- TECHNICAL NOTE
- Open Access
Analytical strategies for the marble burying test: avoiding impossible predictions and invalid p-values
- Stanley E Lazic^{1}Email author
https://doi.org/10.1186/s13104-015-1062-7
© Lazic; licensee BioMed Central. 2015
- Received: 25 July 2014
- Accepted: 17 March 2015
- Published: 11 April 2015
Abstract
Background
The marble burying test is used to measure repetitive and anxiety-related behaviour in rodents. The number of marbles that animals bury are count data (non-negative integers), which are bounded below by zero and above by the number of marbles present. Count data are often analysed using normal linear models, which include the t-test and analysis of variance (ANOVA) as special cases. Linear models assume that the data are unbounded and that the variance is constant across groups. These requirements are rarely met with count data, leading to 95% confidence intervals that include impossible values (less than zero or greater than the number of marbles present), misleading p-values, and impossible predictions. Transforming the data or using nonparametric methods are common alternatives but transformations do not perform well when many zero values are present and nonparametric methods have several drawbacks.
Findings
The problems with using normal linear models to analyse marble burying data are demonstrated and generalised linear models (GLMs) are introduced as more appropriate alternatives.
Conclusions
GLMs have been specifically developed to deal with count and other types of non-Gaussian data, are straightforward to use and interpret, and will lead to more sensible inferences.
Keywords
- Autism
- Count data
- Generalised linear model
- Marble burying test
- Poisson
- Pseudoreplication
- Reproducible research
- Valproic acid
Background
Data derived from the marble burying test have several properties that make the use of normal linear models questionable. First, the data often contain censored observations. These occur when animals bury all of the marbles present and therefore it is unknown whether the animals would have buried more marbles had they been available. Thus, the relationship between the underlying construct (degree of anxiety/compulsiveness/digging) and the measured outcome (number of marbles buried) breaks down once all the marbles are buried. Censored values can bias estimates of treatment effects and lead to a loss of sensitivity; for example, a treatment would appear ineffective if animals in both the treated and control group bury almost all of the marbles. If more marbles are used, then the control group may bury more than the treated group. Methods exist for analysing censored data but will not be considered here because simply increasing the number of marbles used or reducing the time that animals have to bury them can remove, or greatly reduce, the number of censored observations. Sugimoto and colleagues counted the number of marbles buried at 10, 20, and 30 minutes and showed how differences at early time points between control and treated groups tend to diminish by 30 minutes [8]. At 10 minutes the lowest dose group (1 mg/kg fluvoxamine; Figure one in their paper) buried nearly three times as many marbles as the control group, but by 30 minutes they buried the same number. Thus, an incorrect conclusion would have been reached if the final time point was the only one examined. Santini also examined multiple time points and differences between groups at 30 minutes were also smaller compared with earlier time points (Figure two A in their paper) [9]. The number of marbles and the time the animals have to bury them should be optimised to ensure that the test has good sensitivity and to avoid censored observations.
Second, the number of marbles buried are counts, which means they are non-negative integers (0,1,2,3, …) and therefore have a minimum value of zero. Such data will not follow a normal distribution when the counts are small, but will become increasingly normal as the counts get larger. A normal model assumes that the data can take any real number (−∞, +∞), and there is nothing to constrain impossible values. This is a problem because 95% confidence interval (CIs) and predicted values can include negative values, which cannot occur with count data. For example, when using a normal model the numbers {0, 0, 0, 0, 1, 2, 5, 9} have a mean of 2.12 and a symmetric 95% CI of -0.14 to 4.39. The negative values indicate that a normal model is unsuitable, and the reference distribution used to calculate p-values will contain impossible values, making the p-values invalid.
The third property that makes a normal model dubious is that in addition to a minimum value of zero, there is also a maximum value—the total number of marbles available, making the data bounded from above as well as below. Since a normal model has no constraints on possible values that the data may take, confidence intervals can include values greater than the total number of marbles available. One solution is to design the experiment so that the number of marbles used and the time animals have to bury them produces results that are not too close to the upper boundary (which will also ensure censored observations are unlikely).
One approach to deal with unequal variances and skewness is to transform the data; for example, by taking the square-root or logarithm of the counts. While such transformations often improve the distributional properties of the data, they may not work adequately in all situations, in particular when there are many zeros. Log transformation is not recommended for count data [10], partly because the log of zero is undefined, making it necessary to add a small value to all of the counts.
