices can be studied using some
post-processing functions for visualising and analysing the posterior.
It is also possible to make comparative results using the MCMC algorithm
and the SMC algorithm. Unlike MCMC we cannot observe the trace plots and
we do not need to specify a non-zero value for the burn-in. The indexing
in R means that we have to view the `Time+1`

slice of the
output in order to view the posterior once all 100 rankings have been
observed.

Here, we can observe the posterior probabilities of \(\boldsymbol{\rho}\) for a selection of
items can be observed by calling the function `plot()`

.

The posterior distributions of \(\boldsymbol{\rho}\) and \(\alpha\) can be studied with some of the
the analysis tools provided by `SMC-Mallows`

. The posterior
intervals for the consensus ranking of each sushi item are obtained by
calling `compute_posterior_intervals`

.

```
## item parameter mean median conf_level hpdi central_interval
## 1 Item 1 rho 3.601 4 95 % [2,5] [2.000,5.000]
## 2 Item 10 rho 9.982 10 95 % [10] [10.000]
## 3 Item 2 rho 4.235 4 95 % [2,5] [2.000,6.000]
## 4 Item 3 rho 2.191 2 95 % [2,3] [2.000,4.000]
## 5 Item 4 rho 6.548 6 95 % [6,8] [5.000,8.000]
## 6 Item 5 rho 7.713 8 95 % [6,9] [6.000,9.000]
## 7 Item 6 rho 4.048 4 95 % [3,5] [2.000,5.000]
## 8 Item 7 rho 8.936 9 95 % [8,9] [8.000,9.000]
## 9 Item 8 rho 1.000 1 95 % [1] [1.000]
## 10 Item 9 rho 6.746 7 95 % [6,8] [6.000,8.000]
```

We can also rank the sushi items according to their cumulative
probability (CP) consensus and their maximum posterior rankings (MAP).
These are calculated by calling the function
`compute_consensus_rho`

. We demonstrate with the CP
consensus, which is the default.

```
## ranking item cumprob
## 1 1 Item 8 1.000
## 2 2 Item 3 0.872
## 3 3 Item 1 0.451
## 4 4 Item 6 0.601
## 5 5 Item 2 0.955
## 6 6 Item 4 0.553
## 7 7 Item 9 0.937
## 8 8 Item 5 0.919
## 9 9 Item 7 0.983
## 10 10 Item 10 1.000
```

Similarly, we can observe the posterior density and the posterior
intervals for the scale parameter using the functions
`plot()`

and `compute_posterior_intervals`

.

```
## parameter mean median conf_level hpdi central_interval
## 1 alpha 1.697 1.692 95 % [1.447,1.966] [1.446,1.965]
```

In Vitelli et al. (2018), the augmentation scheme for partially observed rankings is done by using an independent sampler conditioned in the observed component of each ranking. The drawback of this approach is that it does not take into account any existing information, such as the current estimates of the consensus ranking and the scale parameter. We have an option to use an alternative augmentation kernel where, for each item, we use a univariate Mallows distribution to select each rank to an item based on the item’s rank in the estimated consensus ranking and the scale parameter. This approach is similar to the importance sampling approximation of the normalisation constant of the Bayesian Mallows model, full details can be found in Vitelli et al. (2018). This particular augmentation method only applies if we are using the footrule and Spearman distance metric.

We augment the missing item ranks in each \(\mathbf{R}\) to create a complete auxiliary ranking \(\tilde{\mathbf{R}}\) in order to perform the remaining steps of the SMC algorithm. We reweight the particles such that are representative of the underlying distribution of the \(|M_t|\) augmented rankings. The particle weights are recalculated as

\[\begin{align*} {w}^{(i)}_t(\boldsymbol{\theta}^{(i)}_{t-1}, \boldsymbol{\theta}^{(i)}_{t}) &= (Z_m(\alpha^{(i)}_{t-1}))^{-(|M_t|-|M_{t-1}|)}\exp \left\{ - \frac{\alpha^{(i)}_{t-1}}{m} \sum_{j= |M_{t-1}|+1}^{|M_t|} d(\tilde{\mathbf{R}}^{(i)}_j, \boldsymbol{\rho}^{(i)}_{t-1} ) \right\} \\ & \times \prod_{j = |M_{t-1}|+1}^{|M_t|} q(\tilde{\mathbf{R}}^{(i)}_j | \mathbf{R}_j, \boldsymbol{\rho}_{t-1}^{(i)}, \alpha_{t-1}^{(i)} ), \end{align*}\]

where \(\alpha^{(i)}_{t-1}\) and \(\boldsymbol{\rho}^{(i)}_{t-1}, \ i=1,\dots,N\) are the current estimated parameter values of the Mallows posterior before we reweight. The distribution \(q()\) represents the probability of creating an augmented ranking \(\tilde{\mathbf{R}}\) given the observed ranking \(\mathbf{R}\) and the current estimated parameters of the posterior.

