2012 e4

1
Stockouts and Restocking:
Monitoring the Retailer From the Supplier’s Perspective
Abstract:
Suppliers and retailers typically do not have identical incentives to avoid stockouts. Thus,
the supplier needs to monitor the retailers restocking efforts with the available data. We
introduce a general model for this purpose and illustrate it using a specific application
provided by a supplier to a national grocery chain. The model distinguishes between store
stockouts (zero inventory in the store) vs. shelf stockouts (an empty shelf, but some inventory
in other parts of the store), thereby identifying the cause of the stockout to be either a supply
chain or a restocking issue. We find that the average stockout rates vary widely between
stores, identifying two stores with stockout rates twice as high as for most other stores.
Moreover, almost all stockouts are shelf stockouts. Thus, the model identifies stores that
may have restocking issues. Stockouts lead to major losses in expected sales. Finally, we also
find major difference in shrinkage between stores, providing useful information to regional
store managers.
Keywords: Out-of-Stock, Inventory, Bayesian Estimation, Particle Filter

2
INTRODUCTION
It is well known that stockouts have a major impact on profits (e.g. Gruen and Corsten
2008, Anderson et al. 2006). Given differences in the retailer’s and the supplier’s profit
functions, the retailer’s and the supplier’s incentives in trying to avoid stockouts are generally
not perfectly aligned. Imagine a situation in which a retailer carries two brands of a certain
product, and consumers exhibit only weak brand preferences. If one of the brands is out of
stock, the consumer will most likely simply buy the other brand rather than go to a different
store or wait for his next shopping trip. Campo et al. (2003) conclude that retailers’ losses
from stockouts can remain very limited, while the suppliers’ losses will be substantial. Thus,
the retailer may expend less effort than desired by the supplier.
Given this discrepancy in interests, it is essential for the supplier to estimate stockouts
in order to monitor the retailer’s stocking efforts. Yet, physically checking the shelves in the
store is typically expensive. Thus, the supplier needs to deduce the occurrence of stockouts
from the available data. More importantly, the supplier also needs to know the root cause
of the stockouts: If stockouts happen because there are no products in the store (a “store
stockout”), the issue lies within the supply chain and the supplier itself may work on fixing
the problem. However, if there is a stockout despite positive inventory in the store (a “shelf
stockout”), the problem lies with the retailer who does not restock the shelves quickly enough.
In the latter case, the supplier may want to urge the retailer to improve his processes.
In this paper, we present a model that allows the supplier to (a) determine the amount
of stockouts and (b) distinguish the two types of stockouts, using only shipment and sales
data that is readily available to the supplier. The model consists of two interrelated models,
a sales model and an inventory model. The sales model yields daily estimates of stockout
probabilities for each product at each store. The inventory model probabilistically tracks
daily inventory levels, allowing for unobserved shrinkage. While the amount of stockouts
could be estimated using a macro-level framework, the micro-level (i.e., daily or even shorter

3
timeframe) analysis is necessary to link stockouts and inventory levels in order to distinguish
the two types of stockouts.
Marketing studies concerning stockouts typically use data with information about prod-
uct availability on the shelf based on physical checks and focus on estimating the effects
of stockouts on demand and the resulting substitution patterns (e.g., Anupindi et al. 1998;
Kalyanam et al. 2007; Musalem et al. 2010). In contrast, our model structure is crucial for
a context where the supplier monitors the retailer. Instead of relying on information about
the occurrence of stockouts as an input to the model, it estimates, among other things, the
probability that a stockout occurred on a given day.
The results of our model not only indicate that the stockout rates vary widely across
stores but our model also shows sales losses to be quite large when products are out of
stock, given that stockouts arise due to unexpected demand shocks. While seven out of 10
stores have stockout rates of less than 6%, two stores have stockouts rates of more than
10%. Thus, the model identifies stores that seem to have management issues leading to the
higher stockout rates. Over 95% of stockouts are shelf stockouts, a situation that is clearly
unacceptable to a supplier. In addition, we find that the loss in expected sales due to a
stockout is on average 60 to 80%, i.e., the the stockouts lead to major reductions of the
supplier’s profits.
GENERAL MODEL
In this section, we present the general model applicable to the problem described in
the introduction. This framework can be used for any situation in which a supplier needs
to monitor the retailer’s restocking efforts. Since the purpose of the model is application-
driven, the distributional assumptions on the functions of the general model should match the
structure of the data generating process of a given application. In contexts as here where
there is relatively little observed information, information on the structural relationships
between variables is essential to estimate the model. This is akin to the developments in the

4
marketing literature on choice models in which the hypothesized structure of the latent choice
process is directly modeled. This allows for insights into the unobserved choice process using
only limited amounts of observed data (i.e., choice outcomes). Rather than solely relying
on theory, the structure in our model can also come from information supplied by store
managers or from inspection of the data, as we show below.
Data Generating Process
Figure 1 gives an overview of the underlying data generating process, presented as the
path of an individual product through the system. The process starts with (observed) ship-
ments to the individual stores. Total inventory can be separated into backroom inven-
tory and shelf inventory. Backroom inventory refers to the products stored in the back of
the store, while shelf inventory refers to the products actually on the shelf at a given time.
Obviously, shipped products first become part of the backroom inventory, before turning into
shelf inventory through restocking. Positive demand during non-zero shelf inventory lead to
observed sales. However, if there is positive demand and zero shelf inventory, a stockout
occurs. Finally, products may also leave inventory without being registered as sold through
shrinkage. Examples for shrinkage are theft or spoilage.
<< Insert Figure 1 about here >>
Zero Inventory = Stockout & Stockout = Zero Inventory. Intuitively, there seems to
be a deterministic relationship between inventory and stockouts. Whenever there is zero
inventory, we must have a stockout; and whenever we have a stockout, we must have zero
inventory. However, neither of these statements is true.
First, notice that the definition of a stockout above invokes the notion of positive demand,
i.e., it is possible to have zero (shelf or total) inventory without observing a stockout. This
definition has the advantage of estimating the “extent of out of stocks that actually matter
to the retailer and the upstream supply chain members” (Gruen et al. 2002, p. 11) because
it focuses on the impact on sales. If there are no units on the shelf, but nobody is trying

5
to purchase the product at that time, neither the retailer’s nor the supplier’s profits are
affected.
Second, a stockout implies zero shelf inventory, but not zero total inventory as there may
well be products in the backroom at the time of a stockout. This subtle difference is at the
heart of the distinction of store stockouts and shelf stockouts.
Stockout = Zero Sales & Stockout = (Sales==Inventory). Similarly, one may intuit a
very simple relationship between stockouts and sales, namely that a stockout implies zero
sales. While it is true that sales are zero for the duration of a stockout, daily sales may not
be zero. In particular, there may be a stockout exactly because positive demand depleted
the available shelf inventory. However, this should not lead one to the erroneous conclusion
that a stockout implies that observed sales are equal to previous total inventory. Once again,
one needs to realize that there may still be backroom inventory despite a stockout.
Model
For the remainder of the paper, we use the subscripts s, v, and t to refer to the different
stores, products(vegetables in our application), and days (time), respectively. However, for
the presentation of the general model, we suppress both s and v.
Depending on whether there was a stockout or not, the observed sales are either equal
to the unobserved demand or less than the unobserved demand. We thus have
(1) SALESt =



