Abstract
Using data from the Hampshire Friendly Society, a sickness insurance institution in southern England, we examine morbidity trends in England between 1870 and 1949. Morbidity prevalence increased between 1870 and around 1890, mainly because of a rise in the average duration of sickness episodes, but after 1890 average durations fell markedly even though the incidence of sickness rose. During the first two decades of the twentieth century, sickness prevalence increased gradually, but this rise was entirely due to the greatly increased duration of claims made by men aged 65 years and over. After the early 1920s, both the incidence and the average duration of sickness claims declined. These trends seem to be measuring ‘objective morbidity’: they vary closely with year-on-year changes in the mortality of men of working age, but do not show any clear relationship with real wages or unemployment. Our conclusions are different from those of earlier research using English sickness insurance data. We believe that one reason for this was a methodological problem with the analysis performed by nineteenth-century actuaries.
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Notes
- 1.
Alternatively, they indicate that a person’s health rendered him or her unable to carry out the duties of their normal employment (Harris et al. 2011, p. 644).
- 2.
The rules of the Hampshire Friendly Society as set out in 1868 used the phrase ‘rendered incapable of gaining his livelihood’ to describe qualifying sickness (Hampshire Friendly Society 1846–77, p. 19).
- 3.
We did compare the trends in sickness incidence using different assumptions and found that they moved in parallel: the choice of assumption did not seem to affect our estimate of the trend.
- 4.
An advantage of the ‘minimum incidence’ assumption is that the difference between the estimates of incidence immediately before and after 1895 is also small.
- 5.
An 11-point moving average seemed to us to offer the best compromise between smoothness and fidelity to the original data. We use the moving averages solely to aid visual interpretation of the graphs.
- 6.
To achieve consistency, we deliberately ignore data for the years from 1895 onwards which indicate that a man had two or more spells of sickness in the same year, and count these as if they were a single spell. In effect, we are transforming the data for the period 1895 onwards so that they are reported in the same way as the data for the period 1870–1894.
- 7.
The effect of the coarser level of detail in the data before 1895 is that the incidence of sickness is underestimated by about 10% compared with the period from 1895 onwards.
- 8.
The real wage series was originally produced by Phelps-Brown and Hopkins (1956). We have preferred this series to more recent variants as it relates specifically to working class men in southern England, into which group most of the HFS members fell. The unemployment data were originally published by Feinstein (1972, pp. T126–T127) and refer to the whole of the United Kingdom (UK). Given the impact of both occupational and regional factors on UK unemployment rates during this period, these statistics may not be an accurate guide to fluctuations in the level of unemployment among members of the Hampshire Friendly Society.
- 9.
Although national insurance was introduced in 1911, the labour market was then severely disrupted by World War I. Our dummy variables assume national insurance started to take effect in 1919 (it was officially introduced earlier but World War I intervened before it could have a widespread impact), and the introduction of state contributory pensions (for workers over the age of 65) took effect in 1926 (Macnicol 1998, p. 214).
- 10.
These measures of mortality fall short of the ideal for our purposes, but in different respects. The Hampshire-specific mortality rate from influenza and bronchitis is geographically a better measure of changes in the disease environment faced by the men in our sample, but includes death rates for infants and children. The national death rates for adult males are a better age match to the men in the sample, but are less geographically focussed. For the mortality data for Hampshire, we only analyse the period 1870–1935 as population data for the late 1930s and early 1940s are likely to be unreliable because of World War II (which led to population movements which were not captured by official statistics as there was no population census in 1941).
- 11.
It is possible that the weak effects of some social and economic covariates (notably unemployment) arise because unemployment rates in Hampshire did not reflect national rates. We have not been able to locate time series of local unemployment rates.
- 12.
Gorsky et al. (2011, pp. 1,781–2) noted that concern that HFS members were using sickness to disguise unemployment was only rarely mentioned in the annual reports of the Society. It might be argued that unemployment itself could lead to ill-health and thus we might expect sickness rates to rise at times of high unemployment. This may be true, but the effect is likely to be too weak to detect in our data, as even in the worst years of the early 1930s, the national unemployment rate did not rise above 16%. Ismay (2015) reminds us that friendly societies were able to exclude from membership individuals known to or suspected to be likely to try to take unfair advantage of being members. She also argues that they fostered a loyalty and a feeling among their members that did much to nullify the moral hazard associated with commercial insurance contracts (although others have suggested that such traditional loyalty became severely strained during the early twentieth century, and Downing (2015) argues that it varied both between societies and between different branches of the same society).