Another approach often used when the assumptions of a parametric test are not met is to switch to nonparametric methods. Nonparametric tests have a number of undesirable features. First, most nonparametric methods analyse ranks rather than the original values, which results in a loss of information. Second, parameter estimates and their confidence intervals are usually not provided by standard statistical software, only p-values. This means that it is not possible to estimate important values such as a maximum response or the EC _{50} (concentration that gives half of the maximal response) in a dose-response design; the only output is an uninformative “the groups are different (p < 0.05)”. Third, only simple designs can be analysed in a straightforward manner with standard software. Covariates cannot be included (e.g. adjusting for baseline) and it is difficult to include multiple explanatory variables in the analysis and to test for interactions. Some work-arounds can b e employed such as testing groups individually, but then the interpretation is more complex and such analyses tend to be conducted and interpreted inappropriately by neuroscientists [11].
The key message of this paper is that rather than using nonparametric tests or transforming the data to fit a normal model, it is often better to use an appropriate parametric model in the first place. In other words, select a model that fits the data rather than mangle the data to fit a normal model. Generalised linear models (GLMs) were developed by Nelder and Wedderburn in the 1970s and are routinely used in many disciplines [12]. As their name suggests, GLMs are more general versions of normal linear models. GLMs can be used for any standard analysis that assumes normality, equal variances, and unbounded data, but they can also be used for binary (0 or 1), proportion (0 to 1), count (non-negative integers), and positively skewed data. A detailed discussion of GLMs is beyond the scope of this article but can be found in references [13-15]. Hilbe provides an introductory book on the analysis of count data suitable for biologis ts [16], as well as a comprehensive but more technical book [17]. Briefly, GLMs use one of several distributions from the exponential family to describe the data. Examples of distributions from the exponential family include the Gaussian (for normal data), the gamma (for data with a positive skew), the Poisson and negative binomial (for count data), and the binomial (for binary and proportion data). The family is the only option that needs to be specified when analysing the data. Two other features of GLMs include the link function, which describes how the mean of the data is related to the explanatory variables and associated parameters, and the inverse link function, which restricts the predicted values to lie within allowable bounds, such as non-negative values for count data. Each distribution has a default or “canonical” link that is typically used, and therefore does not need to be specified during the analysis (other link functions are available but they will not be considered here). One difference with interpreting the results of GLMs is that the coefficients are not on the original scale of the data. If interpretation of coefficients is desired, then the coefficients need to be back-transformed onto the original scale by applying the inverse link function. For example, the link function for a Poisson GLM is the log-link, and therefore taking the antilog returns coefficients on the original scale (i.e. in units of marbles). GLMs are implemented in all major statistics packages and have few “barriers-to-entry”. An appropriate family for the data needs to be selected, but then analysis proceeds as usual.
Methods
Marble burying data from Mehta and colleagues were used, and the experimental details can be found in the original paper [18]; protocols were approved by the University of Pennsylvania Institutional Animal Care and Use Committees and were conducted in accordance with National Institutes of Health guidelines. Briefly, 14 pregnant mice were randomly assigned to receive valproic acid (VPA; n=9) or a saline injection (n=5). VPA was used to generate autistic-like behaviour in the offspring (n _{offspring}=48). Half of the offspring in each condition were also randomly given MPEP (2-methyl-6-phenylethyl-pyrididine), a metabotropic glutamate receptor 5 antagonist, or saline injections. MPEP was hypothesised to correct some of the features induced by VPA and the number of marbles buried out of 20 was an outcome. There are three hypotheses of interest: (1) does VPA affect the number of marbles buried, (2) does MPEP affect the number of marbles buried, and (3) does the existence or strength of MPEP’s effect depend on whether animals were exposed to VPA in utero (interaction effect).