Dt(Xt, γt) with prob. 1 − ρt
SOt(Xt, γt) < Dt(Xt, γt) with prob. ρt
where Dt(·) is the unobserved demand, while SOt(·) denotes sales on a day with a stockout.
Xt is a vector of variables affecting demand on a given day. This vector may include market-
ing variables like price and promotions as well as other factors. γt represents any unobserved

6
demand shocks. Obviously, these same factors may also aﬀect sales on a day with a stockout.
The probability of a stockout on day t, ρt, is a function of shelf inventory being zero or
not (however, as explained in the following paragraph, we only model total inventory) as
well as the unobserved demand shocks, i.e.
(2) ρt = f(I{INVsvt=0}, γt)
While zero inventory does not imply a stockout (see above), it certainly makes it more
likely. Similarly, large demand shocks may make stockouts more likely, as unexpectedly
large demand makes timely restocking less likely.
In order to be able to distinguish between store and shelf stockouts, we need to estimate
total inventory as time progresses. Ideally, one would like to distinguish shelf and backroom
inventory. However, that is not possible with only sales and shipment data. Thus, we restrict
ourselves to estimating total inventory, with the understanding that a stockout implies zero
shelf inventory (but not vice versa).
The progression of inventory is governed by the following equation:
(3) INVt = INVt−1 + SHIPt − SALESt − SHRINKt
In words, inventory at the end of today is equal to yesterday’s closing inventory plus today’s
shipments minus today’s sales and shrinkage. We refer to the balance of shipments minus
sales as the “observed part of inventory” (normalized such that the smallest value is equal
to zero). Since total inventory must be non-negative at all times, the joint distribution
of starting inventory INV0 and the vector of SHRINK, g, needs to ensure non-negative
inventory for all days.
Additional insight into the timing of shrinkage may be gained from directly modeling the
process of reordering. Note that shipment data also reveal order timing, provided that the

7
lag between order and delivery is known. If the timing of orders is influenced by the current
inventory levels (as in a pull strategy), observing orders provides information on inventory
levels, which in turn provides information on shrinkage. Thus, we let
(4) ORDERt ∼ h(INVt, Zt)
where Zt is a vector of other variables that may affect the ordering process. These rela-
tionships between the variables provide information to inform both model structure and
estimation, yielding enough information in our application to obtain estimates that are pre-
cise enough for inferences and managerial decisions.
APPLICATION
We now apply this general model to a specific application. The application is motivated
by a supplier to a national grocery chain. The supplier suspects high stockout rates in the
stores due to the misaligned incentives to avoid stockouts discussed in the introduction. We
first briefly describe the data, before specifying distributional assumptions on the general
model based on interviews with store managers and detailed inspection of the data.
Data
The data come from a supplier to a major US grocery store chain. It consists of daily
sales and shipment data for 181 days (from May 27, 2007 to November 23, 2007) from 10
grocery stores located in different US states for 17 different products in the frozen vegetable
category (for one of the stores we have data for only 16 of the 17 products; thus, in total we
have 169 time-series of 181 days each). These products include, for instance, frozen sweet
corn, frozen sweet peas, and frozen mixed vegetables. However, the grocery stores carry
frozen vegetables not only from this one supplier, but also from a competing supplier.
Summary statistics. Figure 2 is a histogram plot of sales across all stores, products, and
days.1
While the mode of sales is 2 and the mean is 5.05 units, the maximum of observed

8
sales is 64 units of one product in one store on one day, consistent with a category that
exhibits high positive demand shocks that could lead to stockouts. Mean daily sales vary
between stores and products. Tables 1 and 2 give mean daily sales per store and per product,
respectively. Mean daily sales per store range from 2.18 to 7.58. Mean daily sales per product
range from 2.11 for Garlic Peas and Mushrooms to 15.34 for Sweet Corn.
<< Insert Tables 1 and 2 about here >>
Stores receive shipments in multiples of 12, i.e. one case consists of 12 units. Summing
over all stores and products, we observe shipments on about one out of three days (36.4%).
The vast majority of shipments (89.3%) consist of only one case. Another 7.7% are two-case
shipments. The largest observed shipments consist of 6 cases. Not surprisingly, average ship-
ments per store and average shipments per product (in units) closely resemble the numbers
reported for average sales, as shown in tables 1 and 2, respectively.
Data preparation. For each day, we observe the number of units sold as well as the dollar
revenue per product and store. Thus, we calculate prices as dollar revenue divided by the
number of units sold. For days with no units sold, this is undefined. However, since prices
rarely change in the course of the time-series for a given product at a given store, we can fill
in the missing price data with great confidence.
Moreover, we are missing data for a total of 329 individual store-product-day points
(distributed over 15 of the 169 store-product time-series). In addition to this, we also exclude
some of the data for the following reason. The model assumes that the shipments follow
a pull strategy, i.e. products are ordered when inventory runs low. This is a reasonable
assumption for most of the data. However, for part of the data, the shipments are likely
the result of a push strategy, i.e. the national headquarters decided to send shipments
irrespective of current inventory levels, presumably for some kind of display or promotion.

9
In particular, this happens for five products (across all ten stores) towards the end of the
time series. During this period, shipment volumes increase significantly (up to 63 cases per
shipment), and consequently occur less frequently. The rare shipments in this period occur
on the same day for all stores. Since the inventory model informs the timing and amount
of shrinkage by linking the likelihood of orders to the unobserved inventory and shrinkage,
the independence of the shipments from inventory in this period defeats the purpose of the
inventory model. Thus, we restrict ourselves to the more common case of a pull strategy
and exclude the data from the period of the push strategy. For two products, this affects
the last month of the data; for the other three, we must disregard the last two months. In
total, this affects 2,704 of the 30,589 store-product-day combinations in our data.
Similar to the promotions at the end of the data, for some stores and products the
beginning of the time-series seems to be the week after some sort of promotions just ended.
This is evident as significant amounts of inventory are sold off during that time. In those
cases, no shipments are received during that period despite positive sales. Yet, we do not
exclude this data, as those observations still accord with our inventory model in that the
lack of shipments in this period is due to high inventory levels (as it would be when following
a pull-strategy), not to independence from inventory levels. Except for one product at one
store, sales for these periods do not differ from the rest of the data, suggesting that the
promotion had already ended. Figure 3 shows the sales pattern for the exception. Since it is
obvious that the sales in the first week are noticeably higher than for the remainder of the
time despite zero shipments during those days, we assume that this store ran the promotion
one week longer than the other stores. We therefore define a dummy variable for promotion
that equals one for this one week, store, and product, and zero for all other days, stores, and
products.
Model

10
Sales. We start the applied version of the model by specifying the distributional as-
sumptions on the two distributions combining to generate observed sales (cf. equation 1).
Sales are commonly modeled using the Poisson distribution (e.g., Anupindi et al. 1998; Nee-
lamegham and Chintagunta 1999). However, as we model the observed sales as a mixture of
two unobserved distributions, we cannot test a priori whether a Poisson distribution, in par-
ticular its property that the variance is equal to the mean, is appropriate for Dsvt and SOsvt.
We therefore choose to model both Dsvt and SOsvt by the more flexible Conway-Maxwell-
Poisson (CMP) distribution (Conway and Maxwell 1961; Shmueli et al. 2005). The CMP
distribution has two parameters: the first parameter, λ ≥ 0, can be interpreted as a measure
of central tendency, while the second parameter, ν ≥ 0, governs the level of dispersion. If
the decay parameter ν is 1, the CMP reduces to the Poisson distribution; if 0 ≤ ν < 1
(ν > 1) the CMP distribution has thicker (thinner) tails compared to the Poisson, modeling
overdispersed (underdispersed) data relative to the Poisson. Thus, the CMP distribution
allows for modeling both over- and under-dispersed data; the posterior distribution for the
ν parameter indicates whether the commonly used Poisson distribution would have been
appropriate.
We then let the unobserved demand be given by
(5) Dsvt ∼ CMP(λD
svt, νD
)
where λD
svt is a function of the observed demand shifters Xsvt and the unobserved demand
shocks γsvt. In particular, Xsvt includes variations in price as well as seasonality, weekend,
holiday, and promotion effects. We denote these by PRICEsvt, SEASONt, WEt, IDt
(Independence Day), LDt (Labor Day), THt (Thanksgiving), and PROMOsvt, respectively.