For this demonstration we shall assume that we can only observe the
top-5 ranked items from each user in the first 100 rows of the
`sushi_rankings`

dataset.

```
## shrimp sea eel tuna squid sea urchin salmon roe egg fatty tuna tuna roll
## [1,] 2 NA NA 3 4 1 5 NA NA
## [2,] 1 NA NA 4 NA NA 3 5 NA
## [3,] 2 NA 3 4 NA NA NA 1 5
## [4,] 4 NA 5 NA 1 2 NA 3 NA
## [5,] 4 NA NA 5 NA 3 2 NA 1
## [6,] 4 NA 2 NA NA 5 1 NA NA
## cucumber roll
## [1,] NA
## [2,] 2
## [3,] NA
## [4,] NA
## [5,] NA
## [6,] 3
```

We can call the function `smc_mallows_new_users`

with
`type = "partial"`

to run the SMC algorithm for partial
ranking data. The variable `aug_method`

allows you to choose
which data augmentation method to use on the partial ranking data. The
option `"pseudolikelihood"`

is only compatible with when
selecting the distance metric as either `"footrule"`

or
`"spearman"`

.

```
aug_method <- "pseudolikelihood"
metric <- "cayley"
# example of selecting the incorrect combination of metric and aug_method
smc_partial_test <- smc_mallows_new_users(
R_obs = data_partial,
type = "partial",
n_items = n_items,
metric = metric,
leap_size = leap_size, N = N,
Time = Time,
logz_estimate = logz_list$logz_estimate,
cardinalities = logz_list$cardinalities,
mcmc_kernel_app = 5,
num_new_obs = 5,
alpha_prop_sd = 0.5,
lambda = 0.15,
alpha_max = 1e6,
aug_method = aug_method
)
```

This time we call the function again but this time we provide a valid combination of the augmentation method and the distance metric.

```
aug_method <- "pseudolikelihood"
metric <- "footrule"
smc_partial_test <- smc_mallows_new_users(
R_obs = data_partial,
type = "partial",
n_items = n_items,
metric = metric,
leap_size = leap_size, N = N,
Time = Time,
logz_estimate = logz_list$logz_estimate,
cardinalities = logz_list$cardinalities,
mcmc_kernel_app = 5,
num_new_obs = 5,
alpha_prop_sd = 0.5,
lambda = 0.15,
alpha_max = 1e6,
aug_method = aug_method
)
```

The variable `smc_test_partial`

contains a list with three
dimensional matrix of size `N`

by `n_items`

by
`Time+1`

, named `rho_samples`

, and a
`N`

by `Time+1`

matrix called
`alpha_samples`

. The analysis performed in the previous
demonstration can be applied to this scenario. Here, we can observe the
posterior probabilities for a selection of items in \(\boldsymbol{\rho}\) and posterior density
for \(\alpha\) as part of this
demonstration, but we can use other post processing functions to analyse
the output.

We can view this scenario as a observing an updated partial ranking from an known individual, but a previously latent ranking for an unranked items becomes known. For example, an individual may provided a partial ranking for 6 items \((1,2,3,\texttt{NA},\texttt{NA},\texttt{NA})\) and in the SMC algorithm we might have obtained an augmented ranking \((1,2,3,5,6,4)\). However, later on the same individual may have provided more information about their ranking, e.g., \((1,2,3,4,\texttt{NA},\texttt{NA})\), and this ranking is no longer consistent with the augmented ranking. This causes a problem where we have an augmented complete ranking through SMC, conditioned on the original observed partial ranking, that happens to be no longer consistent with the new observed partial ranking. This means that our algorithm has an additional step before the reweighting stage. In order to reweight the particles correctly as a representative sample of the current posterior, we view the new information arriving into the system as an existing individual (and their augmented ranking) leaving the system completely and then re-entering as a new partial ranking with its extra information. For each individual leaving and re-entering the system, we have to make two weight adjustments to account for these actions.