This data set is more complex than it initially appears because VPA was applied to pregnant females and MPEP to the individual offspring of those females. This is called a split-unit or split-plot design because there are two types of experimental units. An experimental unit is the smallest division of sample material that can be randomly and independently assigned to different treatment conditions. The sample size, or “n”, is the number of experimental units and must be determined correctly to obtain valid p-values. When testing the effect of VPA, the experimental units are the pregnant females (n=14); when testing the effect of MPEP, the experimental units are the individual offspring (n=48) [19,20]. The power to detect an effect of VPA is lower than for MPEP because of the smaller sample size, and ignoring the split-unit treatment structure during the analysis can give both too many false positives and false negatives [21-24]. Generalised mixed-effects models could be used for such data but are beyond the scope of this article [25]. Rodent studies using split-unit designs are becoming increasingly popular as new disease models have been developed that apply an intervention to pregnant females to induce pathology in the offspring [26], and due to recent interest in the epigenetic transfer of paternal traits [27]. Since the split-unit structure is rarely taken into account, the statistical results from these studies are largely uninterpretable [24,28]. The implications of the split-unit design will generally be ignored to simplify the comparison of methods. The results of a generalised mixed-effects model are however reported, both to compare with the other models and for those readers who are interested in the effects of VPA and MPEP.
The data were analysed using four parametric GLMs and one nonparametric analysis, but this does not exhaust all possibilities. The first analysis was a standard 2-way ANOVA with VPA and MPEP as factors and is a specific example of a normal linear model (and equivalent to a Gaussian GLM with the identity link). This can be thought of as the “standard analysis” that would most commonly be used and which the other analyses are compared against. With this model it is assumed that the data can be reasonably approximated by a normal distribution, can theoretically take any value, the variances are equal in all groups, and the responses of the individual animals are independent of each other. The second model assumes that the data can be described by a Poisson distribution (with the default log link), which is appropriate for count data. One assumption of a Poisson GLM is that the variance is equal to the mean. This assumption needs to be verified, much like the assumption of equal variances with a normal model. Assumed mean-variance relationships for different GLMs are shown in Figure 2, along with the relationship observed in the actual data. With count data, it often happens that the variance is greater than the mean, and is referred to as overdispersion. Overdispersion can occur for a variety of reasons, including (1) an important variable has been omitted from the model, (2) an important interaction term was not included, (3) the presence of outliers in the data, or (4) a positive correlation between responses, which might indicate a lack of independence of individual responses [17]. If the data are overdispersed, then a Poisson model will have confidence intervals that are artificially precise and p-values that are too small. An estimate of overdispersion is often provided from the output of a GLM analysis but it can also be calculated by dividing the residual deviance by the residual degrees of freedom (which also should be provided in the output of an analysis), and as a rule-of-thumb, if this value is greater than 1.25 [17] or 1.5 [29], then a Poisson model may be unsuitable. The overdispersion in the Mehta data was estimated to be 2.96, which is well above these thresholds. The Poisson model has provided useful information that cannot be obtained with a standard analysis. Why are the data overdispersed? Was an important variable omitted? This is indeed the case, the overdispersion is the result of litter-effects [24]. Multiple litters were used in this experiment and animals within litters tended to bury a similar number of marbles compared with animals between litters. Overdispersion can be remedied in several ways; if it is due to a missing variable or interaction term in the analysis, then including these will remove the overdispersion, and this should be the first option considered. Another method of dealing with overdispersion is to use a quasi-Poisson GLM, which is the third GLM considered. A quasi-Poisson GLM estimates the amount of overdispersion and scales the standard errors upwards to give the appropriate confidence intervals and p-values; the coefficients are identical to the Poisson model, only the uncertainty in the estimates differ. The fourth GLM is a negative binomial model, which is also used for count data but is more flexible than the Poisson in that the variance can exceed the mean. It is therefore an alternative to a quasi-Poisson model when overdispersion is present. The final analysis uses a nonparametric Wilcoxon rank-sums test. Since it is not possible to calculate an interaction effect, tests between individual groups were conducted, and no correction for multiple testing was used.
The results of the different models are compared in four ways. First, 95% confidence intervals are examined to see whether they include impossible values for parameters. Second, visual predictive checks are used to compare data generated from the model to the original data, to see whether there are any discrepancies [30], pp. 158–163. Third, the goodness-of-fit (GOF) of the models was tested. Fourth, models are compared using the Akaike information criterion (AIC), which trades-off goodness-of-fit with model complexity. Since AIC values may be unfamiliar, they are converted into model probabilities, which sum to one for all of the models under consideration and can be (loosely) interpreted as the probability that a model is the best (higher probabilities are better). Such model comparison methods are discussed in detail by Burnham and Anderson [31,32]. P-values are also examined to see how they differ between analyses. The analyses were performed in R (version 3.1.0) [33] and the code can be found in Additional file 1.