11
The measure of central tendency for the Dsvt distribution is then given by
(6)
log(λD
svt)|βsv, γsvt = β0sv + β1svlog(PRICEsvt) + β2svWEt + β3svlog(SEASONt)
+β4svIDt + β5svLDt + β6svTHt + β7svPROMOsvt + γsvt
where
(7) βmsv ∼ N(¯βmsv, σm) for m = 0, 1, ..., 6
and are mutually independent. We do not impose a hierarchy on β7sv since there is only one
store-product for which this parameter is meaningful.
As prices do not change at all for some products at some stores, we normalize the
log(PRICEsvt) such that its mean is equal to zero for each store-product time series to
avoid the problem of separate identification of β0sv and β1sv in those cases. Similarly, the
mean of log(SEASONt) is normalized to 0, where SEASONt is a sine curve with wavelength
of one year fitted to the average sales per day across all stores and products (excluding the
five products for which we excluded the end of the time-series, as this would have biased
the estimate for seasonality, as well as Thanksgiving Day due to its extraordinarily low
sales). log(SEASONt) then ranges from -.11 to .22. WEt, IDt, LDt, and THt are dummy
variables taking the value 1 if day t was a weekend day, Independence Day, Labor Day, or
Thanksgiving Day, respectively, and 0 otherwise. Figure 4 shows the average sales per day
with the fitted sine curve. Moreover, the weekend spikes in sales amount are obvious, as is
the expected low number of sales on Thanksgiving. The impacts of Independence Day and
Labor Day on average sales are not as strong. PROMOsvt is the dummy variable discussed
in the section on data preparation to capture the effect of the promotion.
If within-brand/across-product substitution effects were expected, equation 6 could be

12
extended to include incorporate those effects. However, the category in our application is
known to be one of the least brand-loyal in the U.S. (Mitchell 2011). The grocery chain in our
data does carry a second brand of frozen vegetables, making substitution more likely to be
between-brand/same-product (leading to the discrepancy of interests between supplier and
retailer discussed in the introduction). Since we cannot model between-brand substitution
with our data, we do not incorporate substitution effects in our model.
For the hierarchy on the unobserved demand shocks, γsvt, we use the same multivariate
normal distribution for each daily vector of the 170 store-product shocks, γt. We let the
variance of this distribution be given by a Kronecker product of two matrices S and V , a
10 by 10 and a 17 by 17 matrix, respectively.2
This achieves two goals: (1) It reduces the
dimensionality, so rather than having to work with a 170 by 170 covariance matrix, we can
work with two smaller matrices. And (2) it allows us to introduce some meaningful structure
into the random shocks. In particular, S represents covariances across stores, i.e., it captures
effects that may affect one or more products similarly across several stores like national
advertising campaigns. Similarly, V captures covariances across products, i.e., effects which
affect some or all products within the same store. Depending on the product category,
these correlations may be caused by differences in weather conditions at the different stores,
or other local events like community celebrations. Taking the Kronecker product of these
two matrices parsimoniously combines the two sources of covariance for each store-product
combination. We thus have
(8) γt|ω, S, V ∼ MVN(ω, S ⊗ V ) ∀t
A natural choice for modeling the observed sales in the presence of a stockout would
be to let them equal demand minus some lost sales (e.g., following a binomial distribution
conditional on demand). If demand was Poisson distributed (i.e., νD
= 1), this would result
in a Poisson distribution for SOsvt with parameter (1 − p)λD
svt (where p is the probability of

13
the binomial distribution for lost sales). For νD
= 1, this relationship still holds approxi-
mately. For computational speed, we therefore choose to model SOsvt directly with a CMP
distribution exhibiting the same level of dispersion as Dsvt. In addition, sales should be
higher despite a stockout on days on which demand was higher. We therefore also link the
first parameters of the SOsvt and the Dsvt distributions by a reduction parameter δsv. Taken
together, we get
(9) SOsvt ∼ CMP(δsvλD
svt, νD
)
The final piece of equation 1, the probability of a stockout on a given day, is described
in equation 2. We use a Beta distribution for function f in equation 2, i.e.
(10) ρsvt|asvt, bρ ∼ Beta(asvt, bρ)
where
(11) log(asvt) = a0s + a1sI{INVsvt=0} + a2sγsvt
and
(12) a0s ∼ N(¯a0, τ0)
Inventory. We now turn our attention to inventory, orders, and shrinkage. Rather than
specifying g, the joint distribution for starting inventory and daily shrinkage, directly, we
define it by a set of marginal distributions combined with a set of constraints (equations 17
and 18) to ensure non-negativity of inventory.
The marginal distribution for starting inventory is simply given by a prior based on
conversations with several store managers (see next subsection). The situation is slightly

14
more complicated for the marginal distributions of daily shrinkage.
Careful inspection of the data suggests that shrinkage seems to occur on both the unit-
and the case-level. While unit shrinkage (e.g., due to theft or misplaced items) is the prob-
ably more commonly thought of type of shrinkage, Figure 5 shows a likely instance of case
shrinkage. Figure 5 is a time-series of the observed part of inventory for one product at one
store; it suggests that a case may have disappeared around the apparent jump at day 100
(and potentially even a second case later in the time series). Causes for case shrinkage may
be improperly recorded deliveries or incorrect handling of a case at the point of receipt or
restocking (causing a whole case to thaw and go bad). Thus, we believe that unit shrinkage
is more likely to be caused by the consumer and case shrinkage is more likely to be caused
by store employees.
Total shrinkage on a given day is then given by SHRINKsvt = Su,svt + M · Sc,svt, where
the subscripts u and c refer to unit- and case-wise shrinkage, respectively, and M = 12 is
the number of units per case.
As (unit and case) shrinkage is also discrete and non-negative for all days, the Poisson (or
the NDB or the geometric) distribution again would be a standard choice for the marginal
distribution. However, due to the non-negativity constraint on inventory, the distribution of
shrinkage is actually truncated. Typically, truncation leads to proportionally increased prob-
abilities for the non-truncated values. Since the truncation differs from day to day, depending
on the remainder of the time-series, the probabilities of the non-truncated values would be
increased by different amounts for different days. This would lead to the paradoxical result
that low amounts of shrinkage are more likely on days with low inventory compared to days
with high inventory. In order to avoid that, we could implement a zero-inflated truncated
distribution, i.e. the sum of the truncated probabilities is added to the probability of zero
shrinkage. This keeps the absolute probabilities of positive amounts of shrinkage indepen-