To illustrate how to perform SMC for updated partial rankings, we
will modify a much smaller dataset called `potato_visual`

.
This dataset, which is described in Liu et al.
(2019) and is provided in the
`BayesMallows`

R package, represents the rankings given by 12
assessors on 20 potatoes based on how heavy each potato appeared to be
given their visual appearance. We create several partial subsets of the
complete dataset by removing the lowest ranked item from each assessor
from each previous subset, so that we achieve several partial datasets
which view the top 10, top 11,… top 19 highest ranked items as well as
the complete set of rankings.

```
example_dataset <- potato_visual
n_users <- 12
n_items <- 20
test_dataset <- array(0, c(n_users, n_items, (n_items / 2 + 1)))
test_dataset[, , (n_items / 2 + 1)] <- potato_visual
tt <- 0
for (ii in (n_items - 1):(n_items / 2)) {
tt <- tt + 1
# set n_users line with one more NA
example_dataset[example_dataset > ii] <- NA
# set as new time stamp
test_dataset[, , ((n_items / 2 + 1) - tt)] <- example_dataset
}
```

This means that we now have a 3D array containing 10 partial datasets
and the one complete dataset. The third dimension of the array
represents artificial time. We can view the updated partial rankings
scenario as viewing several 2D slices of the observed dataset
sequentially. So you can see, for example, that at the 5th time point,
we observe the top-14 items from `potato_visual`

.

```
## [,1] [,2] [,3] [,4] [,5] [,6] [,7] [,8] [,9] [,10] [,11] [,12] [,13]
## [1,] 10 NA NA NA 6 NA 4 NA 3 5 12 1 2
## [2,] 10 NA NA NA 11 NA 6 NA 4 3 13 1 2
## [3,] 12 NA NA NA 13 11 7 NA 6 3 8 2 1
## [4,] 9 NA NA NA 10 NA 5 NA 3 4 8 1 2
## [5,] 12 NA NA NA 7 NA 2 NA 3 9 13 1 4
## [6,] 10 NA NA NA 8 NA 6 NA 3 7 11 1 2
## [7,] 9 NA NA NA 10 NA 5 NA 3 8 11 1 2
## [8,] 14 NA NA NA 11 NA 6 NA 4 3 10 1 2
## [9,] 8 NA NA NA 12 13 6 NA 5 3 7 1 4
## [10,] 7 NA NA NA 9 NA 5 NA 3 10 11 1 2
## [11,] 12 NA NA NA 13 NA 7 NA 3 5 11 1 2
## [12,] 14 NA NA NA 12 NA 8 NA 3 4 9 1 2
## [,14] [,15] [,16] [,17] [,18] [,19] [,20]
## [1,] 9 NA 8 7 14 13 11
## [2,] 7 NA 8 5 12 9 14
## [3,] 4 NA 5 9 14 10 NA
## [4,] 7 NA 11 6 13 14 12
## [5,] 5 NA 11 6 8 10 14
## [6,] 4 NA 9 5 13 12 14
## [7,] 6 NA 7 4 14 12 13
## [8,] 7 NA 8 5 12 9 13
## [9,] 2 NA 10 9 NA 14 11
## [10,] 6 NA 8 4 13 12 14
## [11,] 6 NA 10 4 14 8 9
## [12,] 7 NA 6 5 13 10 11
```

We can now run an experiment with the altered
`potato_visual`

dataset by calling the function
`smc_mallows_new_item_rank`

.

```
logz_list <- prepare_partition_function(metric = metric, n_items = n_items)
Time <- dim(test_dataset)[3]
N <- 1000
aug_method <- "pseudolikelihood"
metric <- "footrule"
smc_test_updated_partial <- smc_mallows_new_item_rank(
n_items = n_items,
R_obs = test_dataset,
metric = metric,
leap_size = leap_size, N = N,
Time = Time,
logz_estimate = NULL,
cardinalities = logz_list$cardinalities,
mcmc_kernel_app = 5,
alpha_prop_sd = 0.5,
lambda = 0.15,
alpha_max = 1e6,
aug_method = aug_method
)
```

Unlike the two previous demonstrations, the final slice of the output
occurs at `Time`

instead of `Time+1`

. This is
because we initialised the algorithm with the first slice of the
`test_dataset`

rather than initialising with no observed
data. We observe the posterior probabilities for items in \(\boldsymbol{\rho}\) and posterior density
for \(\alpha\) by using the same
post-processing functions as before.

Berzuini, Carlo, and Walter Gilks. 2001. “RESAMPLE-MOVE Filtering
with Cross-Model Jumps.” In *Sequential Monte Carlo Methods in
Practice*, 117–38. Springer.

Berzuini, Carlo, and Walters R Gilks. 2003. “Particle Filtering
Methods for Dynamic and Static Bayesian Problems.” *OXFORD
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