INTRODUCTION
Amyotrophic lateral sclerosis (ALS) is a neurodegenerative disorder characterized
by upper and lower motor neuron degeneration. It features considerable clinical heterogeneity
in its phenotypic presentation, as well as in the survival.[1]
[2]
[3] Older age, bulbar onset and rate of symptom progression have been consistently reported
as leading to a worse outcome.[4]
More recently, models have been developed to provide an individualized prognosis and
identified prognostic factors.[5] Relatively few studies[6]
[7]
[8]
[9] have been performed in South American about survival and prognostic factors due
to difficulties in epidemiological data retrieval. A population-based study about
the incidence and prevalence of ALS in Uruguay has been conducted,[10] but with no analysis of these factors. The presence of censored survival times to
the right in the follow-up of ALS patients is well known. However, the consequences
of diagnostic delay and inaccurate knowledge of the onset of the disease have not
been taken into account so far.
These characteristics affect the follow-up time from the left and constitute a non-random
selection that produces a truncation bias on the left. The consequence of left truncation
is a bias in the inclusion of patients or the observation of a shorter follow-up time.[11] ([Figure 1])
Figure 1 Types of left truncation. Type-I left truncation can affect the representativeness
of the sample, while type-II affects the exposure time.[11]
The aim of the present study is to describe survival of a cohort of ALS patients in
Uruguay and to evaluate the prognostic role of selected clinical indicators by exploring
statistical analysis strategies.
METHODS
The current is a single-cohort observational study that included 166 patients with
a diagnosis of probable or definite ALS, between January 1, 1996, and December 1,
2006.
From January 1, 2002, until December 31, 2003, a prospective study was conducted considering
the entire Uruguayan population and included patients with probable or definite ALS.
Multiple sources of recruitment (neurologists, neurophysiologists, hospital registries,
the pharmaceutical industry, and death certificates) were used. Some of the cases
were prevalent, with a diagnosis of the disease prior to 2002, while others were incident
cases diagnosed between 2002 and 2003. In the remaining time, between 2004 and 2006,
only patients with ALS referred to the neuromuscular center of our hospital were included,
which were all incident cases.
As aforementioned, recruitment began in 2001, including patients diagnosed with probable
or definite ALS according to the revised El Escorial criteria. In all cases, the patients
were examined and followed up prospectively by neurologists from the ALS research
team to confirm the diagnosis. No primary lateral sclerosis or progressive muscular
atrophy cases were included in the study. The study was approved by the Ethics Committee
at the Teaching Hospital, School of Medicine, Universidad de la República, Uruguay.
The ALS patients were divided into 3 groups: group 1 included patients with diagnosis
before 2002 and who were alive on January 1, 2002 (prevalent cases); the patients
in group 2 were diagnosed between 2002 and 2003, with the exhaustive uptake mentioned
within the entire national population (exhaustive incident cases); and group 3 included
patients diagnosed with ALS from January 1, 2004 to December 1, 2006 who were referred
for assessment to the group of neuromuscular diseases at our hospital (non-exhaustive
incident cases).
Informed consent was obtained from all patients. During December 2016, the dates of
death of the patients were updated through telephone calls to their relatives, and
death certificates were reviewed.
The prognostic variables studied were: age, sex, region of onset (bulbar or spinal),
predominantly upper motoneuron (UMN) or lower motoneuron (LMN) signs, time of diagnostic
delay (time between onset of symptoms and diagnosis of disease), score on the Amyotrophic
Lateral Sclerosis Functional Rating Scale (ALSFRS), riluzole treatment, and disease
progression.
The rate of progression was calculated as the ratio of (40 - ALSFRS at the time of
first examination)/(time period from the symptomatic onset at the time of first examination
in months). Survival analysis was performed using the Kaplan–Meier method, the log-rank
test, and the Cox model.
Different survival multivariate models were applied:
-
Model I, or classic analysis: follow-up was considered as starting on the date of
diagnosis; patients are at risk of dying from the diagnosis.