15
dent of the available inventory; in turn, the probability of whether or not shrinkage occurs
at all depends on inventory, with shrinkage being less likely for lower amounts of inventory.
Therefore, we choose to implement a slightly more complex distribution for shrinkage.
In particular, we model the marginal distribution for both unit and case shrinkage in two
parts. The first part, a Bernoulli distribution, models the occurrence of shrinkage. Shrinkage
is likely to be very rare. Yet, when it occurs it is caused by some event (e.g., a thief being
in the store or a store employee being distracted) that may affect more than just one unit
or case. The second part, a CMP-binomial distribution, models the amount of shrinkage,
given that shrinkage occurred. The CMP-binomial distribution can be derived from the
CMP distribution and can be viewed as the result of correlated Bernoulli draws (rather than
independent Bernoulli draws which would lead to the standard binomial distribution; see
Shmueli et al. 2005). It has three parameters: the number of draws, the probability of a
success, and a parameter governing the association between the draws. If this third parameter
is between 0 and 1, the Bernoulli draws are positively correlated; if it is greater than 1, the
Bernoulli draws are negatively correlated. If it is 1, the Bernoulli draws are independent,
and the CMP-binomial distribution reduces to the standard binomial distribution.
The Bernoulli distribution keeps the probability of shrinkage independent of inventory,
while the CMP-binomial distribution keeps the (marginal) probability of each individual
available unit to be shrunk constant. The relative and absolute probabilities of different
amounts of shrinkage then vary with inventory. We think that it is more appropriate in this
context to keep (marginal) individual probabilities constant than to keep either relative or
absolute probabilities of the overall amount of shrinkage constant.
We thus have
(13) Sc,svt|bc,s, nc,svt, pc,s, νc ∼ Bernoulli(bc,s) · (CMP-Binomial(nc,svt, pc,s, νc) + 1)

16
(14) Su,svt|bu,s, nu,svt, pu,s, νu ∼ Bernoulli(bu,s) · (CMP-Binomial(nu,svt, pu,s, νu) + 1)
where the two Bernoullis are correlated with ηs. This correlation allows for the possibility
that one event can cause both unit and case shrinkage simultaneously. The number of draws
for the CMP-binomial are given by
(15) nc,svt = max(0, floor(
ÎNV svt
M
) − 1)
(16) nu,svt = min(M − 2, ÎNV svt − 1)
where ÎNV svt is the estimate of inventory before the occurrence of shrinkage for this day.
The -1 in equation 15 is due to the +1 in equation 13, i.e. the fact that if shrinkage occurs,
at least 1 case is shrunk. Similarly, the -2 (and the -1) in equation 16 is partly due to the
+1 in equation 14, but also to the fact that, for simplicity and identification, we assume
that the maximum amount of unit shrinkage is M − 1 (otherwise it will be counted as case
shrinkage).
Note that the parameters governing the probability of shrinkage occurring are defined
on a store-level. Thus, the model allows separating out stores that may have greater issues
with shrinkage.
These marginal distributions could result in negative inventory on some days. We there-
fore impose the following constraint on the resulting joint distribution to ensure non-negative
inventory for all days:
(17)
t
1
(Su,svt + M · Sc,svt + SALESsvt) ≤ INV0,sv +
t
1
SHIPsvt ∀s, v, t
However, for some days, we impose a stronger constraint, reflecting the belief that in-
ventory was very likely equal to zero on those days. This addition to the model again was

17
motivated by inspection of the observed part of inventory. See Figure 6 for an example
of a sales pattern and the corresponding observed part of inventory. The combination of
zero sales and the flat line in observed inventory from day 15 to 25 strongly suggests that
inventory in this store was zero on those days (and the store managers agreed). Yet, since
the observed part of inventory is one unit lower than the flat line on day 7, this implies that
there must have been at least one unit and/or one case (if starting inventory was higher
than in our plot) of shrinkage between day 8 and day 15. Thus, this stronger constraint can
greatly help the estimation of the occurrence of shrinkage. We determine the days for which
to use the stronger constraint by the following fairly strict criteria:
1. The future observed part of inventory must be non-decreasing.
2. We observe at least two weeks of future data.
3. The observed part of inventory is not higher than on the previous day.
4. We observe at least two consecutive days with zero sales.
Condition (1) is a necessary condition for zero inventory on a given day. Condition (2)
increases our confidence in asserting zero inventory for a certain day by making sure that we
observe enough future data to make condition (1) meaningful. Condition (3) rules out days
on which we know a shipment was received that may satisfy the other conditions. And finally,
condition (4) further tightens the criteria by requiring that the effect persists for at least
two days, thereby resulting in a visible flat line in a plot of the observed part of inventory,
to make it less likely that the observed pattern is just a coincidence of one day with zero
sales and zero shipments. Based on these criteria, we have 173 days for which to include
the stronger constraint. Referring back to Figure 5, notice that the flat line before day 100
also satisfies the above criteria (as sales are zero for three consecutive days), implying that
shrinkage in the preceding days is very likely.

18
Let all days satisfying the above criteria for store s and product v be denoted by T∗
sv.
The stronger constraint for those days still makes sure that inventory was non-negative, but
it also reﬂects the belief that inventory most likely was zero on those days:
(18)
T∗
sv
1
(Su,svt+M·Sc,svt+SALESsvt)



< INV0,sv + T∗
sv
1 SHIPsvt with prob. 1 − πsvT∗
sv
= INV0,sv + T∗
sv
1 SHIPsvt with prob. πsvT∗
sv
∀T∗
sv and ∀s, v
The top of the right hand side of equation 18 is the same non-negativity constraint as in
equation 17, while the lower part results in zero inventory on T∗
sv.
As discussed in the general model, we can use the placement of orders for further informa-
tion on the exact timing of shrinkage. (Moreover, modeling the ordering process is necessary
to estimate the missing data.) From talking to several local managers of the national grocery
chain, we know that the lag between the placing of the order and the receipt of the shipment
is typically two days. Therefore, the number of cases ordered on a given day is
(19) ORDERsvt =
SHIPsv,t+2
M
From inspection of the data, it is obvious that the distribution of orders is highly under-
dispersed relative to a Poisson distribution. We therefore again use the more ﬂexible CMP
distribution and let the distribution of orders be given by
(20) ORDERsvt|λO
svt, νO
∼ CMP(λO
svt, νO
)
As argued above, λO
svt then depends on inventory, providing information on shrinkage. In
particular, we include a variable calculated from today’s inventory plus tomorrow’s shipment
(which is already ordered and therefore known) relative to the average sales (denoted by
SALESsv) in the two days needed for the shipment to arrive. Moreover, λO
svt also depends

19
on the lagged dummies for the seasonal trend and for weekends (as defined above), as higher
orders should be expected when the store expects a demand spike two days later. The changes
of expected demand level and their effect on orders is captured by these two variables. Thus
we have
(21) log(λO
svt) = κ0sv − κ1sv
INVsvt + SHIPsv,t+1
2 · SALESsv
+ κ2svlog(SEASONsv,t+2) + κ3svWEsv,t+2
and
(22) κlsv ∼ N(¯κl, ςl) for l = 0, 2, 3
and
(23) κ1sv ∼ Gamma(¯κ1, ς1)
Notice the gamma-hierarchy on κ1sv; this ensures that the effect of inventory on orders is
negative, as it should be. See the Limitations section below for a discussion of this restriction.
Priors. Table 3 gives an overview of the priors we choose for the parameters.
<< Insert Table 3 about here >>
Sales: For ¯βm and σm (for m = 0, 1, ..., 6) we specify normal distributions with mean 0 and
variance 100 and an inverse gamma distribution with parameters 2.5 and 2.5, respectively,
as uninformative priors. The prior for β7 for the one store-product combination with the
promotion at the beginning of the time-series is also a normal distribution with mean 0 and
variance 100. Since exp(γsvt) has a multiplicative effect on λsvt, we can reasonably expect that
it is closely clustered around 1. In equation 8, we specify a multivariate normal distribution
for γsvt which is equivalent to a multivariate log-normal distribution for exp(γsvt). Since the
mode of a log-normal distribution is equal to the exponent of its mean minus its variance, we