- 13.
Omran’s model has not gone unchallenged. Weisz and Olszynko-Gryn (2010), for example, argued that it is overdetermined by contemporary development theory. Here, however, we are not concerned with what drives the epidemiologic transition, simply with the fact that it involves a shift in the distribution of causes of death.
- 14.
The HFS data do provide information on the causes of episodes of sickness, but unfortunately for our purposes only from 1895 onwards. Although there is some uncertainty about the underlying causes of the decline in tuberculosis mortality, epidemiological thinking both in the early twentieth century and nowadays favours improved isolation of infected cases and hence reduced transmission rates (Newsholme 1908; Wilson 2005) which would lead to a reduced incidence of this disease. Since tuberculosis was a long-lasting condition, this is likely to have reduced the mean duration of sickness episodes as a whole.
- 15.
Of course, the arrival of the Russian influenza may have resulted in greater awareness of the disease and an increased tendency to report it as a cause of death. Our main point, though, is that the Russian influenza heralded a step change in the incidence of mortality from the disease in England and Wales which lasted for at least two decades.
- 16.
They are, for example, included in the standard book of formulae and tables which all actuarial students of the Institute and Faculty of Actuaries use in the professional examinations (Institute and Faculty of Actuaries 2002).
- 17.
We all knew John in different ways, as both friend and colleague, and are delighted to have this opportunity to express our appreciation of him.
This volume has demonstrated the enormous range of John’s sympathies and research interests. Our own work overlapped with his in relation both the history of health and the history of social insurance. So far as health is concerned, we had common interests in relation both to anthropometric history and the history of morbidity. John was instrumental in providing us with an early platform for our work on the sickness records of the Hampshire Friendly Society when he edited a special issue of Social Science History, and our contribution to this volume arises directly from that. We hope it is a contribution which he would have welcomed and approved.
Over the years, we had the pleasure, individually and collectively, of meeting John at several academic gatherings, including meetings of the Economic History Society, the Society for the Social History of Medicine, and conferences on Economics and Human Biology. It was always a pleasure to hear his own work and he was a sympathetic and perceptive analyst of the work of others. He was also great company.
John was a great supporter of early-career researchers. He offered his support as research mentor and referee and was a source of great inspiration, both intellectually and personally. He had a brilliant mind and a very kind soul, and it was a privilege to have known him.
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Acknowledgements
An earlier version of parts of this paper was presented at the Social Science History Association’s 38th Annual Conference held in Chicago in November 2013. The research on which this paper is based was funded by research grant #RES-062-23-0324 from the United Kingdom Economic and Social Research Council.
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Appendix: Analysis of the Apparent Increase in the Duration of Sickness Among the Independent Order of Oddfellows Between 1866–1870 and 1893–1897
Appendix: Analysis of the Apparent Increase in the Duration of Sickness Among the Independent Order of Oddfellows Between 1866–1870 and 1893–1897
Where does the idea that the increase in morbidity in the second half of the nineteenth century arose because of the increasing duration of episodes of sickness come from? In this Appendix, we focus on Watson’s report, since this is the weightiest piece of evidence.
Watson’s analysis (Watson 1903, pp. 38–9 and 143–59) was based on person-years of sickness. He classified the person-years according to duration since the episode of sickness began using the duration categories 0–6 months, 6–12 months, 12–24 months and over 24 months. He then compared the actual amount of sickness recorded in each of the duration categories with the amount which would have been expected on the basis of Henry Ratcliffe’s investigation of the Oddfellows’ sickness experience in 1866–1870 (Table 4.4). There are two key observations from this table.
-
1.
The increase in morbidity between 1866–1870 and 1893–1897 is very largely a consequence of the increase in the amount of sickness experience recorded at durations over 24 months.
-
2.
This increase occurs in all age categories.
We need to explain both these observations.