-
Model II, or stratified analysis: follow-up was considered as starting on the date
of diagnosis, but the model was stratified by diagnostic delay (time between onset
of symptoms and diagnosis, categorized in quartiles).
-
Model III, or analysis considering truncation and risk of death from the onset of
symptoms: follow-up was considered as starting since the onset of symptoms, and the
model also considers the diagnosis date to enroll patients in the study.
-
Model IV, or analysis considering truncation and risk of death since birth: follow-up
was considered as starting since birth, and the model also considers the date of diagnosis
to enroll patients in the study.
Model II does not directly consider the diagnostic delay, but it differentiates the
presence of patients with a longer evolution time of the disease.
Models III and IV consider the truncated nature of ALS follow-up, because the time
before diagnosis is generally underestimated and not incorporated into the analysis.[11] We have decided to compare analysis strategies with risk origin at the onset of
symptoms or at birth, but with study entry on the date of the diagnosis.
The variables included in the Cox model were chosen according to statistical and clinical
criteria. The median follow up period was calculated through the reverse Kaplan-Meier
method.[12] A significance level of 5% was adopted for the statistical tests. Calculations were
performed using the Stata 16.1 software (StataCorp LLC).
RESULTS
Our cohort included 166 patients, 38% of them women. The mean age at onset was of
58.7 ± 12.4 years. The site of spinal onset was identified in 67% of the sample. Predominant
LMN involvement was observed in 67.5% of the patients. There were 6 cases of familial
ALS (3.6%). The median diagnostic delay in months was of 10.13 and the median overall
follow-up was of 13.6 years.
As for the survival analysis, [Figure 2A] shows the overall survival of the sample since diagnosis, which was of 23 months;
and [Figure 2B] shows the overall survival since symptomatic onset, which was of 37 months. When
calculating survival medians by subgroups within the cohort ([Figure 2C]), group 1 (prevalent cases) presented a survival of 33 months, while group 2 (exhaustive
incident cases), of 22 months, and group 3 (non-exhaustive incident cases), of 14
months. The difference was statistically significant (p = 0.007) regarding the calculation since the date of the diagnosis.
Figure 2 (A) Overall Kaplan-Meier survival estimates (survival since diagnosis); (B) overall Kaplan-Meier survival estimates (survival since symptom onset); (C) Kaplan-Meier survival estimates (since diagnosis) according to the recruitment group.
No statistical differences were found for survival when comparing the men and women
in the cohort using the bivariate analysis of survival and considering the time since
diagnosis (p = 0.76).When analyzing survival according to age groups, subjects younger than 45
years presented a statistically significant higher survival (p = 0.033). The median survival was of 41 months for patients younger than 45 years,
of 22 months for those aged between 45 and 64 years, and of 18 months for the group
aged 65 years or older.
According to the site of onset, patients with spinal forms presented a longer survival
compared to those with bulbar forms, which was statistically significant (24 versus
21 months respectively; p = 0.043). Cases of predominant UMN involvement presented a longer median survival
compared with patients with predominant LMN features (33 versus 20 months respectively;
p = 0.035). There was no statistically significant difference in survival between patients
treated or not with riluzole. When assessing survival according to diagnostic delay
(longer or shorter than 10 months), no statistically significant difference was found
between the two groups (p = 0.059).
The median progression rate for the entire cohort was of 0.8. Therefore, when analyzing
the survival rate according to the progression rate, an inverse relationship is described:
31 months for those with a lower progression rate and 15 months for those with a higher
rate (p = 0.001). Progression rate, riluzole use, and survival were the only variables with
statistically significant differences in the analysis of the three subgroups described
([Table 1]). [Table 2] presents the results of the Cox models for the different multivariate analysis strategies.