20
choose a prior such that, in expectation, the mean is equal to the variance. In addition, the
variance should be fairly small since very large shocks are unlikely. We thus let the prior for
ω be a multivariate normal distribution with mean .001 and a variance conditional on S and
V given by the .001S⊗V . S and V are distributed according to inverse Wishart distributions
that combine, in expectation, to values of .001 on the diagonal of their Kronecker product.
We let the parameters for the inverse Wishart distribution for S be given by 150 and 4 · RS,
where RS is a correlation matrix with .2 for all off-diagonal elements. Similarly, for V the
parameters of the inverse Wishart distribution are 150 and 4 · RV , where RV is a correlation
matrix with .1 for all off-diagonal elements. The difference in off-diagonal elements reflects
our belief that shocks affecting sales of the same product across stores are slightly more likely
than shocks that affect different products in individual stores. Finally, while this choice of
priors for ω, S, and V does not retain conjugacy for S and V , it still allows for simple Gibbs
drawings of S conditional on V and vice versa.
As prior for νD
, we choose a gamma distribution with parameters 2 and 1, resulting in
a mode of 1 (which would reduce the CMP distribution to a Poisson distribution) but also
allowing over- and under-dispersion of the data. For δsv for all s and all v, we choose a beta
distribution with parameters 2 and 2.5. Thus, high values for δsv are unlikely, fitting the
idea that there should be an impact on sales for an empty shelf to be counted as a stockout.
Finally, for the priors related to ρsvt, the probability of stockouts, we let bρ follow a fairly
diffuse gamma distribution with parameters 2 and 1. Since Gruen and Corsten (2008) and
Andersen Consulting (1996) find that average rates of stockouts should not be significantly
greater than 10%, we choose a prior for ¯a0 that, while allowing for some uncertainty, ensures
that exp(a0sv) is small relative to bρ, namely a normal distribution with mean -2.5 and
variance .5. The variance τ0 follows an inverse Wishart distribution with parameters 4
and 2. For a1sv and a2sv (for all s and v) we choose gamma distributions, as both should
definitely have a positive effect. For a1sv, we choose parameters 30 and .125; for a2sv, we
choose parameters 10 and 3 (recall that all γsvt will be fairly small in absolute value).

21
Inventory: We believe that shrinkage should be fairly rare, with unit shrinkage being
somewhat more likely than case shrinkage. Thus, we let the priors for bu,s and bc,s be beta
distributions with parameters (1, 20) and (1, 30), respectively. We use a uniform uncondi-
tional prior for the correlation between unit and case shrinkage. (Note that conditional on
bu,s and bc,s the correlation may not be able to take all values from -1 to +1.) For the
CMP-binomial distribution, we let the probability for each unit or case to be shrunk, i.e.
pu,s and pc,s, be distributed according to a beta distribution with parameters 1 and 20 (all of
the above specifications are independently for all s). For the measure of correlation between
the Bernoulli draws, we use a gamma distribution with parameters 11 and .1 for both νu and
νc. Thus, a standard binomial distribution (i.e., no correlation between the Bernoulli draws)
is the prior mode (ν = 1), but the prior also allows for positive and negative correlation
between the Bernoulli draws.
As mentioned above, the criteria for the days on which we impose the stronger constraint
on the joint distribution for starting inventory and shrinkage are fairly strict, i.e. we are very
confident that inventory was very likely zero. Moreover, this probability of zero inventory,
πsvT∗
sv
, is set against the very small probabilities of shrinkage which also means that it needs
to be fairly large to be effective. Finally, the more days we observe after a day satisfying the
the criteria, the more confident we are that inventory actually was zero on that day. These
reasons lead to a beta distribution with parameters dsvT∗
sv
and 1 as a prior, where dsvT∗
sv
is
the number of days observed after T∗
sv until the next missing data point or the end of the
data.
We use the same uninformative priors for ¯κn and ςn for n = 0, 2, 3 as for ¯βm and σm in
the sales model, namely a normal distribution with mean 0 and variance 100 and an inverse
gamma distribution with parameters 2.5 and 2.5, respectively. For the gamma-hierarchy on
κ1sv, we specify two gamma priors, with ¯κ1 and ς2 both following a gamma distribution with
parameters 4 and .5. For νO
we use a gamma prior with parameters 2 and 2, reflecting
the finding from inspection of the data that orders seem to be underdispersed relative to a

22
Poisson distribution.
Finally, based on discussions with store managers, we choose a prior for starting inventory
with most of its mass around the size of one case, but also allowing for some variation
around it. Rather than using a Poisson distribution or the NBD, we choose a truncated
and discretized normal distribution. Since we observe the tail ends of a promotion for some
products with inventory being sold off (see section on data preparation), starting inventory
has to be in the tail of the marginal distribution in those cases in order to assure non-
negative inventory for all other days. We prefer the discretized normal distribution to the
Poisson and the NBD as its tail decreases more quickly, thereby still providing meaningful
information for starting inventory in those instances. We discretize the normal distribution
by letting the probability for integer x be given by the difference between the normal cdf at
x+0.5 and the normal cdf at x−0.5. Let DTyN denote this discretized normal distribution,
truncated from below at y. The prior for starting inventory (for all stores and products)
then is DT0N(M, (.75M)2
).
MODEL ESTIMATION
The model defined in equations (5) to (23) is estimated using an MCMC algorithm
(Casella and George 1992; Gelfand and Smith 1990). We define a binary auxiliary variable
zsvt to denote the occurrence of a stockout (Tanner and Wong 1987). Note that with several
parameters having to be estimated for all store-product-day combinations (ρsvt, zsvt, γsvt,
Su,svt, Sc,svt) we have far more than 150,000 parameters to estimate! This is feasible thanks
to the hierarchical structure of the model, resembling a frequentist random effects model.
After a burn-in period of 5,000 draws, we use 45,000 draws for the estimation. Convergence
was checked by the Heidelberger and Welch convergence diagnostic as implemented in the
boa package for R (Heidelberger and Welch 1983; Smith 2007).
The hierarchical parameters in equations 7, 12, and 22 are estimated in a Gibbs-sampler
fashion, as they have either conjugate priors, or priors leading to a closed-form condi-