4.1.1 Sickness Episodes in Progress at the Start of the Investigation
The increase in morbidity at longer durations between 1866–1870 and 1893–1897 was characteristic of all age groups: indeed, it was actually stronger among the younger members than among those aged over 65 years. This matters, because it means that whatever was causing it was affecting all age groups. Explanations such as a replacement of acute conditions by chronic degenerative conditions (Riley 1989, p. 172) are unlikely, as if they were the cause, we should expect the increase in morbidity at longer durations to be concentrated among older members. Riley (1989, p. 192) acknowledges that there was an increase in sickness at all ages and describes this as ‘unsettling’, presumably because it suggests that something other than the conventional epidemiological transition is at work. However, he does not suggest what this might be. Perhaps the same need to posit a cause which would affect all age groups stumped Watson?
It is possible to use Watson’s data to obtain some idea as to the proportion of the apparent increase in the duration of sickness between Ratcliffe’s investigation of 1866–1870 and Watson’s investigation of 1893–1897 which might have been due to the different methods employed by the two men.
Watson provides overall data concerning the amount of sickness observed in his investigation. This was, to the nearest person-year, 52,718 at durations 0–3 months, 12,436 at durations 3–6 months, 11,923 at durations 6–12 months, 12,660 at durations 12–24 months and 45,310 at durations over 24 months, making a total of 135,048 person-years (Watson 1903, p. 141). Consider spells of sickness of durations 0–3, 3–6, 6–12, 12–24 months and 24 months and over. Let the number of spells which last for 24 months or more be l24, and the numbers lasting at least 12, 6 and 3 months be l12, l6 and l3, respectively. Let the total number of spells be l0. Let the person-years of experience in each of Watson’s duration categories be p0-3, p3-6, p6-12, p12-24 and p24+, respectively, and let the mean duration of the spells in each duration category be m0-3, m3-6, m6-12, m12-24 and m24+ years.
Using standard life table methods, we can show that the following relationships hold:
Thus, substituting the total number of person-years in each duration category calculated from Watson’s data, we have
This set of five equations with ten unknowns has many solutions, but there are restrictions on the values of some of the unknowns. We know that there are restrictions on the mean durations of spells in each category. Let us assume that m0-3 = 0.125 m3-6 = 0.375 m6-12 = 0.75 and that m12-24 =1.5 (i.e. that spells under 3 months long are, on average, 1.5 months long; those between 3 and 6 months long are, on average, 4.5 months long; that spells lasting between 6 and 12 months are, on average, 9 months long; and that spells lasting between 12 and 24 months are, on average, 18 months long). Then the five equations become
Since l0 ≥ l3 ≥ l6 ≥ l12 ≥ l24 > 0, then eq. (4.A2) implies that l24 ≤ 12, 660. Substituting this into eq. (4.A1) produces
\( {m}_{24+}\ge 2+\frac{45,310}{12,660}=5.58 \), or that the average duration of spells longer than 24 months’ long is at least 5.58 years.
For simplicity, suppose it is 6 years. With m24+ = 6, we can solve eqs. (4.A1)–(4.A5) to give
With other values of m24+ we obtain different solutions (Table 4.5).
We now need to consider the impact of the difference between Watson’s and Ratcliffe’s treatment of the spells ongoing at the start of the period of investigation. A Lexis chart showing the sickness for a set of spells of more than 2 years’ duration illustrates the situation (Figure 4.7). Calendar time is on the horizontal axis, and duration of spell is on the vertical axis. Suppose these spells last m + 2 years and imagine that claims for these sickness spells are made at a rate which is constant over time. Then the person-years of sickness at durations over 2 years during the period of investigation are represented by the area of the rectangle ABDC. This is the person-years calculation used by Ratcliffe in his 1866–1870 investigation. However, the spells under way at the start of the investigation, which are represented by the vertical line AB, will have durations at the start ranging from just above 0 to just under m years distributed uniformly between 0 and m (because of the constant rate of claims). The person-years of sickness before the start of the investigation which they encompass is represented by the area of the triangle ABE. It is this additional sickness which Watson’s approach brings in.