Table 1
Comparison of the clinical and demographic variables of the three groups of patients
|
Factor
|
Group 1 (n = 48)
|
Group 2 (n = 88)
|
Group 3 (n = 30)
|
Total (n = 166)
|
p-value
|
|
Female sex % (n)
|
39.6% (19)
|
34.1% (30)
|
46.7% (14)
|
38.0% (63)
|
NSDμ
|
|
Mean age at onset (years)
|
57.9 ± 13.1
|
59.3 ± 12.2
|
58.4 ± 12.3
|
58.7 ± 12.4
|
NSDχ
|
|
Bulbar onset# % (n)
|
25% (12)
|
33% (29)
|
46.7% (14)
|
33.1% (55)
|
NSDμ
|
|
Lower motoneuron## % (n)
|
68.9% (31)
|
67.1% (57)
|
66.7% (20)
|
67.5% (108)
|
NSDμ
|
|
Riluzole treatment: % (n)
|
52.1% (25)
|
11.4% (10)
|
10.0% (3)
|
22.9% (38)
|
< 0.0001μ
|
|
Mediuan diagnostic delay (months)
|
12.7
|
10.13
|
9.65
|
10.13
|
NSD€
|
|
Median progression rate
|
1.1
|
0.7
|
1.3
|
0.8
|
0.043€
|
|
Median survival since diagnosis (months)
|
32.9
|
22
|
14
|
23
|
0.007★
|
Abbreviation: NSD, no significant difference.
Notes: #Bulbar versus spinal onset; ##lower motoneuron versus upper motoneuron presentation; χanalysis of variance; μChi-squared test; €Kruskal-Wallis test; ★Kaplan–Meyer method.
Table 2
Results of the Cox models for the different analysis strategies
|
Model I
|
Model II
|
Model III
|
Model IV
|
|
Hazard ratio; 95%CI
|
p-value
|
Hazard ratio; 95%CI
|
p-value
|
Hazard ratio; 95%CI
|
p-value
|
Hazard ratio; 95%CI
|
p-value
|
|
Female sex
|
0.74; 0.50–1.10
|
0.14
|
0.77; 0.51–1.16
|
0.20
|
0.70; 0.47–1.04
|
0.078
|
0.83; 0.53–1.29
|
0.405
|
|
Age at onset
|
1.01; 1.00–1.03
|
0.08
|
1.02; 1.00–1.03
|
0.04
|
1.01; 0.99–1.03
|
0.147
|
1.04; 0.96–1.12
|
0.344
|
|
Bulbar onset
|
1.46; 0.97–2.20
|
0.07
|
1.57; 1.01–2.46
|
0.04
|
1.63; 1.07–2.48
|
0.024
|
1.26; 0.79–2.02
|
0.335
|
|
Lower motoneuron
|
1.68; 1.11–2.55
|
0.01
|
1.65; 1.06–2.58
|
0.02
|
1.56; 1.02–2.38
|
0.038
|
1.59; 1.02–2.49
|
0.041
|
|
Riluzole treatment
|
0.65; 0.39–1.07
|
0.09
|
0.50; 0.29–0.85
|
0.01
|
0.57; 0.34–0.95
|
0.031
|
0.75; 0.43–1.31
|
0.316
|
|
Progression rate
|
1.25; 1.14–1.37
|
< 0.01
|
1.40; 1.23–1.59
|
< 0.01
|
1.39; 1.24–1.55
|
< 0.01
|
1.22; 1.07–1.39
|
0.004
|
DISCUSSION
The data herein presented are the first ever produced on survival of a representative
cohort of Uruguayan ALS patients.
The limitations of the present study include the combination of incident and prevalent
cases, as well as the periods of exhaustive and non-exhaustive uptake of patients.
These factors could constitute biases.
The main strengths are the number of patients included, taking into account that ALS
is a disease with low incidence and Uruguay has a population lower than 3.5 million
inhabitants. Most patients were examined and followed up prospectively, and internationally-accepted
diagnostic criteria were used, including only patients with defined or probable ALS,
with a long follow-up period.
When comparing the three groups of patients, there was no statistically significant
difference among them in terms of the following variables: sex, age, site of onset,
predominant UMN or LMN features, and diagnostic delay.