23
tional distribution (for S and V ). Most of the remaining parameters are estimated with
a Metropolis-Hastings sampler.
However, the inventory trajectory, i.e. starting inventory, shrinkage, and missing data, are
estimated using a particle filter (see Arulampam et al. (2002) for a tutorial). In essence, we
construct a discrete representation of the posterior distribution conditional on the current
draw for all other parameters and then take a random draw from it. We implement this
approach because traditional Metropolis-like methods showed very slow mixing behavior
and/or took very long to compute. The particle filter results in a new draw from the
conditional posterior joint distribution for starting inventory, shrinkage, and the missing
sales and shipment data in every step of the MCMC. Thus, we can simply use it as a
drawing method in our Gibbs-sampler estimation.
The discrete representation of the posterior is achieved by recursively filtering a set of
n so-called particles according to importance weights updated in a Bayesian fashion. For
our application, this means we start by drawing the starting inventory for all particles and
update the weights accordingly, then we draw shrinkage (and the potentially missing data)
for day 1 and update the weights, then repeat those steps for day 2, and so on. The
updating of the weights follows a similar rationale as is implemented in the Metropolis-
Hastings sampler. Let φ(·) denote the proposal density from which shrinkage and, if miss-
ing, the missing data (or starting inventory for initializing the particles) are drawn. Let
ψt(Su,svt, Sc,svt, ORDERsvt, SALESsvt|·) denote the probability of observing Su,svt, Sc,svt,
ORDERsvt, and SALESsvt conditional on the other model parameters and the inventory
trajectory of that particular particle up to t, as defined by the model. Then the weight for
particle k after drawing day t is given by updating its weight after day t − 1 by
(24) wk,t = wk,t−1
ψt(·|·)
φ(·)
Thus, the greater the probability of the data as defined by the model, the greater the

24
particle’s weight; at the same time, the higher the probability of drawing the certain values,
the lower the weight. This is very similar in spirit to the Metropolis-Hastings algorithm.
The discrete approximation to the posterior distribution is then given by normalizing the
weights such that they sum to 1.
In order to avoid degeneracy of the algorithm (i.e., the problem that only one particle may
have significant weight compared to all the others) we implement a sequential importance
sampling (SIS) particle filter with resampling (Carpenter et al. 1999). In the SIS with
resampling algorithm, after each updating of the weights an estimate of the effective sample
size (i.e., the number of particles with non-neglegible relative weights), Neff , is calculated
by
(25) Neff =
( k wk,t)2
k w2
k,t
(Liu and Chen 1998). If Neff falls below a pre-specified threshold, a new set of n particles is
resampled randomly from the previous set of particles proportionally to their weights (also,
the weights are reset to 1/n for all particles after resampling). We implement this approach
with n = 150 particles and resample whenever Neff < 15.
In a final step, we randomly draw one of the particles representing the approximation to
the posterior distribution.
RESULTS
The results not only confirm that the suspicion of the supplier was at least partially
justified, but also give an idea of how much the supplier’s profits may be affected by the
situation.3
We first report the results for the structural parameters relating to the likelihood of sales
and to the likelihood of orders. Then we discuss the results for stockouts, before moving on
to the estimates for shrinkage obtained from the particle filter. Finally, we briefly analyze

25
some limitations of our model.
Sales And Orders
Table 4 presents the means and standard deviations for the hyper-parameters for the
CMP distribution for sales without a stockout and for the CMP distribution for orders, as
well as for νD
and νO
.
<< Insert Table 4 about here >>
First note that the CMP distribution for sales is highly overdispersed, as indicated by
the mean of νD
being .53. Thus, a model using the standard Poisson distribution for sales
would have been strongly misspecified. This overdispersion also means that small changes
in λD
svt lead to larger changes in the mean of the demand distribution, a fact to keep in mind
when analyzing the remaining parameters.
The results for ¯β1 to ¯β6 are as expected. Lower prices lead to higher demand (¯β1 = −.31;
values in parentheses refer to the mean); the relatively high variance across stores and
products for the price parameter (σ1 = .39) is due to the problem that for many store-
product combinations prices change only very rarely, if ever. This translates to a price
elasticity of -.4 to -.6 (depending on the amount of average sales, due to the non-linearity
of the CMP distribution for demand). This elasticity is well within the range of commonly
observed price elasticities, though - not surprisingly given the category and low prices -
towards the more inelastic side (see Bijmolt et al. (2005) for a meta-analysis). Also, demand
is 40-60% higher on weekends (¯β2 = .26) and increases as the season progresses towards the
winter time (¯β3 = .55). This translates to an elasticity of demand with respect to season
of about .7 to 1 (however, given the wide range of β3 (σ3 = .62) this may vary more for
individual store-product combinations). It is reassuring, though, that the elasticities are
close (on average) to 1, which would be the expected average value given how SEASON
was defined.
The holidays have different effects on demand: On Independence Day, demand is 30-40%

26
lower on average (¯β4 = −.27); Labor Day, in contrast, is probably regarded as an extra
weekend day to do shopping by many people, leading to a weak (10-20%) positive effect on
demand (¯β5 = .08; the lower bound of the 95% highest probability density interval is .008).
Finally, Thanksgiving has a very strong (80-90%) negative effect since most likely the stores
were closed for at least part of the day (¯β6 = −1.27).
The posterior means of the means for the multivariate normal distribution for the random
shocks (ω) range from .00079 to .00153, with an overall mean of .00104. The variances (i.e.,
the diagonal of S ⊗ V ) have posterior means ranging from .00090 to .00137, with an overall
mean of .00109. Thus, variances are very close to the means, leading to modes around one
for exp(γsvt).
In case of a stockout, δsv governs how much the distribution of sales is shifted to the left
due to the stockout. On average, δsv is .44, translating to a decrease in expected sales of 60
to 80% for each stockout. The posterior means for individual store-product combinations
range from .05 to .76. This means that in the worst case expected sales decrease by 98%,
but even in the best case expected sales drop by 39% due to a stockout!
In contrast to the distribution for sales, the CMP distribution for orders is strongly
underdispersed relative to a Poisson distribution (νO
= 3.85), as expected from inspection
of the data. As a result, the changes in λO
svt have to be fairly large even for small changes
in the mean likelihood of orders. Thus, the size of the ¯κ parameters and the ¯β parameters
cannot be directly compared. But again, as expected, the likelihood of orders increases with
upcoming weekends (¯κ3 = .96), as well as with the season (¯κ2=5.04). However, the effect of
the season is not very consistent across stores and products (ς2 = 29.92); the 95% credible
interval of κ2sv is strictly positive for only 82 of the 170 store-product combinations, and
even strictly negative for seven of them. The posterior means for the hierarchy on κ1sv, the
effect of inventory relative to expected sales, are ¯κ1 = 3.05 and ς1 = .43. Combining them
to get the resulting gamma distribution at each iteration and then averaging over them, the

27
posterior mean and variance of the distribution are 1.29 and .56, respectively.
Stockouts
The average rate of stockouts across stores, products, and days estimated by the model
is 6.37%.4
This rate is below the average rate of about 8% reported by previous studies
(Gruen and Corsten 2008; Andersen Consulting 1996); however, those studies are based on
all instances when no units are on the shelf, whereas our model counts only the instances
in which sales are actually lost. More interestingly, though, we also find a wide range of
stockout rates across stores. Seven of the ten stores actually have an average stockout rate
of less than 6%, the best being only 3.94%; in contrast, two stores have mean stockout rates
of over 10% (10.62% and 11.79%)! This clearly suggests that these stores have significant
potential for improvement. Moreover, combined with the above finding that on average more
than 60% of sales are lost due to a stockout, these stockout rates are high enough for the
supplier to be worried about.
The beta distribution describing the probability of stockouts is strongly U-shaped, with
the posterior mean for ¯a0 being -4.46, for τ0 .35, and for bρ .19. For the individual stores, a0s
ranges from -3.90 to -4.87. Moreover, as to be expected, the likelihood of a stockout increases
significantly if inventory is 0 on a given day (a1s from 1.37 to 1.89). Yet, the average stockout
rate for days with zero inventory is only 34.96%. In addition to our definition of stockouts
as instances with lost sales, this is probably mainly due to the fact that our inventory
measure is only a snapshot at midnight every day. Thus, on a day with zero inventory in our
estimation the store may have had inventory (and therefore sales) until just before midnight.
We are confident that the estimated effect of zero sales on the occurrence of stockouts would
be stronger if data on shorter time intervals were available. However, since we have only
daily data, we cannot confirm this conjecture. Finally, the random shocks also affect the
probability of stockouts, with higher demand shocks leading to a higher chance of a stockout
(a2s from 6.99 to 17.98).