The ratio between Watson’s sickness prevalence and Ratcliffe’s sickness prevalence is equal to \( \frac{0.5{m}^2+5m}{5m}=\frac{0.5m+5}{5}=\frac{m}{10}+1 \). So, if the average length of spells of over 2 years duration is 6 years (the minimum that Watson’s own figures allow), then, relative to Ratcliffe, he has inflated the sickness in the 24 months and over duration segment by 1.4 times. If the average length is 10 years (by no means impossible given Watson’s data), the inflation factor is 1.8 times., and if the average length is 14 years, it will be 2.2 times. Note also that this inflation factor is the same for all age groups provided m is the same for all age groups.
There will also be some inflation in the shorter duration segments, but since m is much smaller in these, the extent of the inflation will be much less: indeed, it cannot be more than 5% in the 12–24-month category and 2.5% in the 0–6- and 6–12-month categories.
4.1.2 Watson’s 12-Month ‘Off’ Period
Watson also adopted a 12-month ‘off’ period when compiling his tables (Watson 1903, p. 15). This suggests that he treated a new sickness within 12 months of the previous one as a continuation of the previous one. According to Riley (1997, pp. 172–3), this was different from the treatment by Ratcliffe in earlier investigations. Riley points out, correctly, that this means that comparisons of the incidence of claims between Ratcliffe and Watson are therefore not possible (Watson will record a lower incidence than Ratcliffe). He fails to mention, however, that altering the definition of the ‘off’ period will also have an impact on the duration of claims, and confound the comparison of durations between the two surveys. Watson described this 12-month ‘off’ period as ‘moderately long’ (1903, p. 15). By comparison with shorter ‘off’ periods, it will tend to inflate the number of claims of long duration.
It therefore seems that the different treatment of spells in progress at the start of the investigation by Ratcliffe and Watson is likely to account for a substantial proportion of the apparent increase in morbidity at longer durations. Since Watson’s data show a rise of some 2.4–2.6 times (Table 4.4), then the changed methods account for a minimum of 30% (using the minimum possible duration of sickness episodes over 2 years long which Watson’s own data allow) and could account for close to 100% of the increase, especially if the rather extended ‘off’ period used by Watson is also factored into the calculations. Moreover, since the impact of this change in method is not necessarily age specific, the notable and ‘unsettling’ fact that the apparent increase was roughly the same for all age groups suggests that the changed methods might be the main reason.
4.1.3 Watson’s Treatment of Sickness Claims Spanning More Than One Calendar Year
According to Riley, Watson treated a claim spanning more than one calendar year as several separate episodes, the second and subsequent episodes starting on each 1 January. This will artificially inflate the number of episodes and, when comparing the incidence of claims between Ratcliffe’s and Watson’s surveys, will act in the opposite direction to Watson’s 12-month ‘off’ period.
It will, however, tend to change the distribution of claims by duration, as longer claims are more likely to cross the end of the calendar year and hence to be counted multiple times. Its effect is to increase the proportion of longer claims and, again, to make it look as if the mean duration of claims is rising faster than it actually is. Its effect, though, is likely to be fairly small. Assuming an exponential distribution of claim durations such that the mean claim duration is x years, then the impact on the longest duration claims involves multiplying the number of such claims by a factor which is less than or equal to (1 + x). So if x is, say, 0.2 years, it will involve inflating the number of long claims by no more than 20%.
4.1.4 Conclusion
Table 4.4 reports standardised morbidity ratios of between 238 and 264 in 1893–1897 for claims of over 24 months duration compared with 1866–1870. It seems possible, and may be more likely than not, that the majority of this increase is accounted for by the different methods used by Watson and Ratcliffe in their computations. There are three specific differences, and all will tend to mean that Watson inflates the proportion of claims of longer duration compared with Ratcliffe. It is possible, therefore, that the increase in the average duration of claims reported by these actuaries is entirely artefactual.Footnote 17
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Hinde, A., Gorsky, M., Guntupalli, A., Harris, B. (2022). Sickness Experience in England, 1870–1949. In: Gray, P., Hall, J., Wallis Herndon, R., Silvestre, J. (eds) Standard of Living. Studies in Economic History. Springer, Cham. https://doi.org/10.1007/978-3-031-06477-7_4
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