The difference in the progression rate of the disease among the groups was also statistically
significant. Group 3 presented the highest progression rate, which might be related
to the fact that these patients were referred to the neuromuscular center and may
present more aggressive forms.
In the case of group 1, there could be a memory bias, in which the symptomatic onset
dates were not reliable, and they were run to the right, explaining a higher rate
of progression when compared to group 2.
However, when comparing survival medians among the subgroups of patients, the highest
survival rate (of 33 months) was presented by the prevalent cases. This could be due
to the absence of poor prognostic factors in these patients who were alive at the
beginning of the prospective study. The incident cases presented an intermediate survival,
similar to that reported for the entire population. Group 3 presented the lowest survival,
which could be linked to a more aggressive disease.
The mean age of ALS onset was of 58.7 years, comparable to what is reported in the
meta-analysis by Marin et al.[13] In studies conducted in Latin American countries,[9]
[14]
[15]
[16] the mean age at onset was of 47.5 in Mexico, of 53.6 years in Brazil, of 54 in Ecuador,
and of 53 years in Cuba, somewhat lower than the mean age in the current study. Consistent
with other studies,[17] 33% of the patients presented bulbar onset of the disease.
The median diagnostic delay was of 10 months, somewhat shorter than is the ranges
reported in the literature,[18] from 8 to 15 months or longer. Several studies[19] have focused on the diagnostic delay in ALS, with little change over the last years,
even after the creation of multidisciplinary ALS centers. In the present study, the
research team actively sought cases, which may have decreased the diagnosis time.
The median survival was of 37 months since the first symptoms and of 23 months since
ALS diagnosis. Some authors compare survival between population studies and in specialized
ALS centers: while some[20] do not find statistical differences, others[21]
[22] report a longer survival in patients cared for in multidisciplinary centers.
A recent study[16] that compares three cohorts of ALS patients in Cuba, Uruguay, and Ireland shows
no differences in survival among the populations. Therefore, we can conclude that
the general survival values in our population are comparable with those of population-based
studies.[23]
[24]
[25] In line with other studies,[4]
[26] a younger age at ALS onset in our patients was significantly associated with better
survival. There was no significant difference in the survival between male and female
subjects in the present study, which is similar to other reports.[4] Bulbar onset of symptoms and predominant LMN involvement were associated with a
shorter survival, which is in agreement with previous population-based studies.[25]
Riluzole use did not show statistical difference in the bivariate analysis of survival.
Although there are reports[4]
[27] that the use of riluzole improves mortality by up to 23% and 15% at 6 and 12 months,
respectively, other authors[28] have found that riluzole therapy was associated with a survival advantage only in
certain stages of the disease.
Borderline significant differences were found in terms of median survival time since
diagnosis after stratifying patients according to the median interval between onset
and diagnosis. Authors[13]
[29] have found that a longer delay leads to a better prognosis, especially with diagnostic
delays of 12 months or longer. Previous studies, as well as the current one, show
that the rate of disease progression appears to be a sensitive clinical prognostic
factor in ALS.[30]
[31]
Truncation is a type of incomplete observation described in survival studies, and
it is linked to the event that defines the subject's “entry” into the study (onset
of the disease). It is completely different from censoring (the other type of incomplete
observation in survival studies), in which the event involved is the dependent variable
(in this case, death).
We propose that ALS, such as other neurodegenerative diseases, has a particular form
of left truncation (type-II truncation), since we do not know the time of the initial
event or risk onset. As can be seen in [Figure 1], type-I left truncation can affect the representativeness of the sample, while type-II
left truncation affects the exposure time. In the case of a neurodegenerative disease
such as ALS, in which the time of onset of the disease is unknown, this type-II truncation
affects all cases.[11]
Taking into account these previous aspects, different multivariate survival analysis
strategies were planned, including two models (III and IV) that consider data truncation.[11] Models I and IV showed similar results, since they coincide in that the predominant
LMN features and the rate of progression are the only significant variables. Given
that model I does not consider left truncation makes us suspect that model IV does
not adequately comply with the correction of temporal information bias. On the other
hand, model II, which stratifies diagnostic delay, recognized a greater number of
risk predictors, all well known as risk factors. Similarly, model III recognized bulbar
onset, the LMN form, riluzole treatment, and progression rate as prognostic factors.