28
Somewhat surprisingly, only 4.46% of the stockouts are store stockouts, while the re-
mainder are shelf stockouts. This rate of store stockouts is significantly lower than what has
been reported in the literature so far (Gruen and Corsten 2008). This may partly be due
to short lead times, relative to case-size and average daily sales. Given that daily sales are
only a fraction of case-size for most products, a store may be able to order and receive more
units before running out of units completely in most instances. However, it may also again
be due to the daily snapshot nature of inventory. Inventory estimates are for midnight of
each day, while stockout estimates are for any time during the day. Thus, if a store had zero
inventory and a stockout during the day but received a new shipment in the evening, this
would currently be counted as a shelf stockout. Again, though, this is not a limitation of
the model, but rather of our data.
For the individual store-product combinations, we find that the relative variance of sales
(defined as the ratio of the variance of sales to the mean of sales) is related to both the
amount of stockouts as well as the ratio of store to shelf stockouts. Not surprisingly, the
higher the relative variance, the higher the stockout rate (correlation coefficient = .343).
Obviously, it is easier to avoid stockouts when sales are more regular. Similarly, the ratio of
store to shelf stockouts increases with higher relative variance of sales (correlation coefficient
= .299), i.e. a larger share of the stockouts are not due to poor shelf replenishment, but to
the lack of units in the store. Due to the higher relative variance, it is more likely that an
unexpected spike in demand occurs that depletes inventory before new shipments can arrive
(recall that the lead time for orders is 2 days).
To illustrate the results of the model, we refer back to the product and store whose
observed part of inventory was depicted in Figure 5. Figure 7 shows the corresponding sales
pattern and the daily mean likelihood of stockout as estimated by the model. As a simple
inspection of the sales pattern may have already suggested, the model finds that a stockout
was most likely during the few days before day 100 with several zero sales days in a row
(not surprisingly, the model also estimates a high likelihood of zero inventory for those days,

29
leading to store stockouts). The results are also reasonable in that in general stockouts are
more likely for days with lower sales. However, the model has better discriminating abilities
than these simple intuitions. For instance, the estimated likelihoods of stockouts for days
with zero sales is on average 60%, but ranges from only 15% to an almost certain 95%.
For days with sales of one unit, stockouts are significantly less likely (25% on average), yet
specific days still have likelihoods of stockouts of over 75%. Even days with two or more
sales may still have likelihoods of over 25%, i.e. on some days with more sales stockouts are
estimated to be much more likely than on some days with no sales at all.
Shrinkage
As for stockouts, we find a fairly wide variation of shrinkage levels across the stores.
Posterior means for bu,s range from .26% to 5.65%, i.e. from less than one incident of unit-
shrinkage per year to more than 20 incidents per year. Similarly, for bc,s the means range
from .10% to .56%; so while case shrinkage is reasonably unlikely for all stores, it is more
than five times as likely in some stores than in others! Moreover, the posterior means for
bu,s and bc,s are strongly correlated (correlation coefficient = .61), showing that the same
stores that have problems with unit shrinkage also have problems with case shrinkage. The
correlations between case and unit shrinkage range from .0666 to .3195.
The parameters for the CMP-binomial distributions for shrinkage given that an incident
occurs are as follows: First, the likelihood of shrinkage for one case/unit is almost indepen-
dent of any other case/unit, as evidenced by the posterior means for νu and νc being .9971
and 1.0044, respectively. The marginal probability of each available unit to be shrunk, pu,s,
ranges from 1.11% to 3.42% across stores (remember that the +1 in equation 14 implies that
at least one unit is shrunk if an incident happens). Similarly, the marginal probability of
each available case to be shrunk ranges from 4.32% to 5.31%. In comparing those rates, one
should keep in mind that at any day there are less cases that can be shrunk than individual

30
units, i.e. these values do not necessarily imply that shrinkage of two cases is more likely
than shrinkage of two units, conditional on occurrence of the respective type of shrinkage.
Finally, as an example of how the particle filter estimates shrinkage, we again refer back to
the store and product depicted in Figure 5. Remember that we believed that true inventory
was likely zero at the flat line before day 100, implying at least some sort of shrinkage. While
the particle filter estimates essentially no case shrinkage up to the flat line, the probability
of case shrinkage increases to almost 30% on day 98 and to about 20% on day 99. After
that, the likelihood for case shrinkage decreases sharply again, yet remains slightly elevated
for the next 10 days relative to the rest of the days. Overall, we estimate at least one case
of shrinkage for this particular store and product in more than 80% of all iterations of the
MCMC. About 30% of the draws result in two cases of shrinkage. Similarly, the likelihood of
unit shrinkage increases slightly before day 98 and then spikes very clearly on day 98. This
shows how the stronger constraint on inventory can help with the estimation of the timing
of shrinkage.
Similarly, for the example given in Figure 6 we expected to observe shrinkage between
day 8 and the flat line starting at day 15. In almost 70% of the draws of the MCMC we
observe unit and/or case shrinkage in that one week, and in more than 85% of the draws in
the time from day 8 to day 20 (as the first few instances of zero sales may also have been
coincidences).
Limitations
As mentioned above, the inventory model assumes that orders are the result of a pull-
strategy, i.e., the amount ordered is inversely related to the current inventory. For this
reason, we excluded some data for which this assumption was clearly invalid (see section
on data preparation), and then estimated the missing data. However, for a few of those
store-product combinations, the assumption seems not to have been satisfied even before
this apparent change in ordering strategy. For those instances, the correlations between the

31
relative inventory measure we use in equation 21 and the placed orders are close to zero,
i.e. orders are independent of inventory (if lead time was indeed two days). With orders
being independent of inventory, there is nothing in the model to prevent very high levels of
inventory (since κ1sv will tend towards zero). In general, this should not be a problem since
the other parameters in equation 21 should be calibrated to result only in appropriately
low amounts of orders. However, keep in mind that (1) the data we excluded was at the
end of the time-series and (2) that the variable for the seasonality increases with time but
shows relatively little variation earlier in the time-series (see Figure 4). Due to the little
variation of seasonality for the days with non-excluded data, a wide range of values for the
coefficient associated with seasonality in equation 21 (i.e., κ2sv) can fit the observed data;
yet, highly positive values lead to high estimated orders for the excluded data. Once high
orders are estimated for the excluded data (i.e., at the end of the time-series), this reinforces
the positive estimates for κ2sv. With orders not depending on inventory levels, we therefore
estimate inventory levels for the last few days in those instances up to double of what we
observe earlier in the time series, which seems unlikely.
This issue is also the reason for specifying a gamma hierarchy for κ1sv in equation 23.
This ensures that the effect of inventory on orders can at least not be positive. Otherwise,
the effect described above could spiral even further out of control. In particular, the higher
estimated orders despite high inventory levels would lead to positive estimates for κ1sv.
However, once κ1sv is positive, the higher inventory levels result in higher estimates for
orders, which results in higher inventory levels, etc.
Thus, it is critical for the performance of the model that the ordering process in fact
follows the assumed pull-strategy. Moreover, knowing the exact lead times between orders
and shipments may be crucial. For instance, it is possible that orders did follow a pull-
strategy even for the discussed store-product combinations, but that the lead-time was not
two days, as for most store-product combinations and as assumed in our model, but one
or three days. This could be enough to cause this issue. Thus, we believe that with more