Availability of supporting data
The data supporting the results are included within the article and can be found in Additional file 2.
Findings
Numeric results for different models
Analysis | VPA | MPEP | VPA × MPEP | GOF | AIC | P _{ model } |
---|---|---|---|---|---|---|
Normal | 0.055 | 0.009 | 0.178 | <0.001 | 259 | <0.001 |
Poisson | <0.001 | <0.001 | 0.223 | <0.001 | 256 | <0.001 |
Quasi-Poisson | 0.045 | 0.006 | 0.490 | <0.001 | NA | NA |
Negative binomial | 0.075 | 0.009 | 0.470 | 0.233 | 223 | 1.000 |
Wilcoxon | ||||||
All: 0.602 | All: 0.002 | NA | NA | NA | NA | |
MPEP: 0.494 | VPA: 0.021 | NA | NA | NA | NA | |
Vehicle: 0.138 | Control: 0.044 | NA | NA | NA | NA |
The p-values for the different analyses are listed in Table 1. The first thing to note is the p-values for the nonparametric Wilcoxon test are much larger than for the parametric methods. Also, it was not possible to test for an interaction between VPA and MPEP. The Wilcoxon test was also absent from previous comparisons because it is not possible to simulate data sets or calculate a GOF or AIC, which highlights the shortcomings of nonparametric methods. The p-values from the normal model were similar to the preferred negative binomial model (Table 1), but this will not always be the case, particularly when there are many low counts. All p-values for the effect of VPA in Table 1 are inappropriately small because the sample size was taken to be the number of animals rather than the number of pregnant females, or equivalently, the number of litters [24,28]. The correct sample size can be disregarded when performing a purely statistical comparison of methods, but in order to obtain the correct biological interpretation, a generalised linear mixed effects model was used. This is similar to the Poisson model but correctly accounts for the split-unit nature of the design. While animals in the VPA group buried 1.6 more marbles than animals in the control group, there was no strong evidence for an effect of VPA (p=0.192). The power to detect this effect is low since the (correct) sample size is now only 14. In addition, there were nearly twice as many VPA as control litters, which is less powerful than having an equal number of litters in the two conditions. MPEP reduced the number of marbles buried by 4.1 (p=0.002), and this effect was similar within the VPA and control groups (little evidence for an interaction effect: p=0.203). Another finding from this analysis is that the effect of MPEP varied across litters. On average, animals in the MPEP condition buried 4.1 fewer marbles, but this ranged from 0 to 10 marbles within the different litters. The variation of MPEP’s effect within litters is larger than expected by chance (p<0.001), assuming that the average effect of 4.1 is constant across litters, and any deviation from this value represents sampling variation. The cause of this variation is unclear.
One may argue that effects are often large and the differences between using a normal, Poisson, or negative binomial model will be small, and the correct overall conclusion will still be reached. It is true that when large effects are present, the correct qualitative conclusion can be obtained (i.e. reject or do not reject the null hypothesis) no matter how inappropriate the analysis. However, if only qualitative results are of interest then a statistical analysis can be dispensed with altogether. The purpose of an analysis is to obtain unbiased estimates of effects and an appropriate measure of uncertainty in those estimates. Models for count data have been developed, are simple to implement and interpret, and available in most statistical packages. These should be the default methods for marble burying data with normal linear models used only if they are justified.
Conclusions
Given that the number of marbles buried are counts, statistical methods that assume normality, equal variances, and unbounded values will often be inappropriate. If a formal statistical analysis is to be conducted, then appropriate methods should be used. Furthermore, the number of marbles used and the time available to bury them should be optimised such that no (or very few) animals bury all of the marbles. This will ensure that censored observations do not bias the estimates and reduce the sensitivity of the assay. These points also generalise to other outcomes that are counts. The American National Institutes of Health (NIH) has recently recognised the need to improve the statistical skills of biologists [40], and expanding one’s repertoire of statistical methods beyond the normal linear model is a useful and straightforward step in that direction.
Declarations
Acknowledgements
The author would like to thank the Siegel lab at the University of Pennsylvania for kindly sharing their data.
Authors’ Affiliations
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Copyright
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