Models II and III seem to better recognize the prognostic factors of mortality in
ALS patients, but it is model III that presents a strategy that incorporates the time
before the diagnostic date and allows an analysis that considers left truncation.
It is important to highlight that the retrospective collection of data, such as the
date of symptom onset and onset site, is a disadvantage of the current study and of
many others that analyze the behavior of this disease. Some symptoms may correspond
to other medical conditions, due to the coexistence of morbid states, especially in
the elderly. The site of the onset of symptoms also has an impact, since those corresponding
to the bulbar level are usually noticed before those of spinal origin.
By considering the diagnosis date, one may obtain greater precision regarding different
prognostic factors. However, the moment a subject first comes under observation usually
does not coincide with the time when the subject becomes at risk of developing the
disease of interest.[32]
The first ALS symptoms generally appear when a significant part of the motor system
has degenerated, and they probably reflect the loss of compensation mechanisms after
a presymptomatic period or subtle symptoms that were not recognized or underestimated.[33] Therefore, it is possible that considering the symptomatic onset enables a closer
approximation to the initial event with respect to the moment of diagnosis. However,
survival analysis studies in ALS generally do not cover the period from the initial
event of the disease until the diagnosis.
The multivariate analysis shows that age at onset as a continuous variable loses value
as an independent prognostic factor, being only a statistically significant variable
when Cox regression is performed stratified by diagnostic delay. Sex continues to
be a non-significant variable, as in the bivariate analysis. Regarding the clinical
form of presentation, it behaved as an independent prognostic factor in all types
of analysis, with shorter survival in cases of predominant LMN involvement. The site
of the onset was also a significant variable in terms of survival, especially in the
analysis that considers the time since the symptomatic onset. As indicated by other
studies,[34] the clinical form is positioned with greater weight in terms of survival with respect
to the site of onset, since it is a more reliable factor based on clinical history,
neurological examination, and the prospective follow-up of the case.
The use of riluzole appears as a statistically significant prognostic factor in the
multivariate analysis, considering the onset and diagnosis dates as well as stratifying
by diagnostic delay. The progression rate behaved as an independent prognostic variable
in all types of multivariate analysis. Therefore, there are clinical presentations
with higher or lower progression beyond the age at onset, clinical features or site
of onset of the disease. Hence, the clinical perception is that, beyond the classical
prognostic factors, there are others that seem to influence the progression of the
disease and that we are not measuring.
We think that bulbar onset and LMN forms are variables associated with a shorter survival,
but their impact or the strength of the association between the variables depends
on whether the time of symptomatic onset or diagnosis is considered, given that the
diagnostic delay in the cases of bulbar onset and LMN forms is significantly shorter.
The same thing occurs with UMN forms of bulbar onset, which usually present with a
prolonged symptomatic history prior to the diagnosis of ALS, with the survival analysis
being different if the onset of symptoms or the date of diagnosis is considered as
the starting point.
We can conclude then that different analysis strategies enable us to observe that,
while some prognostic variables, such as age at onset, use of riluzole, and site of
onset, presented greater variation; others were systematically retained, as the motoneuronal
clinical form and the rate of progression, which showed a great strength of association
with survival in ALS.
Considering the difficulty in measuring time intervals in the diagnostic pathway of
ALS, prospective studies will ideally be performed in order to more accurately measure
the diagnostic delay. However, even in these types of studies the diagnostic delay
remains high, being a problem inherent to those neurodegenerative diseases that require
rigorous diagnostic criteria and do not have diagnostic biological markers.