32
accurate information, this problem could be avoided. Moreover, note that this is not an issue
with complete data, i.e., if there is no missing data to be inferred.
DISCUSSION
Based on the results of our model, the concern of the supplier who provided us with the
data seems justified. At least for some of the stores, the stockout rate is significantly higher
than what is acceptable to a supplier. Both the very small amount of store stockouts and
the fact that other stores have significantly lower stockout rates suggest that this is a store
management (i.e., restocking) issue rather than a problem with timely re-ordering or with
the supply chain. This is useful information to the supplier, as now he can address those
stores specifically.
The model is also useful to the regional or national management level of the grocery
chain, as it now can distinguish well-managed from not so well-managed stores. Choosing
the right amount of inventory is a tricky decision; more inventory leads to higher inventory
holding costs, but lower inventory leads to higher chance of stockouts. Foo (2007) reports
that companies may classify products into different tiers and set target service levels (or
alternatively target stockout rates) for each of them. Obviously, managers choose those
levels they consider to be profit maximizing. Therefore, any deviation (no matter in which
direction!) is detrimental to profits. With our model, management can check whether the
individual retail stores adhere to the pre-specified values.
For further research, it would be interesting to conduct the same analysis for different
products in the same stores to see whether the stockout problems in the less well-managed
stores persist throughout categories. Alternatively, it could be due to the misalignment of
interests between suppliers and retailers mentioned in the introduction, i.e. the retailer has
less incentives to avoid stockouts if there is an alternative brand and switching costs are low.
In this case (1) the stockout rates for the alternative brand should be low (or stockouts of

33
the two brands at least not overlap) and (2) stockout rates for categories with high brand
loyalty should be low.

34
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36
FOOTNOTES
1
All statistics reported in this section include only the used data. See next subsection
for more information.
2
This implies that these shocks are also estimated for the missing store-product combi-
nation. We therefore simply treat the complete time series as missing data to be estimated
in the MCMC.
3
We focus on the most interesting results concerning stockouts and shrinkage in the main
body of the paper. See the appendix for results of the remaining parameters.
4
For the stockout rates, we report the means of the auxiliary variable zsvt indicating the
occurrence of a stockout, rather than the mean of ρsvt; however, results are very close.

37
Table 1: Mean Daily Sales and Shipments by Store
Store 1 2 3 4 5 6 7 8 9 10
Sales 5.36 2.49 6.02 5.70 5.67 2.18 4.09 7.58 6.92 4.62
Shipments 5.28 2.50 6.07 5.68 5.70 2.10 4.00 7.48 6.89 4.62

38
Table 2: Mean Daily Sales and Shipments by Product
Product Sales Shipments Product Sales Shipments
Sweet Corn 15.34 15.27 Asparagus and Gold&White Corn 3.86 3.95
Broccoli Cut 8.43 8.31 Broccoli/Carrots/Peas/Chestnuts 3.82 3.61
Mixed Vegetables 7.92 7.94 Whole Green Beans 3.63 3.56
Sweet Peas 7.71 7.72 Baby Brussel Sprouts 3.56 3.43
Broccoli/Caulifl./Carrots 6.58 6.61 Corn on the Cob 3.17 3.14
Green Beans 5.87 5.82 South-West Corn 2.55 2.61
Broccoli and Cauliflower 4.76 4.79 Garlic Cauliflower 2.38 2.40
Broccoli Florets 4.28 4.08 Garlic Peas and Mushrooms 2.11 2.15
Asian Medley 4.19 4.27

39
Table 3: Overview of Prior Speciﬁcations
Sales Inventory
¯βm for m = 0, 1, ..., 6 N(0,100) bu,s ∀s Beta(1,20)
σm for m = 0, 1, ..., 6 Inv. Gamma(2.5,2.5) bc,s ∀s Beta(1,30)
β7 N(0,100) ηs ∀s Uniform(-1,1)
ω|S, V MVN(.001, .001S ⊗ V ) pu,s ∀s Beta(1,20)
S Inv. Wish.(150, 4 · RS) pc,s ∀s Beta(1,20)
V Inv. Wish.(150, 4 · RV ) νu Gamma(11,.1)
νD
Gamma(2,1) νc Gamma(11,.1)
δsv ∀s, v Beta(2,2.5) πsvT∗
sv
∀T∗
sv and ∀s, v Beta(dsvT∗ , 1)
¯a0 N(-2.5,.5) ¯κl for l = 0, 2, 3 N(0,100)
τ0 Inv. Gamma(4,2) ςl for l = 0, 2, 3 Inv. Gamma(2.5,2.5)
a1sv ∀s, v Gamma(30,.125) ¯κ1 Gamma(4,.5)
a2sv ∀s, v Gamma(10,3) ς1 Gamma(4,.5)
bρ Gamma(2,1) νO
Gamma(2,2)
INV0,sv DT0N(M, (.75M)2
) ∀s, v

40
Table 4: Posterior Means and Asymptotic Standard Errors (in Parentheses) for Sales and
Orders
Sales Model Inventory Model
νD
.5300 (.00076) νO
3.8515 (.00309)
¯β0 .6156 (.00134) ¯κ0 2.7659 (.00951)
¯β1 -.3111 (.00329) ¯κ1 3.0549 (.03166)
¯β2 .2573 (.00028) ¯κ2 5.0405 (.00939)
¯β3 .5474 (.00159) ¯κ3 .9624 (.00094)
¯β4 -.2651 (.00132)
¯β5 .0849 (.00102)
¯β6 -1.2731 (.00502)
σ0 .1847 (.00034) ς0 2.5931 (.03322)
σ1 .3877 (.00397) ς1 .4301 (.00443)
σ2 .0355 (.00003) ς2 29.9161 (.16736)
σ3 .6248 (.00173) ς3 .1348 (.00038)
σ4 .1408 (.00055)
σ5 .1120 (.00037)
σ6 .3391 (.00389)

41
Figure 1: Overview of the Data Generating Process (Observed Data in Bold)

42
Sales
Frequency
0 10 20 30 40 50 60
0100020003000
Figure 2: Histogram of Sales

43
Day
Sales
0 50 100 150
0510152025
Figure 3: Sales Pattern for the Product with Promotion at the Beginning of the Time-Series

44
Day
Sales
0 50 100 150
1234567
Average Sales
SEASON
Independence Day
Labor Day
Thanksgiving
Figure 4: Trend of Average Sales Volume

45
0 50 100 150
0102030405060
Day
Inventory
Figure 5: Observed Part of Inventory for one Product at one Store

46
Day
Inventory
0 50 100 150
010203040
Day
Sales
0 50 100 150
0246810
Figure 6: Observed Part of Inventory and Sales for one Product at one Store

47
Day
Sales
0 50 100 150
05101520
Day
LikelihoodofStockout
0 50 100 150
0.00.20.40.60.8
Figure 7: Sales Pattern and Estimated Likelihood of Stockouts for one Product at one Store

2012 e4

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